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James IV Association of SurgeonsDepartment of Surgery, University of CalgaryDepartment of Surgery, University of Alberta

Department of Surgery, Western UniversityDepartment of Surgery, University of TorontoDépartement de chirurgie, Université de MontréalDepartment of Surgery, University of SaskatchewanDepartment of Surgery, McMaster University

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Vol. 62, No. 2, April/avril 2019canjsurg.ca

Predictors of dysplastic and neoplastic progression of Barrett’s esophagus

High-risk medications in older patients with trauma: a cross-sectional study of risk mitigation

Prevalence of obstructive sleep apnea in male patients with surgically treated maxillary and zygomatic fractures

Effect of predicted travel time to trauma care on mortality in major trauma patients

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The Division of General Surgery at the Vernon Jubilee Hospital in Vernon, BC, seeks a full-time general surgeon to replace one member of a group of 6 FTE in 2020. Applicants must be FRCSC certi�ed or eligible in 2020. Proof of ACLS, CMPA, and BC licensure mandatory. Post-Fellowship training is not required. Applicants must be competent and willing to perform upper/lower endoscopy and surgeries including but not limited to: breast, colorectal, skin malignancies, gallbladder disease, hernias, acute GI conditions, and minor procedures.

VJH is a close-knit, 160-bed facility with a patient care tower opened in 2011 with 5 operating rooms, a 6-bed ICU, FRCP anesthesia, pediatrics, internal medicine, medical oncology, CT, US, nuclear medicine, and MRI (opening 2019). VJH is af�liated with the University of British Columbia and is a training site for the UBC General Surgery Postgraduate program and the UBC undergraduate medical program. Vernon is a jewel of a small city, dotted by lakes, cradled by mountains in the North Okanagan Valley, surrounded by orchards, vineyards, farms, and a world-class ski hill: Silver Star/Sovereign Lakes. Other surgical specialties at VJH include ENT, obstetrics-gynecology, ophthalmology, orthopedics, and urology.

Applications should consist of a current curriculum vitae and three letters of reference for receipt by July 31, 2019. Incomplete applications will not be considered. Successful short-listed candidates will be contacted in August 2019 for interviews. Please submit applications to:

Dr. H. HwangVernon Jubilee Hospital

2101 32 St., Vernon, BC V1T 5L2Email [email protected]

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Otolaryngologist OpportunityTrail, BC

An opportunity is available for an Otolaryngologist to practice at Kootenay Boundary Regional Hospital in Trail.

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© 2019 Joule Inc. or its licensors Can J Surg, Vol. 62, No. 2, April 2019 73

EDITORIAL • ÉDITORIAL

76 Are morbidity and mortality conferences becoming a lost art? C.G. Ball

77 Les revues de morbidité et mortalité sont-elles un art qui se perd? C.G. Ball

RESEARCH • RECHERCHE

78 Use of the Corail stem for revision total hip arthroplasty: evaluation of clinical outcomes and cost T.J. Wood, M. Alzahrani, J.D. Marsh, L.E. Somerville, E.M. Vasarhelyi, B.A. Lanting

83 Relation between socioeconomic variables and surgical, systemic and radiation treatment in a cohort of patients with breast cancer in an urban Canadian centre J. Li, S.D. Cornacchi, F. Farrokhyar, N. Johnston, S. Forbes, S. Reid, N. Hodgson, S. Lovrics, K. Lucibello, P. Lovrics

93 Predictors of dysplastic and neoplastic progression of Barrett’s esophagus S. Alnasser, R. Agnihotram, M. Martel, S. Mayrand, E. Franco, L. Ferri

100 High-risk medications in older patients with trauma: a cross-sectional study of risk mitigation E. Lester, M. Dykstra, C. Grant, V. Fawcett, B. Tsang, S. Widder

105 Prevalence of obstructive sleep apnea in male patients with surgically treated maxillary and zygomatic fractures S. Lupi-Ferandin, T. Galic, N. Ivkovic, R. Pecotic, Z. Dogas

111 Surgical site infection following abdominal surgery: a prospective cohort study A. Alkaaki, O.O. Al-Radi, A. Khoja, A. Alnawawi, A. Alnawawi, A. Maghrabi, A. Altaf, M. Aljiffry

118 Does oxidized zirconium make a difference? Midterm cohort survivorship of symmetric posterior condyle posterior-stabilized total knee arthroplasty A.R. Demcoe, E.R. Bohm, D.R. Hedden, C.D. Burnell, T.R. Turgeon

123 Effect of predicted travel time to trauma care on mortality in major trauma patients in Nova Scotia G. Tansley, N. Schuurman, M. Bowes, M. Erdogan, R. Green, M. Asbridge, N. Yanchar

REVIEW • REVUE

131 Limberg �ap versus Karydakis �ap for treating pilonidal sinus disease: a systematic review and meta-analysis P. Gavriilidis, E. Bota

Vol. 62, No. 2, April/avril 2019canjsurg.ca

canjsurg.ca

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74 J can chir, Vol. 62, No 2, avril 2019 © 2019 Joule Inc. or its licensors

DISCUSSIONS IN SURGERY DISCUSSIONS EN CHIRURGIE

139 Laparoscopic colectomy: trends in implementation in Canada and globally M. Hoogerboord, J. Ellsmere, A. Caycedo-Marulanda, C. Brown, S. Jayaraman, D. Urbach, S. Cleary

142 Deployment of second-generation resuscitative endovascular balloon occlusion of the aorta for unresponsive hypotension in a polytrauma patient T. Paradis, O. Bekdache, D. Bracco, J. Grushka, T. Razek, D. Lasry, A. Beckett

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COEDITORS CORÉDACTEURS

Chad G. Ball, MD, Calgary [email protected]

Edward J. Harvey, MD, Montreal [email protected]

ASSOCIATE EDITORS RÉDACTEURS ASSOCIÉS

BARIATRIC SURGERYScott Gmora, MD, Hamilton

BREAST SURGERYMuriel Brackstone, MD, London

CARDIOVASCULAR SURGERYMichel Carrier, MD, Montreal Michael Chu, MD, London

CRITICAL CARERaymond Kao, MD, London

DIGITAL MEDIAAmeer Farooq, MD, Calgary

ENDOCRINE SYSTEMJesse Pasternak, MD, Toronto

EVIDENCE-BASED MEDICINEMichelle Ghert, MD, Hamilton Kelly Vogt, MD, London

GASTROINTESTINAL AND COLORECTAL SURGERYMarcus Burnstein, MD, Toronto Jason Park, MD, Winnipeg

GLOBAL SURGERYDan Deckelbaum, MD, Montreal Vanessa Fawcett, MD, Edmonton

GENERAL SURGERYAndrew Beckett, MD, Montreal Hamish Hwang, MD, Vernon

HEPATOBILIARY AND PANCREATIC SURGERYShiva Jayaraman, MD, Toronto Peter Kim, MD, Vancouver

MILITARY MEDICINECarlos J. Rodriguez, MD, Bethesda

ORTHOPEDIC SURGERYGraham Elder, MD, Sault Ste. Marie Karl-André Lalonde, MD, Ottawa Prism Schneider, MD, Calgary

PEDIATRIC SURGERYMary Brindle, MD, Calgary

SPORTS MEDICINEPaul Martineau, MD, Montreal

SURGICAL ONCOLOGYGeoff Porter, MD, Halifax

THORACIC SURGERYRichard Malthaner, MD, London Jonathan Spicer, MD, Montreal

TRAUMA SURGERYWilliam Dust, MD, Saskatoon Rob Leeper, MD, London Sandy Widder, MD, Edmonton

VASCULAR SURGERYApril Boyd, MD, Winnipeg Tom Forbes, MD, Toronto

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The Canadian Journal of Surgery aims to contribute to the effective continuing medical education of Canadian surgical specialists and to provide surgeons with an effective vehicle for the dissemination of observations in the areas of clinical, basic science and education research.Readers can find CJS online at canjsurg.ca.Submission of new manuscripts can be made at http://mc.manuscriptcentral.com/cjs.

Le Journal canadian de chirurgie vise à dispenser une éducation médicale continue efficace aux spécialistes en chirurgie au Canada, et fournir aux chirurgiens un mécanisme efficace pour diffuser les constatations de la recherche clinique, fondamentale et éducative.Les lecteurs trouveront en direct le JCC à l’adresse canjsurg.ca.Nous favorisons l’envoi électronique de manuscrits. Veuillez visiter le http://mc.manuscriptcentral.com/cjs.

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The views expressed in this editorial are those of the author and do not necessarily reflect the position of the Canadian Medical Association or its subsidiaries.

Are morbidity and mortality conferences becoming a lost art?

S urgical morbidity and mortality (M&M) confer-ences have a history that is both deep and central to many of the core principles we hold dear to our

profession. More specifically, well-run M&M confer-ences employ a peer review process that analyzes adverse patient outcomes on multiple levels. This evaluation includes a discussion of the mechanics of the negative event itself; suggestion of modifications in behaviour, analysis, decision-making and/or technique; and �nally system improvements to prevent similar errors from occurring in the future.

The origin of M&M conferences possessed a challenging ignition in 1904 when Ernest Codman famously suggested that surgeon competence must be evaluated and reported in a structured and repetitive manner.1 While this altruistic beginning led to his �ring from the Massachusetts General Hospital, Codman’s ideas were relatively quick to be par-layed into the American College of Surgeons case reporting system for adverse patient outcomes in 1916.1 It could be argued that Codman’s concept reached its zenith in 1983 when the Accreditation Council for Graduate Medical Edu-cation (the American equivalent of our Royal College of Physicians and Surgeons of Canada) mandated the presence of weekly M&M conferences to achieve and maintain accreditation for all surgical residency training programs.1

As many of us travel throughout the world to engage in postresidency surgical fellowships, visit our colleagues’ institutions, pursue mid-career academic sabbaticals, and/or deliver formal visiting professorships, it becomes quickly apparent that the precise delivery of Codman’s M&M con-ferences varies dramatically from site to site.2 In some insti-tutions, M&M conferences are city-wide endeavours that include multiple hospitals sharing their adverse patient out-comes, whereas in others, it remains limited to a small sur-gical subspecialty group discussing patients and diagnoses speci�c to a very narrow slice of medicine. In a few institu-tions, the discussion is focused on CanMEDS learning objectives that are indirect to the event itself. Irrespective of the delivery, however, an environment that is nonpunitive to the individual surgeon, in conjunction with a focus on quality improvement in patient care (i.e., future prevention of a similar error) is essential.

It has been increasingly evident in recent years that the central importance of weekly M&M conferences (and the associated quality-improvement opportunities) seems to be less revered and accepted among today’s generation of sur-geons. While the postulates for this observation are fasci-nating, a recurring theme now includes the impact of “big data” quality improvement opportunities such as the National Surgical Quality Improvement Program (NSQIP). More speci�cally, an increasingly popular narrative is that NSQIP and its comparators have essentially replaced the need for formal small-group M&M conferences. While these large and powerful quality-improvement platforms are rapidly becoming ingrained within our surgical practices for everything from trauma to emergency general surgery to highly selected subspecialty work, the local wisdom and experience available in a typical M&M conference remains critical to both surgeon improvement and peak perfor-mance. As a result, it seems that the future of Codman’s call for the continued evaluation of competence must involve both comparisons of surgeon performance to larger groups of colleagues at the individual and program levels (i.e., via big data), as well as the incorporation of local expertise and sage advice in the form of collegial and real-time discussion at a formal M&M conference.3 In other words, never has the role of the M&M conference been so important to the mission of continued quality improvement in patient care at each of our institutions and hospital systems.

Chad G. Ball, MDCoeditor, Canadian Journal of Surgery

Competing interests: None declared.

DOI: 10.1503/cjs.004119

References

1. Gregor A, Taylor D. Morbidity and mortality conference: its pur-pose reclaimed and grounded in theory. Teach Learn Med 2016; 28:439-47.

2. Campbell W. Surgical morbidity and mortality meetings. Ann R Coll Surg Engl 1988;70:363-5.

3. McNamara DA, Hall HM, Hardin EA. Rethinking the modern cardiology morbidity and mortality conference: harmonizing qual-ity improvement and education. J Am Coll Cardiol. 2019;73:868-72.

edit-april.indd 76 2019-03-19 1:43 PM



Les opinions exprimées dans cet éditorial sont celles de l’auteur et ne représentent pas nécessairement celles de l’Association médicale canadienne ou ses filiales.

Les revues de morbidité et mortalité sont-elles un art qui se perd?

L ’histoire des revues de morbidité et mortalité (RMM) en chirurgie est fort riche et elle est indissociable de plusieurs principes fondamentaux pour notre profes-

sion. Plus précisément, les RMM rigoureuses font appel à un processus d’examen par les pairs qui analyse à divers niveaux les issues défavorables chez les patients. Cet examen inclut une discussion sur la mécanique des complications elles-mêmes et des suggestions de modifications com-portementales, analytiques, décisionnelles ou techniques devant aboutir à l’amélioration du système a�n de prévenir d’autres erreurs similaires.

Les RMM ont débuté de manière laborieuse en 1904 : Ernest Codman s’est fait connaître pour avoir suggéré de soumettre les chirurgiens à une évaluation de leurs compé-tences sous forme de rapports structurés périodiques1. Bien que motivé par un altruisme louable, Codman s’est fait mettre à la porte de l’Hôpital général du Massachusetts pour ses idées, qui n’en ont pas moins assez rapidement suscité l’adhésion de l’American College of Surgeons, soit dès 1916, pour la préparation des rapports de cas lors de complications chez les patients1. On pourrait dire que les idées du Dr Codman ont atteint leur apogée en 1983 lorsque l’Accreditation Council for Graduate Medical Education (l’équivalent américain de notre Collège royal des médecins et chirurgiens du Canada) a exigé la tenue de RMM heb-domadaires pour l’obtention et le maintien de l’agrément de tous les programmes de résidence en chirurgie1.

Nous sommes nombreux à voyager un peu partout sur la planète pour diverses raisons : stages de spécialisation post-résidence, visites des établissements de collègues, sabbatiques académiques en milieu de carrière ou postes de professeur invité, et il devient vite apparent que les RMM suggérées par Codman varient grandement d’un endroit à l’autre2. Pour certains établissements, les RMM sont des événements à dimension urbaine, qui regroupent plusieurs hôpitaux pour des discussions sur les complications rencontrées, tandis qu’ailleurs, elles restent limitées à l’équipe de chirurgie qui discute des cas et des diagnostics spéci�ques à un champ très restreint de la médecine. Dans quelques établissements, la discussion tourne autour des objectifs d’apprentissage CanMEDS, qui ont un lien indirect avec l’événement lui-même. Indépendamment du format, toutefois, il est essentiel de mettre l’accent sur un environnement non punitif pour le chirurgien en tant qu’individu, sans négliger l’amélioration de la qualité des soins (c.-à-d., la prévention d’erreurs similaires).

Ces dernières années, on constate de plus en plus que le caractère fondamental des RMM hebdomadaires (et des occasions d’amélioration de la qualité qui en découlent) sem-ble moins respecté et accepté des chirurgiens de la relève. Les postulats de cette observation sont fascinants, mais on retien-dra désormais un thème récurrent, soit l’impact des occasions d’amélioration de la qualité qui reposent sur les « données massives » (big data), comme le programme national d’amélioration de la qualité des soins chirurgicaux (NSQIP). Plus spécifiquement, selon un scénario de plus en plus fréquent, le NSQIP et d’autres initiatives semblables remplacent désormais les RMM formelles en petits groupes. Même si ces plateformes volumineuses et ef�caces pour l’amélioration de la qualité s’intègrent rapidement à notre pratique chirurgicale peu importe le contexte, de la traumatologue à la chirurgie générale d’urgence en passant par des interventions hautement spécialisées, les connaissances et expériences mises en commun durant les RMM typiques restent cruciales pour l’amélioration des chirurgiens et leur rendement optimal. Il semble donc que l’avenir de l’évaluation continue des compétences proposée par Codman passe obligatoirement par une comparaison du rendement des chirurgiens par rapport à de grands groupes de collègues, tant à l’échelle des individus que des programmes (par le biais des données massives) et par l’inclusion de l’expertise et de l’expérience à l’échelle locale sous forme de discussions collégiales en temps réel dans le cadre de RMM formelles3. En d’autres mots, le rôle des RMM n’a jamais été aussi indispensable à la mission de chacun de nos établissements et systèmes hospitaliers, qui est d’assurer l’amélioration continue de la qualité des soins.

Chad G. Ball, MDCorédacteur en chef, Journal canadien de chirurgie

Intérêts concurrents : Aucun déclaré.

DOI: 10.1503/cjs.004519

Références

1. Gregor A, Taylor D. Morbidity and mortality conference: its pur-pose reclaimed and grounded in theory. Teach Learn Med 2016; 28: 439-47.

2. Campbell W. Surgical morbidity and mortality meetings. Ann R Coll Surg Engl 1988;70:363-5.

3. McNamara DA, Hall HM, Hardin EA. Rethinking the modern cardiology morbidity and mortality conference: harmonizing quality improvement and education. J Am Coll Cardiol. 2019;73:868-72.

edit-april-fr.indd 77 2019-03-19 9:33 AM



Use of the Corail stem for revision total hip arthroplasty: evaluation of clinical outcomes and cost

Background: With the growing number of total hip arthroplasty (THA) procedures performed, revision surgery is also proportionately increasing, resulting in greater health care expenditures. The purpose of this study was to assess clinical outcomes and cost when using a collared, fully hydroxyapatite-coated primary femoral stem for revision THA compared to commonly used revision femoral stems.

Methods: We retrospectively identi�ed patients who underwent revision THA with a primary stem between 2011 and 2016 and matched them on demographic variables and reason for revision to a similar cohort who underwent revision THA. We extracted operative data and information on in-hospital resource use from the patients’ charts to calculate average cost per procedure. Patient-reported outcomes were recorded preoperatively and 1 year postoperatively.

Results: We included 20 patients in our analysis, of whom 10 received a primary stem and 10, a typical revision stem. There were no signi�cant between-group differ-ences in mean Western Ontario and McMaster Universities Osteoarthritis Index score, Harris Hip Score, 12-Item Short Form Health Survey (SF-12) Mental Com-posite Scale score or Physical Composite Scale score at 1 year. Operative time was signi�cantly shorter and total cost was signi�cantly lower (mean difference –3707.64, 95% con�dence interval –5532.85 to –1882.43) with a primary stem than with other revision femoral stems.

Conclusion: We found similar clinical outcomes and signi�cant institutional cost savings with a primary femoral stem in revision THA. This suggests a role for a primary femoral stem such as a collared, fully hydroxyapatite-coated stem for revision THA.

Contexte : Avec le nombre croissant d’interventions pour prothèse de hanche (PTH) effectuées, la chirurgie de révision est aussi proportionnellement en hausse, ce qui entraîne des coûts supérieurs pour le système de santé. Le but de cette étude était d’évaluer les résultats cliniques et le coût associés à l’emploi d’une prothèse fémorale primaire à collerette entièrement recouverte d’hydroxyapatite pour la révision de PTH, comparativement à d’autres prothèses d’usage courant utilisées pour les révisions.

Méthodes : Nous avons identi�é rétrospectivement les patients ayant subi une révi-sion de PTH avec une prothèse primaire entre 2011 et 2016 et nous les avons assortis selon les caractéristiques démographiques et le motif de la révision à une cohorte simi laire soumise à une révision de PTH. Nous avons extrait les données sur l’opération et sur l’utilisation des ressources hospitalières à partir des dossiers des patients pour calculer le coût par intervention. Les résultats déclarés par les patients ont été notés avant l’intervention et 1 an après.

Résultats : Nous avons inclus 20 patients dans notre analyse, dont 10 ont reçu une prothèse primaire et 10, une révision de prothèse typique. On n’a noté aucune dif-férence signi�cative entre les groupes pour ce qui est du score WOMAC (Western Ontario and McMaster Universities Osteoarthritis Index) moyen pour l’arthrose, du score de Harris pour la hanche, ou des sous-échelles santé mentale ou santé physique à 1 an du questionnaire SF-12 (12-Item Short Form Health Survey). L’intervention a duré signi�cativement moins longtemps et le coût a été signi�cativement moindre (différence moyenne –3707,64, intervalle de con�ance de 95 % –5532,85 à –1882,43) avec une prothèse primaire qu’avec les autres prothèses de révision.

Conclusion : Nous avons observé des résultats cliniques similaires et des économies signi�catives pour l’établissement avec la prothèse primaire utilisée pour la révision de PTH. Cela donne à penser que la prothèse fémorale primaire, par exemple, à col-lerette et entièrement recouverte d’hydroxyapatite, aurait un rôle à jouer pour la révi-sion de PTH.

Thomas J. Wood, MD Mohammad Alzahrani, MD, MSc Jacquelyn D. Marsh, PhD Lyndsay E. Somerville, PhD Edward M. Vasarhelyi, MD, MSc Brent A. Lanting, MD, MSc

Accepted May 11, 2018; Published online Feb. 1, 2019

Correspondence to: T. Wood 339 Windermere Rd London ON N6A 5A5 [email protected]

DOI: 10.1503/cjs.002318

stem-wood.indd 78 2019-03-19 12:26 PM

RESEARCH

Can J Surg, Vol. 62, No. 2, April 2019 79

A lthough total hip arthroplasty (THA) is considered a widely successful operation, the increasing num-ber of procedures has resulted in a growing num-

ber of revision operations.1,2 The most common causes for revision surgery include instability, mechanical loosening and infection.1 Revision THA is associated with longer hospital stays and substantially higher costs.1 In addition, such cases are challenging owing to structural bone loss, and various stem options, such as extensively porous, cemented, long proximally coated or modular stems are often required to obtain �xation.3,4

In revision THA, it is critical to obtain good �xation between the implant–bone interface, which is often challenging as, in many cases, the femoral canal is smooth and sclerotic.5 The use of cementless femoral stems in revision THA is gaining popularity owing to the poor initial outcomes following use of cemented components.6 However, subsidence with proximally coated stems and higher revision rates have made sur-geons turn to fully porous coated stems.4,6,7 The Corail femoral component (DePuy Synthes) is a cementless, tapered, fully hydroxyapatite-coated titanium stem with good clinical outcomes for both hip fracture man-agement and THA for arthritis.8–12 The tapered design has been reported to avoid medullary canal blocking and distributes the stress, and the hydroxyapatite pro-vides maximal osseointegration.11 However, there are concerns regarding loose hydroxyapatite particles, which may lead to premature wear and osteolysis.12 Although few studies have evaluated this stem in revi-sion THA, results have been positive, with good long-term function.5

In contrast, modular stems are often used in revision settings as they can independently address distal �xation and bypass bone loss as well as optimize leg length, offset and stability proximally.13 However, this added ease with modularity is also associated with increased costs and adds a potential site of failure compared to monolithic stems.14 The results with modular stems have been positive in terms of patient-reported outcomes as well as complica-tions and reoperation for mechanical failure.3,13,15,16

The purpose of this study was to assess clinical outcomes and cost with the primary collared fully hydroxyapatite-coated Corail femoral stem for revision THA compared to commonly used revision femoral stems. We hypothesized that patient-reported outcomes at 1 year are similar to those with revision femoral stems. The substantial cost sav-ings with the Corail stem could potentially justify the use of a primary femoral stem in a revision setting where proximal bone stock is deemed acceptable intraoperatively.

METHODS

We retrospectively identi�ed patients who underwent revision THA with a Corail stem between 2011 and

2016. We compared this cohort with a group of patients who underwent revision THA with a revision stem, matched for age, body mass index and reason for revision based on an inclusive database search. We extracted operative data and information on in-hospital resource use from the patients’ charts to calculate the average cost per procedure. Total procedure-related costs included implant cost and length of time in the operating room (from time in to time out). We obtained unit costs using administrative data from the case-costing department at our institution.

Patients prospectively completed several health-related quality-of-life instruments, including the West-ern Ontario and McMaster Universities Osteoarthritis Index (WOMAC), Harris Hip Score and 12-Item Short Form Health Survey (SF-12) Mental and Phys-ical Composite Scales, at each visit with their surgeon. These instruments were also administered preopera-tively and 1 year postoperatively. We recorded preop-erative proximal femoral bone loss for the 2 groups and classi�ed this using the Paprosky classi�cation.17 Insti-tutional ethics board approval was obtained for the study.

Statistical analysis

We used descriptive statistics to summarize the num-ber of reoperations, operating room time, implant costs and quality of life for the 2 groups. We per-formed independent sample t tests to compare patient-reported outcomes between groups, with p < 0.05 set as the level of signi�cance. We used nonparametric boot-strapping to compare differences in mean costs between groups. All costs are reported in 2017 Canad-ian dollars.

RESULTS

We included 20 patients in our analysis, of whom 10 received a primary collared Corail stem and 10 received a revision stem. Demographic characteristics, reason for revision and preoperative proximal femoral bone loss were similar between the 2 groups (Table 1). There were no sig-ni�cant differences in preoperative patient-reported out-comes (Table 1) (p > 0.05).

At 1 year postoperatively, there were no signi�cant differences in mean WOMAC score (p = 0.8), mean Harris Hip Score (p = 0.8), mean SF-12 Mental Com-posite Scale score (p = 0.9) or mean SF-12 Physical Composite Scale score between the groups (p = 0.3) (Table 2).

There were no complications with the Corail stem. One patient underwent reoperation owing to persistent infection. Five complications occurred in the control group, 1 case each of pulmonary embolism, deep vein

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80 J can chir, Vol. 62, No 2, avril 2019

Table 1. Baseline characteristics of patients who underwent revision total hip arthroplasty

Characteristic

Mean ± SD*

p valueCorail group

n =10Control group

n = 10

Age 70 ± 8.16 71 ± 9.05 0.8

Male:female ratio 70:30 50:50

Body mass index 31 ± 3.5 31 ± 6.05 > 0.99

Reason for revision Infection (3), aseptic loosening (3),

pseudotumour (2), other (2)

Infection (4), aseptic loosening (3), pseudotumour (2), other (1)

Type of stem Corail Restoration modular (Stryker) (6), AML monoblock (DePuy Synthes) (1),

Reclaim modular (DePuy Synthes) (1), S-ROM modular (DePuy Synthes) (1), LPS universal fluted modular (DePuy

Synthes) (1)

Paprosky classification of preoperative femoral bone deficiency

Type I: 7 Type II: 2

Type IIIA: 1

Type I: 6 Type II: 3

Type IIIA: 1

Harris Hip Score 47.4 ± 16.27 53.3 ± 13.71 0.4

12-Item Short Form Health Survey Physical Composite Scale score

26.6 ± 3.08 32.7 ± 8.48 0.1

12-Item Short Form Health Survey Mental Composite Scale score

51.3 ± 13.89 49.5 ± 13.05 0.8

Western Ontario and McMaster Universities Osteoarthritis Index score

45.8 ± 19.68 55.6 ± 22.70 0.4

SD = standard deviation. *Except where noted otherwise.

Table 2. Patient-reported outcomes at 1 year postoperatively

Outcome

Mean score ± SD

Mean difference (95% CI)Corail group Control group

Harris Hip Score 79.0 ± 13.29 80.8 ± 17.01 1.8 (–17.97 to 21.57)

12-Item Short Form Health Survey Physical Composite Scale

38.6 ± 10.66 43.1 ± 7.97 4.5 (–5.15 to 14.15)

12-Item Short Form Health Survey Mental Composite Scale

57.1 ± 5.92 56.8 ± 4.43 –0.3 (–5.66 to 5.06)

Western Ontario and McMaster Universities Osteoarthritis Index

79.2 ± 16.18 81.0 ± 12.76 1.8 (–13.17 to 16.77)

CI = confidence interval; SD = standard deviation.

Table 3. Operating room and implant costs

Variable

Mean ± SD

Mean difference (95% CI)Corail group Control group

Operating room time, min

171 ± 25.14 217 ± 45.09 46 (12.1 to 79.9)

Total cost, $* 5691.17 ± 1006.10 10 721.75 ± 3490.62 –3707.64 (–5532.85 to –1882.43)

CI = confidence interval; SD = standard deviation. *Time in operating room and implant.

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RESEARCH


thrombosis, sciatic nerve palsy, dislocation and greater tro-chanter fracture. There were no reoperations in the con-trol group.

The mean operative time was signi�cantly shorter in the Corail group than in the control group (Table 3). Mean total operating room costs were signi�cantly lower in the Corail group (mean difference –828.34, 95% con�-dence interval –1420.67 to –270.56), as were mean implant costs (mean difference –2879.301, 95% con�dence interval –4293.82 to –1464.79).

DISCUSSION

In this retrospective study evaluating the clinical outcomes and costs for patients undergoing revision THA with a pri-mary total hip femoral stem (Corail) compared to a matched control group of patients undergoing THA with a revision femoral stem, we found that patient-reported out-comes at 1 year postoperatively were not signi�cantly dif-ferent between the 2 groups. There were no complications in the Corail group, compared to 5 complications in the control group; complications in the latter group may be overrepresented owing to the small sample. Furthermore, operating room time and costs and implant costs also sig-ni�cantly favoured the Corail primary stem when used for revision THA. Taken together, the results suggest that there is a role for the use of a primary femoral stem such as the Corail in revision THA, especially in patients with minimal bone loss.

The Corail femoral stem is a tapered titanium stem with a proximal �ared design allowing for a 3-dimensional metaphyseal �t.11 Stability is achieved through initial press fit and secondarily via fixation through the bone–hydroxyapatite interface.11,12 The extensive hydroxyapatite coating increases biological �xation and allows for even stress distribution and good long-term survival.5,12 In a retrieval study of 165 patients, Coathup and colleagues18 found significantly more ingrowth and attachment to bone with the hydroxyapatite porous coating than with plasma spray or grit-blasted stems. Furthermore, the seal between the bone and the implant is purportedly strong with the Corail primary stem, which slows down wear particle migration, limiting osteolysis.12,18 This stem has been shown to have good results when used in the primary setting, with survivorship of 96.3% at 23 years.12 Reikerås5 reported that, of 66 consecutive patients who underwent revision THA in which the Corail primary stem was used, only 1 required revision for mechanical failure. Although 8 hips that had a proximal fracture intraoperatively requir-ing wire fixation, long-term results were good up to 27 years.

In comparison, the matched cohort consisted of typ-ical revision modular THA stems including the Restora-tion modular (Stryker), S-ROM (DePuy Synthes), Reclaim (DePuy Synthes) and LPS universal fluted

modular (DePuy Synthes) in addition to an AML mono-block stem (DePuy Synthes). These stems enable the surgeon to engage the diaphysis to obtain axial and rota-tional stability distally.3 A distinct advantage of using a modular stem is obtaining a secure �t distally indepen-dent of the proximal body, which allows for optimization of limb length, stability and offset.3 This is especially important when there is a mismatch between bone loss proximally and distally.15 This was shown by Restrepo and colleagues,13 who used a Restoration modular stem in 118 patients; limb length was restored to within 5 mm in 75% of patients and offset to within 2 mm in 65% of patients. Dzaja and colleagues3 reported improvement in the WOMAC score and Harris Hip Score in 55 patients who underwent revision THA in which the Restoration modular stem was used, with 2 patients requiring revi-sion owing to infection and subsidence. Similarly, in 161 revision THA procedures in which the Restoration modular stem was used, Riesgo and colleagues16 found an overall reoperation rate of 14.9% and an aseptic loosening rate of 2% with mean follow-up of 6.1 years. However, there are risks of junction failure and cor-rosion in addition to cost concerns when using modular femoral stems, especially in patients with minimal bone loss.14

Limitations

The limitations of this study include the small sample and the short follow-up period. Large prospective studies are warranted to support our results. However, the goal of this study was to compare early failures with a primary total hip femoral stem used in a revision setting and the cost differences as compared to revision femoral stems. Although it is recognized that a primary femoral compo-nent should be used only in speci�c circumstances, we mitigated this limitation by matching to revision cases with similar indications for revision surgery. In patients in whom the calcar is intact and there is minimal bone loss, with overall robust proximal femoral bone quality, a pri-mary femoral stem can be appropriately used for revision THA.

CONCLUSION

Our results suggest a role for a primary femoral stem such as the Corail for revision THA. We found similar clinical outcomes and substantial institutional cost savings com-pared to a matched control group who received revision femoral stems. Therefore, taking into account patient factors including bone deficiency, a collared, fully hydroxyapatite-coated stem such as the Corail primary femoral stem can play a role in revision THA, with com-parable early clinical results to those with other revision femoral stems.

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Af�liations: From the Division of Orthopaedic Surgery, Department of Surgery, London Health Sciences Centre, University Campus, Western University, London, Ont. (Wood, Alzahrani, Vasarhelyi, Lanting); the Department of Orthopaedic Surgery, Imam Abdulrahman Bin Faisal University, Dammam, Saudi Arabia (Alzahrani); the School of Physical Therapy, Faculty of Health Sciences, Western University, London, Ont. (Marsh); and the Bone and Joint Institute, Western University, London, Ont. (Marsh, Vasarhelyi, Lanting).

Competing interests: E. Vasarhelyi and B. Lanting have received con-sulting fees and research support from Smith & Nephew, DePuy Synthes and Stryker. No other competing interests declared.

Contributors: T. Wood, L. Somerville, E. Vasarhelyi and B. Lanting designed the study. T. Wood, J. Marsh, L. Somerville and E. Vasarhelyi acquired the data, which all authors analyzed. T. Wood, M. Alzahrani and L. Somerville wrote the article, which all authors reviewed and approved for publication.

References

1. Bozic KJ, Kurtz SM, Lau E, et al. The epidemiology of revision total hip arthroplasty in the United States. J Bone Joint Surg Am 2009;91: 128-33.

2. Kurtz S, Ong K, Lau E, et al. Projections of primary and revision hip and knee arthroplasty in the United States from 2005 to 2030. J Bone Joint Surg Am 2007;89:780-5.

3. Dzaja I, Lyons MC, McCalden RW, et al. Revision hip arthroplasty using a modular revision hip system in cases of severe bone loss. J Arthroplasty 2014;29:1594-7.

4. Malkani AL, Lewallen DG, Cabanela ME, et al. Femoral component revision using an uncemented, proximally coated, long-stem prosthe-sis. J Arthroplasty 1996;11:411-8.

5. Reikerås O. Femoral revision surgery using a fully hydroxyapatite-coated stem: a cohort study of twenty two to twenty seven years. Int Orthop 2017;41:271-5.

6. Barrack RL, Folgueras AJ. Revision total hip arthroplasty: the femo-ral component. J Am Acad Orthop Surg 1995;3:79-85.

7. Mulliken BD, Rorabeck CH, Bourne RB. Uncemented revision total hip arthroplasty: a 4-to-6-year review. Clin Orthop Relat Res 1996; 325:156-62.

8. Cawley DT, Curtin PD, Lohan D, et al. The Corail® stem for the treatment of displaced femoral neck fractures — a viable alternative. Hip Int 2011;21:243-50.

9. Drobniewski M, Borowski A, Synder M, et al. Results of total cementless hip joint arthroplasty with Corail stem. Ortop Traumatol Rehabil 2013;15:61-8.

10. Hallan G, Lie SA, Furnes O, et al. Medium- and long-term perfor-mance of 11,516 uncemented primary femoral stems from the Nor-wegian Arthroplasty Register. J Bone Joint Surg Br 2007;89:1574-80.

11. Xu J, Xie Z, Zhao J, et al. Results of a hydroxyapatite-coated femoral stem (Corail) in Chinese: a minimum 10-year follow up. Springerplus 2016;5:1983.

12. Vidalain JP. Twenty-year results of the cementless Corail stem. Int Orthop 2011;35:189-94.

13. Restrepo C, Mashadi M, Parvizi J, et al. Modular femoral stems for revision total hip arthroplasty. Clin Orthop Relat Res 2011;469:476-82.

14. Sivananthan S, Lim CT, Narkbunnam R, et al. Revision hip arthro-plasty using a modular, cementless femoral stem: intermediate-term follow-up. J Arthroplasty 2017;32:1245-9.

15. Stimac JD, Boles J, Parkes N, et al. Revision total hip arthroplasty with modular femoral stems. J Arthroplasty 2014;29:2167-70.

16. Riesgo AM, Hochfelder JP, Adler EM, et al. Survivorship and com-plications of revision total hip arthroplasty with a mid-modular fem-oral stem. J Arthroplasty 2015;30:2260-3.

17. Sheth NP, Nelson CL, Paprosky WG. Femoral bone loss in revision total hip arthroplasty: evaluation and management. J Am Acad Orthop Surg 2013;21:601-12.

18. Coathup MJ, Blunn GW, Flynn N, et al. A comparison of bone remodeling around hydroxyapatite-coated, porous-coated and grit-blasted hip replacements retrieved at post-mortem. J Bone Joint Surg Br 2001;83:118-23.

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stem-wood.indd 82 2019-03-19 12:26 PM



Relation between socioeconomic variables and surgical, systemic and radiation treatment in a cohort of patients with breast cancer in an urban Canadian centre

Background: Studies have shown an association between socioeconomic status and breast cancer treatment. We examined the relation between socioeconomic status and the treatment of breast cancer (surgical, systemic and radiation) in a universal health care system.

Method: Data from a single urban Canadian centre were collected for consecutive patients who received a diagnosis of breast cancer from January 2010 to December 2011. Variables included patient and disease factors, surgery type, systemic and radia-tion treatment, and breast reconstruction. Socioeconomic variables were obtained from 2006 Canadian census data. We used multivariable logistic regression to identify predictors of breast cancer treatment.

Results: A total of 721 patients were treated for breast cancer during the study period. Socioeconomic variables were not related to type of breast surgery for breast cancer. Age less than 50 years, having a �rst-degree relative with breast cancer and income status were predictors of breast reconstruction. Employment status was a con-sistent predictor of systemic and radiation treatment.

Conclusion: Employment consistently predicted systemic and radiation treatment, and age and income were predictors of breast reconstruction in a universal health care system. Further research is required to determine precisely how socioeconomic factors affect care and to minimize possible disparities in delivery of health care services.

Contexte : Des études ont montré un lien entre la situation socio-économique et le traitement du cancer du sein. Nous avons analysé ce lien entre la situation socio-économique et le traitement (chirurgie, chimiothérapie, radiothérapie) du cancer du sein dans un système de santé universel.

Méthodes : Les données d’un seul centre urbain canadien ont été compilées pour les patientes consécutives ayant reçu un diagnostic de cancer du sein entre janvier 2010 et décembre 2011. Les variables incluaient des facteurs propres aux patientes et à la mala die, le type de chirurgie, la chimiothérapie, la radiothérapie et la reconstruction mammaire. Les variables socio-économiques proviennent des données du recense-ment canadien de 2006. Nous avons utilisé la régression logistique multivariée pour identi�er les prédicteurs du traitement du cancer du sein.

Résultats : En tout, 721 patientes ont été traitées pour un cancer du sein durant la période de l’étude. Les variables socio-économiques n’ont pas in�ué sur le type de chirurgie mammaire pour cancer du sein. L’âge inférieur à 50 ans, un cancer du sein chez une parente au premier degré et le revenu ont été des prédicteurs de la recon-struction mammaire. La situation professionnelle a été un prédicteur �able du traite-ment systémique et de la radiothérapie.

Conclusion : L’emploi a été un prédicteur �able du traitement systémique et de la radiothérapie, et l’âge et le revenu ont été des prédicteurs de la reconstruction mam-maire, dans un système de santé universel. Il faudra approfondir la recherche pour déterminer plus précisément l’in�uence des facteurs socio-économiques sur les soins et pour réduire les possibles disparités dans leur prestation.

Jennifer Li, MD Sylvie D. Cornacchi, MSc Forough Farrokhyar, PhD Neil Johnston, MSc Shawn Forbes, MD, MSc Susan Reid, MD Nicole Hodgson, MD, MSc Sarah Lovrics, BA Kristen Lucibello, MSc Peter Lovrics, MD

Results from this project were presented as a poster presentation at the Canadian Surgery Forum conference, Toronto, Ont., Sept. 10, 2016.

Accepted Apr. 17, 2018; Published online Feb. 1, 2019

Correspondence to: P. Lovrics G802, St. Joseph’s Healthcare 50 Charlton Ave E Hamilton ON L8N 4A6 [email protected]

DOI: 10.1503/cjs.009217

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B reast cancer is the most commonly diagnosed cancer and the second-leading cause of cancer mortality in North American women.1 Breast cancer is treated

with a multidisciplinary approach involving surgery, radia-tion, chemotherapy and antihormonal treatments. It is a highly curable type of cancer if diagnosed early.

There is evidence that nonbiological factors may affect the presentation, diagnosis, treatment and outcome of breast cancer in women. Many studies have shown a posi-tive relation between higher socioeconomic status and improved breast cancer survival in the United States,2–5 the United Kingdom,6 the Netherlands,7 Australia8 and Can-ada.9–12 In addition, compared to patients of higher socio-economic status, those of lower socioeconomic status were found to be less often in screening programs,13–15 to receive adjuvant chemoradiation less often16–18 and to receive the diagnosis of breast cancer at a later stage.19 Conversely, Gorey and colleagues20,21 compared 2 areas in Canada and 2 areas in the US and found a positive association between socioeconomic status and breast cancer survival in Hawaii and Iowa but not in Manitoba and Ontario. They impli-cated universal health coverage in Canada in their discus-sion to explain this result. Regionally, socioeconomic status was found to have an impact on overall health and longev-ity in a study in Hamilton, Ontario.22 The incidence of breast cancer in the inner city and urban core was lower than in the more af�uent suburban areas, where socioeco-nomic status was higher, but rates of breast cancer mortal-ity were higher in the inner city.23

Breast reconstruction is available for patients undergo-ing mastectomy and is offered in an immediate or delayed fashion. Several studies showed that reconstruction was more likely to be chosen by women who were younger and white and who resided in an urban setting where there was better access to plastic surgeons.24–26 Higher socioeco-nomic status has also been reported to be associated with a higher rate of breast reconstruction following mastec-tomy.27 However, a report from Nova Scotia suggested that this might not be the case in Canada.28

The current study aimed to assess the relation between socioeconomic status, surgical and nonsurgical treatment of breast cancer, and the rate of breast reconstruction in an urban Canadian centre where care is publicly funded. The city of Hamilton is an amalgamated city with an older postindustrial core and more af�uent suburbs. The potential value of such research is to identify any discrep-ancy in care that is associated with factors related to socio-economic status. Universal health care is designed to be equal and accessible to everyone; however, there remains underlying inequality in access to health care services that is not fully understood. The findings from this study should aid in increasing awareness of determinants of health and consumption of health care services, and potentially foster changes in public policy, resource allo-cation and education.

METHODS

Study cohort

Consecutive patients with a diagnosis of breast cancer (sur-gical and nonsurgical cases, including invasive cancer and ductal carcinoma in situ) living in Hamilton from January 2010 to December 2011 were included in this retrospective chart review. Cases were identi�ed through health record queries of hospital databases and the Regional Cancer Centre database in Hamilton. Male patients, patients with a diagnosis of benign breast disease and patients who resided outside of Hamilton or had their primary treat-ment for breast cancer outside Hamilton were excluded. Local research ethics board approval was obtained for the study.

Data collection

Trained data abstractors reviewed hospital and cancer cen-tre charts and extracted the following information: patient demographic characteristics (age at diagnosis, marital sta-tus, obesity [body mass index > 30]), employment status, smoking status, comorbidities, presence of �rst-degree rel-ative with breast cancer, previous breast cancer, method of tumour identi�cation (asymptomatic screening v. symp-tomatic investigation), location of primary imaging (Ontario Breast Screening Program, hospital or nonhospi-tal clinic), disease stage at diagnosis, type of breast and nodal surgery (breast-conserving surgery or mastectomy, sentinel lymph node biopsy, axillary lymph node dissection or status on breast reconstruction), pathological features (tumour size, nodal status, receptor status), receipt of neo-adjuvant or adjuvant therapy (chemotherapy including trastuzumab, radiation to breast, chest wall and nodes, and hormonal therapy) and, for nonsurgical cases, type of palli-ative treatment. Employment status was coded as employed versus not employed (i.e., unemployed, receiving disability support, homemaker or retired).

Census data

We linked patients’ postal codes to 2006 Canadian census data to extract the following data: education level (propor-tion of census tract population who completed college or university), immigration status (proportion of census tract population who immigrated in the previous 5 yr) and aver-age income of the census tract of patient residence. We divided income into quintiles.


For data analysis, we performed descriptive statistics. Cate-gorical variables were reported as counts and percentages and were compared with the use of the χ2 test or the Fisher


RESEARCH


exact test. Continuous variables were reported as mean and standard deviation and were compared with the use of a t test for independent samples. We performed multivari-able logistic regression analysis to identify patient, patho-logical and socioeconomic status variables that had an impact on surgical and nonsurgical treatment of breast cancer, and rate of breast reconstruction. Odds ratios (ORs) with 95% con�dence intervals (CIs) and Hosmer–Lemeshow goodness-of-fit values are reported. We checked multicolinearity using correlation analysis. To maximize power, we performed univariable analysis and entered variables with a value less than 0.1 into the multi-variable regression. Statistical signi�cance was set at p < 0.05. Data analyses were performed with SPSS Statistical Software version 23.0 (IBM Corp.).

RESULTS

A total of 721 cases met the inclusion criteria and were reviewed. Table 1 outlines the demographic and clinical characteristics of the study cohort. Breast-conserving sur-gery was performed in 393 patients (54.5%), nodal surgery in 545 (75.6%) and reconstruction after mastectomy in 67/276 (24.3%). Most patients presented with early-stage disease, but 176 (24.4%) presented with stage 3 or 4 dis-ease. Almost half (337 [46.7%]) presented with screen-detected cancers. Income quintile, completion of college or university, and immigration status showed relatively similar distributions across the various categories.

Socioeconomic status versus surgical treatment of breast cancer

Table 2 outlines patient, disease and socioeconomic status factors for type of breast surgery received. Univariate analysis showed that mastectomy rates were signi�cantly higher among women less than 51 years of age, those with stage 3 or 4 disease and those with a previous history of breast cancer. Mastectomy rates did not vary with other factors, such as higher education or income quintile. Of the 276 patients who underwent mastectomy, 67 (24.3%) had breast reconstruction (immediately in 40 cases [60%] and delayed in 27 [40%]). Breast reconstruction was done significantly more frequently in women who were younger, employed, healthier (fewer comorbidities) and more educated, had earlier-stage disease and lived in higher-income census tracts. Multivariable analysis showed that only early disease stage (stage 0–1: OR 22.75, 95% CI 12.8–40.5; stage 2: OR 9.44, 95% CI 5.3–16.6) and a previous history of breast cancer (OR 5.25, 95% CI 2.9–9.4) were independent predictors of breast-conserving surgery versus mastectomy (Table 3). Socioeconomic sta-tus factors were not signi�cant predictors. After baseline differences were adjusted for, multivariable logistic regres-sion analysis showed that patients younger than 50 years

were 30 times more likely to have reconstruction, and those with a �rst-degree relative with breast cancer were twice as likely to have reconstruction. Furthermore, with each increase in income quintile, the odds of having breast reconstruction doubled for women aged 51–70 years (Table 3).

Table 4 outlines patient, disease and socioeconomic sta-tus factors compared with the type of nodal surgery. Of the 721 patients, 545 (75.6%) had nodal surgery, of whom 317 (58.2%) had sentinel lymph node biopsy and 228 (41.8%) had axillary lymph node dissection. Univariate analysis showed that age, having a �rst-degree relative with breast cancer and disease stage predicted sentinel lymph node biopsy; socioeconomic status factors (e.g., income quintile, higher education) were not signi�cant. On multi-variable analysis, predictors of not having nodal surgery were age more than 70 (OR 1.54, 95% CI 0.9–2.5), not employed (OR 1.85, 95% CI 1.1–3.1), nonsmoker (OR 1.97, 95% CI 1.3–3.1) and previous breast cancer (OR 3.66, 95% CI 2.2–6.2) (Table 3). Independent predic-tors of sentinel lymph node biopsy versus axillary lymph node dissection included no previous breast cancer (OR 3.00, 95% CI 1.3–7.0) and early stage disease (stage 0–1: OR 166.26, 95% CI 64.8–426.6; stage 2: OR 26.58, 95% CI 11.1–63.9).

Socioeconomic status versus systemic and radiation treatment for breast cancer

Table 5 outlines the demographic and socioeconomic sta-tus factors compared to systemic and radiation therapies. A total of 256 patients (35.5%) received chemotherapy, either in a neoadjuvant or an adjuvant setting, and 342 patients (87.0%) who had breast-conserving surgery received adjuvant radiation to the breast. Hormonal ther-apy was taken by 390 patients (72.8%) with positive recep-tor status. On multivariable analyses, predictors for che-motherapy included age less than 51 (OR 5.25, 95% CI 2.9–9.4), later-stage disease (stage 2: OR 7.50, 95% CI 4.40–13.1; stage 3–4: OR 36.64, 95% CI 19.5–68.6), being employed (OR 4.84, 95% CI 2.9–7.9) and being married (OR 1.90, 95% CI 1.2–3.0) (Table 3). Predictors of radia-tion included being employed (OR 3.06, 95% CI 1.1–8.3) and having no comorbidities (OR 9.34, 95% CI 1.2–71.2). Finally, predictors for hormonal therapy included being employed (OR 3.55, 95% CI 2.1–60.5) and later-stage dis-ease (stage 2: OR 7.45, 95% CI 2.1–5.8; stage 3–4: OR 14.00, 95% CI 5.8–33.8). No relation was identi�ed between socioeconomic status factors and these adjuvant therapies.

DISCUSSION

There are numerous treatment options in the management of breast cancer. Recommendations are based on many


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Table 1 (part 2 of 2). Demographic characteristics, disease factors and type of treatment among women with breast cancer

VariableNo. (%) of patients

n = 721% of census tract population who immigrated in previous 5 yr > 30 199 (27.6) 21–30 224 (31.1) < 21 295 (40.9) NOS 3 (0.4)Type of breast surgery Breast-conserving 393 (54.5) Mastectomy 276 (38.3) No surgery 52 (7.2)Type of nodal surgery Sentinel lymph node biopsy 317 (44.0) Axillary lymph node dissection* 228 (31.6) No nodal surgery 176 (24.4)Breast reconstruction after mastectomy (n = 276)

67 (24.3)

Tumour stage Tis 74 (10.3) T1 291 (40.4) T2 223 (30.9) T3–T4 133 (18.4)Nodal stage N0 365 (50.6) N1–N3 228 (31.6) NX 127 (17.6) NOS 1 (0.1) Disease stage (TNM) 0–1 323 (44.8) 2 222 (30.8) 3–4 176 (24.4)Estrogen receptor/progesterone receptor status† (n = 637) Positive 536 (84.1) Negative 101 (15.8)Tumour grade I 145 (20.1) II 362 (50.2) III 175 (24.3) NOS 39 (5.4)Histologic type† (n = 647) Ductal 529 (81.8) Lobular‡ 78 (12.0) Other (e.g., mucinous, tubular) 32 (4.9) NOS 8 (1.2)Chemotherapy§ Yes 256 (35.5) No 461 (63.9) NOS 4 (0.6)Hormonal therapy§ Yes 395 (54.8) No 322 (44.7) NOS 4 (0.6)Radiation therapy 460 (63.8)Recurrent breast cancer 50 (6.9)No treatment 5 (0.7)Neoadjuvant treatment 79 (11.0)

NOS = not otherwise specified. *Includes cases with sentinel lymph node biopsy and axillary lymph node dissection, and axillary lymph node dissection alone †For invasive cases only. ‡Lobular or mixed ductal/lobular. §Includes neoadjuvant and adjuvant.

Table 1 (part 1 of 2). Demographic characteristics, disease factors and type of treatment among women with breast cancer

VariableNo. (%) of patients

n = 721

Age at diagnosis, yr

≤ 50 149 (20.7)

51–70 364 (50.5)

> 70 208 (28.8)

Obese (body mass index > 30)

Yes 255 (35.4)

No 435 (60.3)

NOS 31 (4.3)

First-degree relative with breast cancer

Yes 159 (22.0)

No 519 (72.0)

NOS 43 (6.0)

Smoking status

Smoker/previous smoker 289 (40.1)

Never smoked 426 (59.1)

NOS 6 (0.8)

No. of comorbidities

0 183 (25.4)

1–2 296 (41.0)

≥ 3 242 (33.6)

Previous breast cancer 87 (12.1)

Location of primary imaging

Ontario Breast Screening Program 258 (35.8)

Hospital 314 (43.6)

Nonhospital clinic 140 (19.4)

NOS 9 (1.2)

Method of tumour identification

Asymptomatic screening 337 (46.7)

Symptomatic investigation 383 (53.1)

NOS 1 (0.1)

Imaging in previous 2 yr

Yes 317 (44.0)

No 313 (43.4)

NOS 91 (12.6)

Marital status

Married 447 (62.0)

Not married 272 (37.7)

NOS 2 (0.3)

Employment status

Employed 277 (38.4)

Not employed 430 (59.6)

NOS 14 (1.9)

Income quintile

1 (lowest) 147 (20.4)

2 141 (19.6)

3 148 (20.5)

4 140 (19.4)

5 (highest) 139 (19.3)

NOS 6 (0.8)

% of census tract population who completed college or university

> 40 304 (42.2)

31–40 237 (32.9)

< 31 177 (24.5)

NOS 3 (0.4)


RESEARCH


Table 2. Univariate analysis of demographic, clinical and socioeconomic factors versus surgical treatment of breast cancer and reconstruction following mastectomy

Variable

No. (%) of patients*

p value


p value

Breast-conserving surgery n = 393

Mastectomy n = 276

Breast reconstruction

n = 67

No breast reconstruction

n = 209


< 51 68 (17.3) 76 (27.5) 0.004 37 (55.2) 39 (18.7) < 0.001

51–70 219 (55.7) 126 (45.6) 29 (43.3) 97 (46.4)

> 70 106 (27.0) 74 (26.8) 1 (1.5) 73 (34.9)

Obese

Yes 146 (38.6) 95 (35.7) 0.4 21 (31.8) 74 (37.0) 0.4

No 232 (61.4) 171 (64.3) 45 (68.2) 126 (63.0)

Employment

Employed 152 (39.6) 118 (43.1) 0.4 49 (75.4) 69 (33.0) < 0.001

Not employed 232 (60.4) 156 (56.9) 16 (24.6) 140 (67.0)

Marital status

Married 256 (65.1) 172 (62.8) 0.5 44 (66.7) 128 (61.8) 0.6

Not married 137 (34.9) 102 (37.2) 22 (33.3) 79 (38.2)


Yes 92 (24.5) 54 (21.0) 0.3 18 (40.9) 36 (18.6) 0.1

No 283 (75.5) 203 (79.0) 26 (59.1) 157 (81.3)

Smoking status

Smoker/previous smoker 152 (38.8) 117 (42.8) 0.3 31 (47.0) 86 (41.5) 0.4

Never smoked 240 (61.2) 156 (57.1) 35 (53.0) 121 (58.4)

No. of comorbidities†

0 105 (26.7) 71 (25.7) 0.8 30 (44.8) 41 (19.6) < 0.001

≥ 1 288 (73.3) 205 (74.3) 37 (55.2) 168 (80.4)

Income quintile

1 78 (20.1) 60 (21.8) 0.2 9 (13.4) 51 (24.5) 0.01

2 66 (17.0) 63 (22.9) 9 (13.4) 54 (26.0)

3 84 (21.6) 55 (20.0) 15 (22.4) 40 (19.2)

4 77 (19.8) 55 (20.0) 19 (28.4) 36 (17.3)

5 83 (21.4) 42 (15.3) 15 (22.4) 27 (13.0)


> 40 159 (40.7) 121 (44.0) 0.07 39 (58.2) 82 (39.4) 0.02

31–40 143 (36.6) 78 (28.4) 16 (23.9) 62 (29.8)

< 31 89 (22.8) 76 (27.6) 12 (17.9) 64 (30.8)

% of census tract population who immigrated in previous 5 yr

> 0.99

> 30 107 (27.4) 75 (27.3) 14 (20.9) 61 (29.3) 0.1

21–30 124 (31.7) 88 (32.0) 19 (28.4) 69 (33.2)

< 21 160 (40.9) 112 (40.7) 34 (50.7) 78 (37.5)

Previous breast cancer

Yes 29 (7.4) 51 (18.5) < 0.001 9 (13.4) 42 (20.1) 0.2

No 364 (92.6) 225 (81.5) 58 (86.6) 167 (79.9)

Disease stage (TNM)

0–1 241 (61.3) 80 (29.0) < 0.001 30 (44.8) 50 (23.9) 0.004

2 130 (33.1) 83 (30.1) 17 (25.4) 66 (31.6)

3–4 22 (5.6) 113 (40.9) 20 (29.8) 93 (44.5)

*Numbers do not sum to total in all cases because of missing data. †Includes chronic obstructive pulmonary disease, rheumatic disease, coronary artery disease, myocardial infarction, coronary heart failure, diabetes, kidney disease, major psychiatric illness, morbid obesity (body mass index > 40), other cancer diagnosis, osteoporosis, hypertension, hypercholesterolemia and neurodegenerative diseases (e.g., Parkinson’s disease).


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potential factors: patient factors, tumour factors, local experience and guidelines. Many factors also can in�uence decisions that patients make. In our urban Canadian cen-tre, we examined the relation between socioeconomic sta-tus and breast cancer treatment and found that socioeco-nomic status was related only to reconstruction rates after mastectomy and not to other surgical interventions. In addition, patients more than 70 years of age were less likely than younger patients to undergo axillary surgery, includ-

ing sentinel node biopsy and axillary dissection. This might be explained by the �ndings from a randomized controlled trial that axillary surgery can be avoided in older women receiving hormonal therapy, with no adverse onco-logic outcomes.29

A US study showed that patients in lower socioeco-nomic classes were less likely than those in higher socio-economic classes to undergo breast-conserving surgery owing to larger tumour size at the time of diagnosis.30 Such results were not observed in our study. This may be partially attributable to Canada’s universal health cover-age but also to an ongoing effort to make screening more accessible, especially for women in lower socioeconomic status neighbourhoods.25,26 In the Hamilton region, a “Screen for Life” bus was initiated in 2012, and more than 600 patients were screened in its �rst year of opera-tion. The program targeted neighbourhoods with low socioeconomic status and poor access to screening cen-tres, and cultural groups less inclined to accept cancer screening.31,32 Such programs have also been used in northern Ontario, where access to health care is limited, and in urban neighbourhoods with lower socioeconomic status.

International studies have shown that patients with lower socioeconomic status are less likely than those with higher socioeconomic status to receive adjuvant chemora-diation.11,12,33 Most of the studies were conducted in 2-tier health care systems; however, a recent study from Ontario showed that socioeconomic status affected breast cancer treatment and outcomes in a universal health care sys-tem.12 We also found that patients who were not employed (including homemakers and those who were unemployed, receiving disability support or retired) were less likely than employed patients to receive systemic and radiation treatment in a publicly funded health care sys-tem. The explanation for this is unclear, as systemic and radiation therapies are generally covered by public health insurance; there may be limitations in coverage of certain drugs, and there may be other, ancillary costs (such as transportation and costs related to side effects of treat-ment) associated with these therapies that discourages treatment. It also unknown whether these therapies are recommended but not accepted by patients, or whether different recommendations are made by oncologists based on employment status, as these details were not included in our data collection. A more detailed study focused on this issue would elucidate whether patients with low socioeconomic status are receiving fewer systemic and radiation therapies owing to access versus patient choice versus recommendation by the medical or radiation oncologist.

Interestingly, married patients were 1.9 times more likely to receive chemotherapy than unmarried patients, presumably owing to increased social support and social networks. A retrospective cohort study using the National

Table 3. Multivariable analyses exploring predictors of breast cancer treatment

Variable OR (95% CI)

Predictors of breast-conserving surgery v. mastectomy (n = 591)

No previous breast cancer 5.25 (2.9–9.4)

Stage 0–1 disease 22.75 (12.8–40.5)

Stage 2 disease 9.44 (5.3–16.6)

Stage 3–4 disease Reference

Hosmer–Lemeshow p value = 0.7

Predictors of breast reconstruction (n = 276)

Age < 50 yr at diagnosis 30.3 (9.2–99.2)

Age 51–70 yr by income quintile 2.0 (1.5–2.8)

First-degree relative with breast cancer 2.7 (1.1–6.5)


Predictors of no nodal surgery v. nodal surgery (n = 627)

Age > 70 yr 1.54 (0.9–2.5)

Not employed 1.85 (1.1–3.1)

Nonsmoker 1.97 (1.3–3.1)

Previous breast cancer 3.66 (2.2–6.2)


Predictors of SLNB v. ALND (n = 493)

No previous breast cancer 3.00 (1.3–7.0)

Stage 0–1 disease 166.26 (64.8–426.6)

Stage 2 disease 26.58 (11.1–63.9)



Predictors of chemotherapy (n = 623)

Age < 51 yr 5.25 (2.9–9.4)


Stage 2 disease 7.50 (4.40–13.1)

Stage 3–4 disease 36.64 (19.5–68.6)

Employed 4.84 (2.9–7.9)

Married 1.90 (1.2–3.0)


Predictors of radiation therapy for breast-conserving surgery cases (n = 347)

Employed 3.06 (1.1–8.3)

No comorbidities 9.34 (1.2–71.2)


Predictors of hormonal therapy for ER/PR-positive cases (n = 468)

Employed 3.55 (2.1–60.5)


Stage 2 disease 7.45–(2.1–5.8)

Stage 3–4 disease 14.00 (5.8–33.8)


ALND = axillary lymph node dissection; CI = confidence interval; ER/PR = estrogen/progesterone receptor; OR = odds ratio; SLNB = sentinel lymph node biopsy.


RESEARCH


Cancer Institute Surveillance, Epidemiology, and End Results (SEER) database34 showed that, compared to married women, unmarried women were more likely to receive their breast cancer diagnosis at a later stage, were less likely to receive therapy and had higher mortality,

which reaffirms the importance of social support for patients to complete diagnostic and therapeutic care.

Socioeconomic status has been shown to in�uence the frequency of breast reconstruction in the US, Australia, England and Denmark.18–21,35 In 2008, the rate of

Table 4. Univariate analysis of patient demographic and socioeconomic factors versus nodal surgery

Variable


p value


p valueNodal surgery

n = 545No nodal surgery

n = 176

Sentinel lymph node biopsy

n = 317

Axillary lymph node dissection†

n = 228


< 51 129 (23.7) 20 (11.4) < 0.001 62 (19.6) 67 (29.4) 0.008

51–70 283 (51.9) 81 (46.0) 181 (57.1) 102 (44.7)

> 70 133 (24.4) 75 (42.6) 74 (23.3) 59 (25.9)

Obese

Yes 206 (39.0) 49 (30.2) 0.04 117 (38.2) 89 (40.1) 0.7

No 322 (61.0) 113 (69.8) 189 (61.8) 133 (59.9)

Employment

Employed 233 (43.5) 44 (25.7) < 0.001 133 (42.8) 100 (44.4) 0.7

Not employed 303 (56.5) 127 (74.3) 178 (57.2) 125 (55.6)

Marital status

Married 354 (65.2) 93 (52.8) 0.003 210 (66.7) 144 (63.2) 0.4

Not married 189 (34.8) 83 (47.2) 105 (33.3) 84 (36.8)


Yes 120 (22.9) 39 (25.3) 0.5 80 (26.2) 40 (18.3) 0.03

No 404 (77.1) 115 (74.7) 225 (73.8) 179 (81.7)

Smoking status

Smoker/previous smoker 240 (44.3) 49 (28.3) < 0.001 131 (41.6) 109 (48.0) 0.1

Never smoked 302 (55.7) 124 (71.7) 184 (58.4) 118 (52.0)

No. of comorbidities

0 149 (27.3) 34 (19.3) 0.03 83 (26.2) 66 (28.9) 0.48

≥ 1 396 (72.7) 142 (80.7) 234 (73.8) 162 (71.0)

Income quintile

1 112 (20.8) 35 (19.9) 0.9 54 (17.2) 58 (25.7) 0.2

2 110 (20.4) 31 (17.6) 65 (20.8) 45 (19.9)

3 108 (20.0) 40 (22.7) 66 (21.1) 42 (18.6)

4 106 (19.7) 34 (19.3) 65 (20.8) 41 (18.1)

5 103 (19.1) 36 (20.4) 63 (20.1) 40 (17.7)


> 40 230 (42.4) 74 (42.0) > 0.99 135 (42.8) 95 (41.8) 0.2

31–40 179 (33.0) 58 (33.0) 111 (35.2) 68 (30.0)

< 31 133 (24.5) 44 (25.0) 69 (21.9) 64 (28.2)

% of census tract population who immigrated in previous 5 yr

> 30 151 (27.8) 48 (27.3) 0.3 84 (26.7) 67 (29.5) 0.8

21–30 161 (29.7) 63 (35.8) 96 (30.5) 65 (28.6)

< 21 230 (42.4) 65 (36.9) 135 (42.8) 95 (41.8)

Previous breast cancer

Yes 47 (8.6) 40 (22.7) < 0.001 27 (8.5) 20 (8.8) 0.9

No 498 (91.4) 136 (77.3) 290 (91.5) 208 (91.2)

Disease stage (TNM)

0–1 233 (42.8) 90 (51.1) 0.005 205 (64.7) 28 (12.3) < 0.001

2 185 (33.9) 37 (21.0) 105 (33.1) 80 (35.1)

3–4 127 (23.3) 49 (27.8) 7 (2.2) 120 (52.6)

*Numbers do not sum to total in all cases because of missing data. †Includes cases with sentinel lymph node biopsy and axillary lymph node dissection, or axillary lymph node dissection alone.


RECHERCHE


Table 5. Univariate analysis of patient demographic and socioeconomic factors versus neoadjuvant and adjuvant therapies

Variable


p value


p value

No. (%) of patients

p valueChemotherapy

n = 256

No chemotherapy

n = 461

Radiation therapy† n = 342

No radiation therapy n = 51

Hormonal therapy‡ n = 390

No hormonal therapy n = 146

Age at diagnosis, yr < 51 99 (38.7) 50 (10.8) < 0.001 63 (18.) 5 (9.8) < 0.001 89 (22.8) 16 (11.0) < 0.001 51–70 135 (52.7) 228 (49.4) 199 (58.) 20 (39.2) 210 (53.8) 52 (35.6) > 70 22 (8.6) 183 (39.7) 80 (23.) 26 (51.0) 91 (23.3) 78 (53.4)Obese Yes 91 (36.2) 161 (37.0) 0.8 130 (39.) 16 (34.0) 0.49 140 (37.0) 51 (37.2) > 0.99 No 160 (63.7) 274 (63.0) 201 (61.) 31 (66.0) 238 (63.0) 86 (62.8)Employment Employed 156 (61.7) 120 (26.7) < 0.001 143 (43.) 9 (18.0) 0.001 169 (43.9) 27 (19.1) < 0.001 Not employed 97 (38.3) 330 (73.3) 191 (57.) 41 (82.0) 216 (56.1) 114 (80.8)Marital status Married 185 (72.3) 259 (56.4) < 0.001 228 (67.) 28 (4.9.) 0.1 242 (62.2) 83 (57.2) 0.3

Not married 71 (27.7) 200 (43.6) 114 (33.) 23 (45.1) 147 (37.8) 62 (42.8)First-degree relative with breast cancer Yes 47 (18.8) 111 (26.2) 0.03 79 (24.) 13 (31.7) 0.2 85 (22.8) 36 (27.5) 0.3 No 203 (81.2) 313 (73.8) 255 (76.) 28 (68.3) 287 (77.2) 95 (72.5)Smoking status

Smoker/previous smoker

118 (46.6) 169 (36.9) < 0.001 137 (40.) 15 (30.0) 0.2 178 (45.8) 50 (34.7) 0.02

Never smoked 135 (53.4) 289 (63.1) 205 (60.) 35 (70.0) 211 (54.2) 94 (65.3)No. of comorbidities 0 93 (36.3) 90 (19.5) < 0.001 100 (29.) 5 (9.8) 0.003 97 (24.9) 28 (19.2) 0.2 ≥ 1 163 (63.7) 371 (80.5) 242 (71.) 46 (91.2) 293 (75.1) 118 (80.8)Income quintile 1 49 (19.2) 98 (21.5) 0.2 67 (20.) 11 (21.6) 0.9 76 (19.6) 32 (22.4) 0.2 2 46 (18.0) 94 (20.6) 57 (17.) 9 (17.6) 69 (17.8) 36 (25.2) 3 45 (17.6) 102 (22.4) 71 (21.) 13 (25.5) 85 (21.9) 29 (20.3) 4 56 (22.0) 83 (18.2) 68 (20.) 9 (17.6) 80 (20.6) 23 (16.1) 5 59 (23.1) 79 (17.3) 74 (22.) 9 (17.6) 78 (20.1) 23 (16.1)% of census tract population who completed college or university > 40 116 (45.5) 186 (40.5) 0.4 146 (43.) 13 (25.5) 0.05 101 (26.0) 49 (34.0) 0.01 31–40 82 (32.2) 154 (33.6) 118 (35.) 25 (49.0) 123 (31.6) 59 (41.0) < 31 57 (22.4) 119 (25.9) 76 (22.) 13 (25.5) 165 (42.4) 36 (25.0)% of census tract population who immigrated in previous 5 yr > 30 65 (25.5) 133 (29.0) 0.6 89 (26.) 18 (35.3) 0.1 101 (26.0) 43 (29.9) 0.6 21–30 83 (32.5) 140 (30.5) 105 (31.) 19 (37.2) 123 (31.6) 46 (31.9) < 21 107 (42.0) 186 (40.5) 146 (43.) 14 (27.4) 165 (42.4) 55 (38.2)Previous breast cancer Yes 22 (8.6) 64 (13.9) 0.04 26 (8.) 3 (5.9) 0.7 36 (9.2) 24 (16.4) 0.02 No 234 (91.4) 397 (86.1) 316 (92.) 48 (94.1) 354 (90.8) 122 (83.6)Disease stage (TNM) 0–1 41 (16.0) 280 (60.7) < 0.001 213 (62.) 28 (54.9) 0.3 121 (31.0) 96 (65.8) < 0.001 2 89 (34.8) 131 (28.4) 112 (33.) 18 (35.3) 145 (37.2) 40 (27.4)

3–4 126 (49.2) 50 (10.8) 17 (5.) 5 (9.8) 124 (31.8) 10 (6.8)

*Numbers do not sum to total in all cases because of missing data. †To breast and/or nodes in patients who had breast-conserving surgery. ‡In patients who were positive for estrogen/progesterone receptor.


RESEARCH


immediate reconstruction was 11.7% in Ontario.30 Ham-ilton is considered to have good access to plastic surgery, and the rate of immediate reconstruction in our cohort was 15%. We found a strong association between socio-economic status and rate of breast reconstruction in our study cohort: the odds of breast reconstruction doubled with each increase in income quintile among women aged 51–70. Past studies have suggested that patients with lower socioeconomic status have more comorbid-ities,36 which makes them less ideal candidates for tissue-based reconstruction. One study showed that the rate of complications following breast reconstruction was also higher in patients in the lowest income quintile.37 Implant-based reconstruction can be a less morbid pro-cedure than autologous reconstruction, but it requires multiple visits to the clinic and repeat surgery to exchange the expander to a permanent implant once the ideal size is reached. This may entail a �nancial burden in terms of prolonged time off work and travel costs, making it less attractive for women with lower socioeco-nomic status. Reconstruction after mastectomy has been associated with better quality of life.38 Increased public awareness and policy change are needed to make this a more available option for patients of lower socioeco-nomic status.

Strengths and limitations

Our demographic and clinical data were obtained directly from chart review, which allowed for more accurate and detailed information, with minimal missing data (about 10%, less than that in socioeconomic status studies using amalgamated data39,40). Because of the regional nature of breast cancer management in our centre, the chart review allowed us to retrieve extensive and complete data. How-ever, it was a retrospective review and does not have the potential advantages of a prospective, amalgamated data-base. We were unable to obtain individual patient infor-mation for income, education and immigration status but, rather, used the neighbourhood average from the census tract. Furthermore, we used 2006 census data, as the 2011 census was a short form that did not contain variables of interest for this study. However, it is unlikely that there was marked change in economic status within the few years from the census to the data collection period (2010–2011). If there were absolute changes in income, it is even less likely that these changes would have affected quintile distributions.

CONCLUSION

Despite universal health care funding, socioeconomic sta-tus appeared to affect breast cancer care in our Hamilton cohort in terms of neoadjuvant and adjuvant therapies, and breast reconstruction. Future directions include eval-

uation of socioeconomic status with the rate of screening and stage at diagnosis to see whether socioeconomic sta-tus has an impact before initiation of breast cancer treat-ment. It would also be interesting to evaluate whether there is any correlation between socioeconomic status, accessibility to a primary care physician, primary care practice models, and early breast cancer detection and diagnosis. Further research is required to more precisely determine how socioeconomic factors affect care, to iden-tify such disparities and make all forms of breast cancer treatment more accessible to those with lower socioeco-nomic status.

Af�liations: From the Department of Surgery, McMaster University, Hamilton, Ont. (Li, Cornacchi, Farrokhyar, Forbes, Reid, Hodgson, Lucibello, Lovrics); the Department of Surgical Oncology, Hamilton Health Sciences and Juravinski Hospital and Cancer Centre, Hamil-ton, Ont. (Forbes, Rid, Hodgson, Lovrics); the Department of Sur-gery, St. Joseph’s Healthcare, Hamilton, Ont. (Lovrics); and the Department of Medicine, McMaster University, Hamilton, Ont. (Johnston).

Funding: Funding was provided by a grant from McMaster Surgical Associates. The funding source played no role in the design, conduct or reporting of this study.


Contributors: J. Li, S. Cornacchi, F. Farrokhyar, N. Johnston, S. Forbes, S. Reid and P. Lovrics designed the study. J. Li, S. Cornacchi, N. Johnston, N. Hodgson, S. Lovrics and K. Lucibello acquired the data, which J. Li, S. Cornacchi, F. Farrokhyar, S. Reid and P. Lovrics analyzed. J. Li, S. Cornacchi and P. Lovrics wrote the article, which all authors reviewed and approved for publication.

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20. Gorey KM, Kliewer E, Holowaty EJ, et al. An international compar-ison of breast cancer survival: Winnipeg, Manitoba and De Moines, Iowa metropolitan areas. Ann Epidemiol 2003;13:32-41.

21. Gorey KM, Holowaty EJ, Fehringer F, et al. An international com-parison of cancer survival: metropolitan Toronto, Ontario and Honolulu, Hawaii. Am J Public Health 2000;90:1866-72.

22. DeLuca PF, Buist S, Johnston N. The code red project: engaging communities in health system change in Hamilton, Canada. Soc Indic Res 2012;108:317-27.

s23. Johnston N, Buist S, DeLuca PF. Code Red. Day 2: spotlight on breast cancer. Hamilton Spectator. Available: http://thespec-codered.com/?p=866 (accessed 2014 Jan. 22).

24. Christian CK, Niland J, Edge SB, et al. A multi-institutional analysis of the socioeconomic determinants of breast reconstruction: a study of the National Comprehensive Cancer Network. Ann Surg 2006; 243:241-9.

25. Zhong T, Fernandes KA, Saskin R, et al. Barriers to immediate breast reconstruction in the Canadian universal health care system. J Clin Oncol 2014;32:2133-41.

26. Forte T, Porter G, Rahal R, et al. Geographic disparities in surgery for breast and rectal cancer in Canada. Curr Oncol 2014;21:97-9.

27. Azzopardi J, Walsh D, Chong C, et al. Surgical treatment for women with breast cancer in relation to socioeconomic and insurance status. Breast J 2014;20:3-8.

28. Barnsley GP, Sigurdson L, Kirkland S. Barriers to breast recon-struction after mastectomy in Nova Scotia. Can J Surg 2008;51: 447-52.

29. Rudenstam CM, Zahrieh D, Forbes JF, et al. Randomized trial com-paring axillary clearance vs. no axillary clearance in older patients with breast cancer: �rst results of International Breast Cancer Study Group Trial 10-93. J Clin Oncol 2006;24:337-44.

30. Nguyen BC, Alawadi ZM, Roife D, et al. Do socioeconomic factors and race determine the likelihood of breast conserving surgery? Clin Breast Cancer 2016;16:e93-7.

31. Chapman T. Cancer screening bus rolls onto Hamilton streets. CBC News 2013 June 26. Available: www.cbc.ca/news/canada/hamilton/news/cancer-screening-bus-rolls-onto-hamilton-streets-1.1362640 (ac -cessed 2016 Mar. 20).

32. Craggs S. Hamilton cancer-screening bus “saved my life.” CBC News 2014 Aug. 3. Available: www.cbc.ca/news/canada/hamilton/news/hamilton-cancer-screening-bus-saved-my-life-1.2724044 (accessed 2016 Mar. 20).

33. Hershman DL, Tsui J, Wright JD, et al. Household net worth, racial disparities and hormonal therapy adherence among women with early-stage breast cancer. J Clin Oncol 2015;33:1053-9.

34. Osborne C, Ostir GV, Du X, et al. The in�uence of marital status on the stage at diagnosis, treatment, and survival of older women with breast cancer. Breast Cancer Res Treat 2005;93:41-7.

35. Platt J, Zhong T, Moineddin R, et al. Geographic variation immedi-ate and delayed breast reconstruction utilization in Ontario, Canada and plastic surgeon availability: a population-based observational study. World J Surg 2015;39:1909-21.

36. Alder NE, Newman K. Socioeconomic disparities in health: path-ways and policies. Health Aff (Millwood) 2002;21:60-76.

37. Jeevan R, Browne JP, Pereira J, et al. Socioeconomic deprivation and inpatient complication rates following mastectomy and breast recon-struction surgery. Br J Surg 2015;102:1064-70.

38. Dean C, Chetty U, Forrest AP. Effects of immediate breast recon-struction on psychosocial morbidity after mastectomy. Lancet 1983;1: 459-62.

39. Kim HM, Goodman M, Kim BI, et al. Frequency and determinants of missing data in clinical and prognostic variables recently added to SEER. J Registry Manag 2011;38:120-31.

40. Du XL, Key CR, Dickie L, et al. Information on chemotherapy and hormone therapy from tumor registry had moderate agreement with chart reviews. J Clin Epidemiol 2006;59:53-60.




Predictors of dysplastic and neoplastic progression of Barrett’s esophagus

Background: It is unknown why some cases of Barrett’s esophagus progress to inva-sive malignant disease rapidly while others do so more slowly or not at all. The aim of this study was to identify demographic and endoscopic factors that predict dysplastic and neoplastic progression in patients with Barrett’s esophagus.

Methods: Patients with Barrett’s esophagus who were assessed in 2000–2010 were assessed for inclusion in this retrospective study. Demographic and endoscopic vari-ables were collected from an endoscopy database and the medical chart. Dysplastic and neoplastic progression was examined by time-to-event analysis. We used Cox proportional hazard regression modelling and generalized estimating equation meth-ods to identify variables that were most predictive of neoplastic progression.

Results: A total of 518 patients had Barrett’s esophagus con�rmed by endoscopy and pathology and at least 2 surveillance visits. Longer Barrett’s esophagus segment (≥ 3 cm) (odds ratio [OR] 1.2, 95% con�dence interval [CI] 1.1–1.3) and increased age (≥ 60 yr) (OR 3.5, 95% CI 1.7–7.4) were independent predictors of progression from nondysplasia to dysplastic or neoplastic grades. Presence of mucosal irregulari-ties (OR 8.6, 95% CI 2.4–30.4) and increased age (OR 5.1, 95% CI 1.6–16.6) were independent predictors of progression from nondysplasia to high-grade dysplasia or adenocarcinoma.

Conclusion: Increased age, longer Barrett’s segment and presence of mucosal irregu-larities were associated with increased risk of dysplastic and neoplastic progression. In addition to dysplasia, these factors may help stratify patients according to risk of neo-plastic progression and be used to individualize surveillance. More prospective studies with larger samples are required to validate these results.

Contexte : On ignore pour quelle raison certains cas d’œsophage de Barrett évoluent rapidement vers une maladie maligne envahissante, tandis que d’autres progressent lentement ou se stabilisent. Le but de cette étude était d’identifier les facteurs démographiques et endoscopiques prédicteurs d’une progression dysplasique et néo-plasique chez les patients porteurs d’un œsophage de Barrett.

Méthodes : Des patients présentant un œsophage de Barrett ayant été examinés entre 2000 et 2010, ont été évalués en vue de leur participation à cette étude rétrospective. Les variables démographiques et endoscopiques ont été recueillies à partir d’une base de données endoscopiques et des dossiers médicaux. La progression dysplasique et néoplasique a été évaluée par analyse du délai de survenue de l’événement. Nous avons utilisé le modèle de la régression de Cox (risques proportionnels) et les équations d’estimation généralisée a�n d’identi�er les variables les plus prédictives d’une progres-sion néoplasique.

Résultats : En tout, 518 patients présentaient un œsophage de Barrett con�rmé par examen endoscopique et anatomopathologique et comptaient au moins 2 visites de surveillance. La présence de segments d’œsophage de Barrett plus longs (≥ 3 cm) (rapport des cotes [RC] 1,2, intervalle de con�ance à 95 % [IC] 1,1–1,3) et un âge avancé (≥ 60 ans) (RC 3,5, IC à 95 % 1,7–7,4) ont été des prédicteurs indépendants de progression d’un grade non dysplasique vers un grade dysplasique. La présence d’irrégu larités muqueuses (RC 8,6, IC à 95 % 2,4–30,4) et l’âge avancé (RC 5,1, IC à 95 % 1,6–16,6) ont été des prédicteurs indépendants de progression de la non- dysplasie vers une dysplasie de haut grade ou l’adénocarcinome.

Conclusion : L’âge avancé, des segments d’œsophage de Barrett plus longs et la présence d’irrégularités muqueuses ont été associés à un risque accru de progression dysplasique et néoplasique. En plus de la dysplasie, ces facteurs peuvent faciliter la

Saleh Alnasser, MD MSc Raman Agnihotram, PhD Myriam Martel, MSc Serge Mayrand, MD Eduardo Franco, MPH, DrPH Lorenzo Ferri, MD PhD

Accepted May 11, 2018

Correspondence to: L. Ferri Departments of Surgery and Oncology McGill University Montreal General Hospital 1650 Cedar Ave, Rm L8-505 Montreal QC H3G 1A4 [email protected]

DOI: 10.1503/cjs.008716

predict-alnasser.indd 93 2019-03-19 12:36 PM

RECHERCHE


T he past 3 decades have seen a dramatic increase in North America in the rate of esophageal adenocar-cinoma (EAC).1–3 Barrett’s esophagus is a well-

known premalignant esophageal condition that has the potential to progress to invasive adenocarcinoma.4,5 Although controversial, the most accepted de�nition in North America for Barrett’s esophagus follows the position supported by the American College of Gastroenterology: any length in the distal esophageal epithelium that can be recognized through endoscopy as columnar-type mucosa and con�rmed by biopsy to have intestinal metaplasia.6

The key for curable management of invasive adenocar-cinoma in patients with known Barrett’s esophagus is the detection and eradication of cancer while it is still at an early stage. Indeed, enrolment into an endoscopic surveil-lance biopsy program is the current recommendation for patients with Barrett’s esophagus.6 However, not all patients with Barrett’s esophagus are being identi�ed, as their symptoms are controlled with medications and the disorder eventually progresses to EAC without their undergoing any surveillance. At present, the recommenda-tions for the period of surveillance in patients with nondys-plastic Barrett’s esophagus ranges between 2 and 3 years.6–8 Because the rate of progression from nondysplasia to can-cer has been estimated to be 0.1%–0.5% per year,9,10 most patients may not bene�t from a routine endoscopic biopsy surveillance program, which makes the clinical effective-ness and cost-effectiveness of this program question-able.11,12 On the other hand, in some patients, there is a very rapid progression to malignant disease within the 2- or 3-year period recommended for surveillance.

Given these circumstances, a risk strati�cation for neo-plastic progression among patients with Barrett’s esopha-gus would be more ideal and would make the surveillance program more cost-effective in the long term, as it would allow for individualization of intensive surveillance for those at high risk.

In current practice, the management of patients with Barrett’s esophagus is based on the pathologic �ndings of the endoscopic biopsy sample. However, to be able to strat-ify the risk of dysplastic or neoplastic progression, addi-tional predictors are required to identify patients at higher risk for progression. When these predictors are used in conjunction with endoscopic biopsy, they may improve the ef�ciency and cost-effectiveness of surveillance programs.

The aim of this study was to identify readily available demographic and endoscopic factors that can predict the progression to any dysplastic or neoplastic grade from a baseline of nondysplastic Barrett’s esophagus. The primary objective was to identify the predictors of progression from nondysplastic Barrett’s esophagus epithelium to high-

grade dysplasia or cancer. The secondary objective was to calculate the annual incidence of each type of progression.

METHODS

All patients with a diagnosis of Barrett’s esophagus who were evaluated at the McGill University Health Centre, Montréal, between January 2000 and December 2010 were assessed for inclusion. We reviewed demographic and endo-scopic data for each patient con�rmed by endoscopy and pathology to have Barrett’s esophagus. We maintained strict criteria for Barrett’s esophagus: endoscopic (columnar- lined mucosa) and pathologic (intestinal metaplasia) crite-ria were required. For identi�cation of predictors of pro-gression, patients were included only when there was 1) a diagnosis of Barrett’s esophagus that was con�rmed on both pathologic and endoscopic �ndings, 2) 2 or more endoscopic surveillance visits carried out in order to assess progression status and 3) no prior history of esophageal cancer or esophageal/gastric resection.

Patients were identified through a comprehensive pathology database of the McGill University Health Cen-tre. All patients with intestinal metaplasia of the esophagus were captured through this database. Their demographic and endoscopic �ndings were then obtained from both an endoscopy database and the medical chart. Demographic variables included age and sex; endoscopic variables included endoscopy date, information on hiatal hernia presence, presence of esophagitis, presence of ulcer, Barrett’s segment length, presence of mucosal irregular-ities (nodules, irregular mucosa or polyps) and presence of stricture. Smoking, alcohol use and obesity were not included as potential variables owing to lack of adequate documentation of these variables in the older charts.

We classified patients into 3 groups based on their baseline pathologic status at the �rst endoscopy: 1) non-dysplastic Barrett’s esophagus, 2) inde�nite of dysplasia or low-grade dysplasia or 3) high-grade dysplasia or EAC. To identify predictors of dysplastic and neoplastic progres-sion, we analyzed 2 different stages of progression: from nondysplastic Barrett’s esophagus to any dysplastic or neo-plastic grade, and from nondysplastic Barrett’s esophagus to high-grade dysplasia or EAC.


We used the statistical software program Stata 12 (Stata-Corp) for all analyses. We calculated the incidence rate (cases per 1000 person-years) for each type of progression outcome being investigated among the patients included in the �nal analysis. Length of time for progression from

strati�cation des patients selon le risque de progression néoplasique et servir à indi-vidualiser la surveillance. Il faudra procéder à d’autres études prospectives auprès d’échantillons de population plus volumineux pour valider ces résultats.


RESEARCH


nondysplastic Barrett’s esophagus to premalignant/ malignant lesions was initially examined by time-to-event analysis. We used Kaplan–Meier curves and log-rank tests to see the survival difference among selected categories.

We used age-adjusted Cox regression models to esti-mate risk factors for progression. For multivariate model-ling, we applied an empirical estimation in which all the independent variables were examined for their covariate effects. We analyzed demographic and endoscopic char-acteristics using the low-risk category as the referent group. Two-sided p values < 0.05 were considered to be statistically signi�cant. Owing to the nature of the data and availability at multiple observations per patient, we used generalized estimating equation models to calculate how the progression at an index visit was predictive from the variables of the previous visit. We examined all inde-pendent variables to estimate their covariate in�uence in order to identify factors that were most predictive of progression.

RESULTS

From January 2000 to December 2010, 1054 patients with intestinal metaplasia of the esophagus were identi�ed in the pathology database. A total of 303 patients were excluded because there was no documentation of columnar-lined mucosa on endoscopy or we were unable to locate the endoscopy report, leaving 751 patients who were con�rmed to have both pathologic and endoscopic diagnosis of Barrett’s esophagus. Of the 751, 518 had 2 or more surveillance visits and were thus eligible for the pro-gression analysis.

Baseline demographic and endoscopic characteristics of the 518 patients are shown in Table 1. Among the 518, 458 had Barrett’s esophagus without any degree of dys-plasia, and in 60, the disorder had progressed to any dys-plastic grade or cancer. The incidence of progression to any dysplastic grade or cancer was 19.8 cases per 1000 person-years (95% con�dence interval [CI] 14.2–27.6).

Table 1. Baseline characteristics of patients with Barrett’s esophagus

Characteristic

No. (%) of patients

All patients n = 518

Nondysplastic n = 458

Indefinite of dysplasia/ low-grade dysplasia

n = 33

High-grade dysplasia/EAC

n = 27

Age, yr

< 60 263 (50.8) 240 (52.4) 15 (45.4) 8 (29.6)

≥ 60 255 (49.2) 218 (47.6) 18 (54.5) 19 (70.4)

Sex

Male 372 (71.8) 319 (69.7) 28 (84.8) 25 (92.6)

Female 146 (28.2) 139 (30.3) 5 (15.2) 2 (7.4)

Hiatal hernia

No 209 (40.3) 187 (40.8) 11 (33.3) 11 (40.7)

Yes 143 (27.6) 124 (27.1) 11 (33.3) 8 (29.6)

Missing 166 (32.0) 147 (32.1) 11 (33.3) 8 (29.6)

Esophagitis

No 269 (51.9) 234 (51.1) 17 (51.5) 18 (66.7)

Yes 83 (16.0) 77 (16.8) 5 (15.2) 1 (3.7)

Missing 166 (32.0) 147 (32.1) 11 (33.3) 8 (29.6)

Ulcer

No 318 (61.4) 281 (61.4) 18 (54.5) 19 (70.4)

Yes 34 (6.6) 30 (6.6) 4 (12.1) 0 (0.0)

Missing 166 (32.0) 147 (32.1) 11 (33.3) 8 (29.6)

Length of Barrett’s esophagus segment, cm

< 3 49 (9.4) 40 (8.7) 5 (15.2) 4 (14.8)

≥ 3 cm 98 (18.9) 81 (17.7) 8 (24.2) 9 (33.3)

Missing 371 (71.6) 337 (73.6) 20 (60.6) 14 (51.9)

Mucosal irregularities

No 338 (65.2) 305 (66.6) 20 (60.6) 13 (48.1)

Yes 14 (2.7) 6 (1.3) 2 (6.1) 6 (22.2)

Missing 166 (32.0) 147 (32.1) 11 (33.3) 8 (29.6)

Stricture

No 341 (65.8) 301 (65.7) 21 (63.6) 19 (70.4)

Yes 11 (2.1) 10 (2.2) 1 (3.0) 0 (0.0)

Missing 166 (32.0) 147 (32.1) 11 (33.3) 8 (29.6)

EAC = esophageal adenocarcinoma.


RECHERCHE


High-grade dysplasia or EAC developed in 10 patients, with an overall progression rate of 5.5 cases per 1000 person-years (95% CI 2.9–10.3).

The cumulative incidence of progression to dysplastic or neoplastic lesions by age and by length of the Barrett’s esoph-agus segment is shown in Figure 1 and Figure 2, respectively. The cumulative incidence of progression to high-grade dys-plasia/EAC by age and by presence of mucosal irregularities is shown in Figure 3 and Figure 4, respectively.

Univariate analyses of the effect of potential factors on progression from nondysplastic Barrett’s esophagus to any dysplastic or neoplastic grade showed that length of the Barrett’s esophagus segment of 3 cm or greater was a significant predictor of progression (p = 0.03). Presence of ulcer (p = 0.01) and presence of mucosal irregularities (p = 0.001) were signi�cant predictors of progression from nondysplastic Barrett’s esophagus to high-grade dysplasia/EAC.

Multivariate Cox regression analyses showed that only length of the Barrett’s esophagus segment of 3 cm or greater was an independent predictor of progression from nondysplastic Barrett’s esophagus to any dysplastic or neo-plastic grade (hazard ratio [HR] 2.2, 95% CI 1.1–4.7) (Table 2). The presence of mucosal irregularities was an independent predictor of progression from nondysplasia to high-grade dysplasia/EAC (HR 4.6, 95% CI 1.1–18.3) (Table 3). Other factors carried no predictive value for dysplastic or neoplastic progression.

Generalized estimating equation analysis gave almost identical results to the Cox regression analyses. Length of the Barrett’s esophagus segment of 3 cm or greater (odds ratio [OR] 1.2, 95% CI 1.1–1.3) and increased age (≥ 60 yr) (OR 3.5, 95% CI 1.7–7.4) were the only independent pre-dictors of progression to any dysplastic and neoplastic grade (Table 4). Presence of mucosal irregularities (OR 8.6, 95% CI 2.4–30.4) and increased age (OR 5.1,

Fig. 1. Cumulative incidence of progression from nondysplastic Barrett’s esophagus to dysplastic or neoplastic grades by age. There was no difference between age groups (p = 0.1).

< 60 yr ≥ 60 yr

0.00

0.10

0.20

0.30

0.40

Inci

denc

e of

pro

gres

sion

0 24 48

Length of follow-up, mo72 96 120

Fig. 2. Cumulative incidence of progression from nondysplastic Barrett’s esophagus to dysplastic or neoplastic by length of Barrett’s esophagus segment. Longer length was a significant predictor of progression (p = 0.03).

< 3 cm

≥ 3 cm

0.00

0.10

0.20

0.30

0.40

Inci

denc

e of

pro

gres

sion

0 24 48

Length of follow-up, mo72 96 120

Fig. 3. Cumulative incidence of progression from nondysplastic Bar-rett’s esophagus to high-grade dysplasia or esophageal adenocarci-noma by age. There was no difference between age groups (p = 0.07).

< 60 yr≥ 60 yr

0 12 24 36 48 60

Inci

denc

e of

pro

gres

sion

Length of follow-up, mo

0.00

0.05

0.10

0.15

0.20

Fig. 4. Cumulative incidence of progression from nondysplastic Barrett’s esophagus to high-grade dysplasia or esophageal adenocar-cinoma by presence of mucosal irregularities. Presence of mucosal irregularities was a significant predictor of progression (p = 0.001).

No mucosal irregularitiesMucosal irregularities

0.20

0.15

0.10

0.05

0.00

0 12 24 36 48 60

Inci

denc

e of

pro

gres

sion

Length of follow-up, mo


RESEARCH


95% CI 1.6–16.6) were the only signi�cant independent predictors of progression to high-grade dysplasia/EAC.

DISCUSSION

In this study, we examined 2 different types of dysplastic and neoplastic progression in patients with Barrett’s esoph-agus. Consistent with previous studies,13,14 we found that most patients remained in a nondysplastic condition over a median follow-up period of almost 5 years. However, for those who progressed from nondysplastic Barrett’s esopha-gus to any dysplastic or neoplastic grade, only a longer Barrett’s esophagus segment and age 60 years or more were independent predictors. Presence of mucosal irregu-larities and age 60 years or more were independent predic-tors of progression from nondysplastic Barrett’s esophagus to high-grade dysplasia/EAC. This suggests that patients with these risk factors should perhaps be prioritized for shorter intervals of surveillance, and those deemed at low risk based on our �ndings could perhaps undergo surveil-lance at longer intervals.

Age as a risk factor for dysplastic and neoplastic pro-gression of Barrett’s esophagus is considered controversial, as it has been reported in some studies15,16 but not in oth-ers.17–19 In the present study, age was an independent pre-dictor in the generalized estimating equation model but not in the multivariate Cox regression analysis. However, in Cox regression, the HR was greater than 1, which sug-gests that age likely has some impact, albeit not statistically signi�cant. This is likely a type II error that would have become signi�cant in the multivariate analysis if we had had a larger sample.

Sex did not play a signi�cant role in dysplastic or neo-plastic progression in the current study. There is disagree-ment in the literature about the role of sex in progression of Barrett’s esophagus.16,19,20 Nevertheless, a higher male to female ratio among patients with EAC21 may possibly be explained by the �nding that Barrett’s esophagus develops at a younger age in men than in women.21,22

Length of the Barrett’s esophagus segment is frequently described as a risk factor for malignant progression; the longer the segment, the greater the risk of progression.23,24 In our study, patients with a Barrett’s epithelial segment of 3 cm or greater were more likely to progress to dysplastic and neoplastic grades than those with a length less than 3 cm. However, this factor was not a signi�cant predictor of progression to high-grade dysplasia/EAC. This may have been due to the small number of patients who pro-gressed on this path. This suggest that patients with a longer Barrett’s esophagus segment should undergo more intense surveillance. However, surveillance of patients with a shorter Barrett’s esophagus segment should not be limited.

In previous studies, the presence of a hiatal hernia was more common among patients who progressed to dysplastic

Table 2. Multivariate analysis of potential risk factors for progression from nondysplastic Barrett’s esophagus to dysplastic or neoplastic grades (Cox regression)

Variable HR (95% CI)

Age, yr

< 60 1.0

≥ 60 1.9 (0.9–3.9)


< 3 1.0

≥ 3 2.2 (1.1–4.7)

CI = confidence interval; HR = hazard ratio.

Table 3. Multivariate analysis of potential risk factors for progression from nondysplastic Barrett’s esophagus to high-grade dysplasia or esophageal adenocarcinoma (Cox regression)

Variable HR (95% CI)

Age, yr

< 60 1.0

≥ 60 3.2 (0.8–12.6)


No 1.0

Yes 4.5 (1.1–18.3)

Ulcer

No 1.0

Yes 3.6 (0.9–14.6)


Table 4. Multivariate analysis of potential risk factors for progression (generalized estimating equation analysis)

Variable

OR (95% CI)

Progression to any dysplasia

Progression to high-grade

dysphasia/EAC

Age, yr

< 60 1.0 1.0

≥ 60 3.5 (1.7–7.4) 5.1 (1.6–16.6)

Sex

Female 1.0 1.0

Male 2.1 (0.8–5.6) 2.6 (0.6–11.1)

Hiatal hernia

No 1.0 1.0

Yes 0.8 (0.5–1.3) 0.9 (0.5–1.3)

Esophagitis

No 1.0 1.0

Yes 0.6 (0.3–1.4) 1.2 (0.4–3.2)

Ulcer

No 1.0 1.0

Yes 1.9 (0.9–4.5) 0.9 (0.2–1.17)


No 1.0 1.0

Yes 3.3 (0.8–13.8) 8.6 (2.4–30.4)


< 3 1.0 1.0

≥ 3 1.2 (1.1–1.3) 1.1 (0.9–1.17)

CI = confidence interval; EAC = esophageal adenocarcinoma; OR = odds ratio.


RECHERCHE


or neoplastic grades than among those who did not progress owing to disruption of the physiologic antire�ux barrier, which creates a greater probability of dysplastic and neoplas-tic progression.17,25 The presence of a hiatal hernia was not signi�cant enough to serve as an independent predictor in our study and that by Sikkema and colleagues.26 One possi-ble reason for this discrepancy is the lack of documentation of hiatal hernia in endoscopy reports.

Esophagitis can play an important role in dysplastic and neoplastic progression, as the underlying inflammation may increase the possibility of abnormal proliferation and neoplastic mutation.27 In the current study, esophagitis was not a strong predictor of progression. One possible reason for this is the aggressive treatment of esophagitis with pro-ton pump inhibitors, which reduce the in�ammatory pro-cess while at the same time reducing progression. Simi-larly, complications of esophagitis such as esophageal ulcers and strictures were not signi�cant predictors of pro-gression. Their role in dysplastic and neoplastic progres-sion in patients with Barrett’s esophagus is controversial. Some studies28,29 showed that patients with esophageal ulcers or strictures were more likely to progress to high-grade dysplasia and cancer than those without, whereas the study by Sikkema and colleagues26 showed no link. In the current study, neither factor was an independent predictor of dysplastic or neoplastic progression.

Several studies have con�rmed the presence of mucosal irregularities, including nodules, nodular mucosa and pol-ypoid lesions, to be associated with a higher risk of pro-gression to high-grade dysplasia and EAC.30–32 In the pres-ent study, the presence of mucosal irregularities was an independent predictor of progression from nondysplastic Barrett’s esophagus to high-grade dysplasia/EAC only and not of progression to any dysplastic or neoplastic grade. This may have been due to the presence of nodules, which can increase the risk of high-grade dysplasia or EAC spe-ci�cally. Another possible reason may be the frequent use of endoscopic mucosal resection instead of conventional biopsy of a large suspicious area that might include a focus of cancer or high-grade dysplasia.

Limitations

The retrospective nature of our study and the lack of strict control over different variables are disadvantages compared to a prospective study. In addition, endoscopy reports were missing for some patients and/or for some visits for differ-ent patients. The retrospective nature of the study also made the identi�cation of lifestyle and clinical variables challenging and sometimes impossible. We were unable to investigate the role of smoking, alcohol or obesity owing to poor documentation of these variables in the older charts; in addition, the pattern of these variables may be modi�ed at the time of diagnosis of Barrett’s esophagus. There is also the possibility of recall bias, as most of our patients received

the diagnosis several years ago. Furthermore, we were unable to address the role of molecular markers in progres-sion of Barrett’s esophagus because of the retrospective nature of the study and the fact that these molecular mark-ers were not routinely tested at our institution. Some patients have no symptoms and may progress to the dys-plastic or even neoplastic stage without being identi�ed until the late neoplastic stage. These patients may be dif�-cult to identify or stratify for risk as they are asymptomatic and will not undergo any kind of investigation. They will not bene�t from risk strati�cation. Another limitation is the lack of a standardized biopsy protocol at our institution, and some endoscopists performed surveillance biopsies without following the proper guidelines. In general, this lack of attention to protocol reduces the chance of detecting dys-plasia or cancer in patients with Barrett’s esophagus, and this may have reduced the overall number of patients with progression in our cohort. There is also the possibility that some patients who had follow-up visits in our centre may have moved to another city, progressed, sought medical advice in other centres or a different city, or died, and some may have simply withdrawn from the surveillance program offered to them. These factors may have affected the results.

CONCLUSION

Our study has shown that a longer length of the Barrett’s esophagus segment, increased patient age and the presence of mucosal irregularities are all predictive of neoplastic progression of Barrett’s esophagus. These factors may be used together with the pathologic grade to individualize the frequency of surveillance based on the estimated risk. However, more prospective studies with larger samples are required to validate these results.

Af�liations: From the Department of General Surgery, McGill Uni-versity Health Centre, Montreal, Que. (Alnasser); the Department of Epidemiology and Biostatistics – Cancer Epidemiology, McGill Univer-sity, Montreal, Que. (Agnihotram, Franco); and the Ringgold Standard Institution – Gastroenterology, McGill University Health Centre, Montréal, Que. (Martel, Mayrand, Ferri).


Contributors: S. Alnasser and L. Ferri designed the study. S. Alnasser, R. Agnihotram, M. Martel and S. Mayrand acquired the data, which S. Alnasser, R. Agnihotram and E. Franco analyzed. S. Alnasser and R. Agnihotram wrote the article, which L. Ferri revised and all authors reviewed and approved for publication.

References

1. Devesa SS, Blot WJ, Fraumeni JF Jr. Changing patterns in the inci-dence of esophageal and gastric carcinoma in the United States. Cancer 1998;83:2049-53.

2. El-Serag HB, Mason AC, Petersen N, et al. Epidemiological differ-ences between adenocarcinoma of the oesophagus and adenocarci-noma of the gastric cardia in the USA. Gut 2002;50:368-72.

3. Blot WJ, Devesa SS, Kneller RW, et al. Rising incidence of adenocar-cinoma of the esophagus and gastric cardia. JAMA 1991;265: 1287-9.


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4. Vakil N, van Zanten SV, Kahrilas P, et al. The Montreal de�nition and classification of gastroesophageal reflux disease: a global evidence-based consensus. Am J Gastroenterol 2006;101:1900-20.

5. Overholt BF, Lightdale CJ, Wang KK, et al. Photodynamic therapy with por�mer sodium for ablation of high-grade dysplasia in Barrett’s esophagus: international, partially blinded, randomized phase III trial. Gastrointest Endosc 2005;62:488-98.

6. Wang KK, Sampliner RE; Practice Parameters Committee of the American College of Gastroenterology. Updated guidelines 2008 for the diagnosis, surveillance and therapy of Barrett’s esophagus. Am J Gastroenterol 2008;103:788-97.

7. Boyer J, Laugier R, Chemali M, et al. French Society of Digestive Endoscopy SFED guideline: monitoring of patients with Barrett’s esophagus. Endoscopy 2007;39:840-2.

8. Fitzgerald RC, Di Pietro M, Ragunath K, et al.; British Society of Gastroenterology. British Society of Gastroenterology guidelines on the diagnosis and management of Barrett’s oesophagus. Gut 2014;63: 7-42.

9. O’Connor JB, Falk GW, Richter JE. The incidence of adenocarci-noma and dysplasia in Barrett’s esophagus: report on the Cleveland Clinic Barrett’s Esophagus Registry. Am J Gastroenterol 1999;94: 2037-42.

10. Drewitz DJ, Sampliner E, Garewal HS. The incidence of adenocar-cinoma in Barrett’s esophagus: a prospective study of 170 patients followed 4.8 years. Am J Gastroenterol 1997;92:212-5.

11. Provenzale D, Schmitt C, Wong JB. Barrett’s esophagus: a new look at surveillance based on emerging estimates of cancer risk. Am J Gas-troenterol 1999;94:2043-53.

12. Garside R, Pitt M, Somerville M, et al. Surveillance of Barrett’s oesophagus: exploring the uncertainty through systematic review, expert workshop and economic modelling. Health Technol Assess 2006; 10:1-142, iii-iv.

13. Reid BJ, Blount PL, Feng Z, et al. Optimizing endoscopic biopsy detection of early cancers in Barrett’s high-grade dysplasia. Am J Gastroenterol 2000;95:3089-96.

14. Schnell TG, Sontag SJ, Chejfec G, et al. Long-term nonsurgical management of Barrett’s esophagus with high-grade dysplasia. Gas-troenterology 2001;120:1607-19.

15. Gopal DV, Lieberman DA, Magaret N, et al. Risk factors for dyspla-sia in patients with Barrett’s esophagus (BE): results from a multi-center consortium. Dig Dis Sci 2003;48:1537-41.

16. Gatenby PA, Caygill CP, Ramus JR, et al. Barrett’s columnar-lined oesophagus: demographic and lifestyle associations and adenocarci-noma risk. Dig Dis Sci 2008;53:1175-85.

17. Avidan B, Sonnenberg A, Schnell TG, et al. Hiatal hernia size, Barrett’s length, and severity of acid re�ux are all risk factors for esophageal adenocarcinoma. Am J Gastroenterol 2002;97:1930-6.

18. De Jonge PJ, Steyerberg EW, Kuipers EJ, et al. Risk factors for the development of esophageal adenocarcinoma in Barrett’s esophagus. Am J Gastroenterol 2006;101:1421-9.

19. Oberg S, Wenner J, Johansson J, et al. Barrett esophagus: risk factors for progression to dysplasia and adenocarcinoma. Ann Surg 2005; 242:49-54.

20. Bani-Hani K, Sue-Ling H, Johnston D, et al. Barrett’s oesophagus: results from a 13-year surveillance programme. Eur J Gastroenterol Hepatol 2000;12:649-54.

21. Van Soest EM, Dieleman JP, Siersema PD, et al. Increasing inci-dence of Barrett’s oesophagus in the general population. Gut 2005; 54:1062-6.

22. Van Blankenstein M, Looman CW, Johnston BJ, et al. Age and sex distribution of the prevalence of Barrett’s esophagus found in a pri-mary referral endoscopy center. Am J Gastroenterol 2005;100:568-76.

23. Iftikhar SY, James PD, Steele RJ, et al. Length of Barrett’s oesopha-gus: an important factor in the development of dysplasia and adeno-carcinoma. Gut 1992;33:1155-8.

24. Weston AP, Sharma P, Mathur S, et al. Risk strati�cation of Barrett’s esophagus: updated prospective multivariate analysis. Am J Gastroen-terol 2004;99:1657-66.

25. Weston AP, Badr AS, Hassanein RS. Prospective multivariate analy-sis of clinical, endoscopic, and histological factors predictive of the development of Barrett’s multifocal high-grade dysplasia or adeno-carcinoma. Am J Gastroenterol 1999;94:3413-9.

26. Sikkema M, Looman CW, Steyerberg EW, et al. Predictors for neo-plastic progression in patients with Barrett’s esophagus: a prospective cohort study. Am J Gastroenterol 2011;106:1231-8.

27. Moons LM, Kusters JG, Bultman E, et al. Barrett’s oesophagus is characterized by a predominantly humoral in�ammatory response. J Pathol 2005;207:269-76.

28. Montgomery E, Bronner MP, Greenson JK, et al. Are ulcers a marker for invasive carcinoma in Barrett’s esophagus? Data from a diagnostic variability study with clinical follow‐up. Am J Gastroenterol 2002;97:27-31.

29. Hillman LC, Chiragakis L, Clarke AC, et al. Barrett’s esophagus: macroscopic markers and the prediction of dysplasia and adenocarci-noma. J Gastroenterol Hepatol 2003;18:526-33.

30. Buttar NS, Wang KK, Sebo TJ, et al. Extent of high‐grade dysplasia in Barrett’s esophagus correlates with risk of adenocarcinoma. Gas-troenterology 2001;120:1630-9.

31. Thurberg BL, Duray PH, Odze RD. Polypoid dysplasia in Barrett’s esophagus: a clinicopathologic, immunohistochemical, and molecular study of �ve cases. Hum Pathol 1999;30:745-52.

32. Arnold GL, Mardini HE. Barrett’s esophagus‐associated polypoid dysplasia: case report and review of the literature. Dig Dis Sci 2002; 47:1897-900.




High-risk medications in older patients with trauma: a cross-sectional study of risk mitigation

Erica Lester, MD, MSc Mark Dykstra, MD Chantalle Grant, MD Vanessa Fawcett, MD, MPH Bonnie Tsang, MD Sandy Widder, MD, MHA, MSc

Presented at the Trauma Association of Canada Annual Scientific Meeting and Conference, Feb. 23–24, 2017, Vancouver, British Columbia.

Accepted June 4, 2018

Correspondence to: E. Lester Department of Surgery University of Alberta Hospital 8440 112 Street NW Edmonton AB T6G 2B7 [email protected]

DOI: 10.1503/cjs.017117

Background: The Beers Criteria for Potentially Inappropriate Medication Use in Older Adults is a framework that can assess the nature of high-risk medication use. The objective of this study was to use the Beers Criteria to assess the prevalence and nature of polypharmacy, the magnitude of medication changes during the hospital stay and the impact of Beers Criteria medications on outcomes in older patients with trauma.

Methods: We used the Alberta Trauma Registry to conduct a retrospective review of patients aged 65 years or older with major trauma (Injury Severity Score ≥ 12) who were admitted to a level 1 trauma centre between January 2013 and December 2014. We analyzed changes in medication prescriptions during the hospital stay using descriptive statistics and assessed the association between Beers Criteria med-ications and relevant outcomes using multivariable regression analysis.

Results: There was no signi�cant change in the number of Beers Criteria medica-tions prescribed before and after admission. The adjusted odds ratio for 30-day mor-tality for each additional Beers Criteria medication prescribed was 2.02 (95% con�-dence interval [CI] 1.16–3.51) versus 1.24 (95% CI 1.04–1.59) for each additional medication of any type. The corresponding adjusted incidence rate ratios for length of stay were 1.15 (95% CI 1.03–1.30) versus 1.05 (95% CI 1.01–1.10).

Conclusion: Beers Criteria medications were not discontinued during trauma admis-sions. Using the Beers Criteria as a process indicator for quality improvement in trauma care may provide interdisciplinary trauma teams an opportunity to audit patient medi-cations and stop potentially harmful medications in a vulnerable population.

Contexte : Les critères de Beers sur les médicaments potentiellement inappropriés chez les adultes âgés constituent un cadre qui permet d’évaluer la nature d’une pharmacothérapie à risque élevé. L’objectif de cette étude était d’utiliser les critères de Beers pour évaluer la prévalence et la nature de la polypharmacologie, l’ampleur des changements de prescriptions en cours d’hospitalisation et l’impact des médica-ments potentiellement inappropriés sur l’évolution de l’état de personnes âgées vic-times de traumatismes.

Méthodes : Nous avons utilisé le Registre albertain des traumatismes pour procéder à une revue rétrospective des patients de 65 ans et plus victimes d’un traumatisme grave (indice de gravité des blessures ≥ 12) admis dans un centre de traumatologie entre janvier 2013 et décembre 2014. Nous avons analysé les changements de médi-caments prescrits durant le séjour hospitalier au moyen de statistiques descriptives et nous avons évalué le lien entre les médicaments potentiellement inappropriés et les variables pertinentes au moyen d’une ana lyse de régression multivariée.

Résultats : On n’a noté aucun changement signi�catif entre les médicaments poten-tiellement inappropriés prescrits avant et après l’admission. Le rapport des cotes ajusté pour la mortalité à 30 jours pour chaque médicament potentiellement inappro-prié prescrit a été de 2,02 (intervalle de con�ance [IC] à 95 % 1,16–3,51) contre 1,24 (IC à 95 % 1,04–1,59) pour chaque médicament additionnel, de tout type. Les rap-ports des taux d’incidence ajustés correspondants pour la durée de l’hospitalisation ont été de 1,15 (IC à 95 % 1,03–1,30) contre 1,05 (IC à 95 % 1,01–1,10).

Conclusion : Les médicaments potentiellement inappropriés (selon les critères de Beers) n’ont pas été cessés durant les admissions pour traumatisme. L’utili sation des critères de Beers comme indicateur de processus pour l’amélioration de la qua lité des soins en traumatologie pourrait fournir aux équipes interdisciplinaires une occasion de véri�er les médicaments prescrits et de cesser ceux qui sont nuisibles à une population vulnérable.

high-lester.indd 100 2019-03-19 11:31 AM

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T he proportion of the population over the age of 65 years is growing throughout the developed world.1 In Canada, the proportion of the population

in this age group increased from 8% to 16% between 1970 and 2015, and is expected to reach 23% by 2030.1,2 Trau-matic injuries in older people have also increased: in 2009, nonintentional injury was the �fth-leading cause of hospi-tal admission and the sixth-leading cause of death among Canadians aged 65–74.3–5 Older adults experience higher rates of mortality and morbidity than those of average age.6–10 Substantial resource use is associated with trauma in older people, both directly, owing to health care expen-ditures, and indirectly, through increased care needs, pro-ductivity losses and premature loss of life.11,12 The Ameri-can College of Surgeons has acknowledged these issues and has developed dedicated geriatric trauma management guidelines as part of its Trauma Quality Improvement Program.13,14

The incidence and severity of traumatic injuries in this age group has numerous associations, including comor-bidities, sensorium deterioration, substance abuse and polypharmacy.14–18 Polypharmacy has been the focus of multiple studies that have identi�ed associations between preinjury medications and subsequent morbidity and mortality.14,16,19,20 The updated Beers Criteria for Poten-tially Inappropriate Medication Use in Older Adults lists medications for which the risks of use may outweigh the bene�ts in patients over the age of 65.21,22 These guide-lines were developed and updated through systematic lit-erature reviews by an expert interdisciplinary panel appointed by the American Geriatrics Society.22 The medications are organized into subcategories based on organ system/therapeutic category. The criteria are prog-nostic for increased hospital admissions and length of stay, and their use has been effective in reducing poly-pharmacy in older adults.23,24 Since older patients who have experienced injury are more than 3 times more likely to be readmitted with recurrent trauma than similar older patients with no trauma history, reducing the risk of reoc-currence through reduction of high-risk medications is a logical prevention measure.25 The American College of Surgeons has recommended the use of the Beers Criteria to make pharmacotherapy decisions during trauma admis-sions in this population.14

To our knowledge, there is no literature exploring the role of the Beers Criteria as a process measure for inpa-tients with trauma.21,26 The objective of this study was to use the Beers Criteria as a framework to assess the preva-lence and nature of polypharmacy, the magnitude of medi-cation changes during the hospital stay and the impact of Beers Criteria medications on outcomes (length of hospital stay and 30-day mortality) in older patients who experi-enced trauma. This information will inform the suitability of the Beers Criteria as a guideline and indicator for medi-cation review in this patient population.

METHODS

This study was approved by the Health Research Ethics Board at the University of Alberta. We performed a retro-spective study of patients aged 65 years or more with trauma who were admitted to a level 1 trauma centre within 24 hours of injury between January 2013 and December 2014. Patients were identi�ed from the Alberta Trauma Registry, a prospectively maintained database of patients with major trauma (Injury Severity Score ≥ 1227). Out-of-province patients and patients who died during ini-tial resuscitation were excluded. Chart review collected data on demographic characteristics, injury mechanism, hospital and intensive care unit length of stay, 30-day mor-tality, in-hospital complications, clinical service that dis-charged the patient, �nal disposition, and pre- and post-admission prescription medications.


We used basic descriptive statistics to delineate medication use and change in the number of Beers Criteria medica-tions prescribed before and after the injury. It was deter-mined that a negative binomial approach was the optimal modelling approach for predicting length of stay. We adjusted the outcome variable length of stay for preadmis-sion medications (independent variable of interest) as well as the covariables age, sex, Injury Severity Score, number of operations during the hospital stay and admission to the intensive care unit. Table 1 outlines the models. Model 1 adjusted for all covariables. Model 2 repeated this process but instead adjusted for number of Beers Criteria medica-tions only, rather than all medications at admission. Model 3 considered the effect of each subcategory of the Beers Criteria list, adjusting for a series of binary variables corresponding to the presence of a medication from each class, along with all other independent variables. Model 4 considered binary variables representing Beers and non-Beers medications, along with all other independent vari-ables. To explore the impact these medications had on

Table 1. Statistical model description

Model no. Outcome variable Independent variable of interest

1 Length of stay Number of preadmission medications

2 Length of stay Number of preadmission Beers Criteria medications

3 Length of stay Binary representation of Beers Criteria categories

4 Length of stay Binary representations of presence of Beers and non-Beers medications

5 30-day mortality Number of preadmission medications

6 30-day mortality Number of preadmission Beers Criteria medications

Beers Criteria = Beers Criteria for Potentially Inappropriate Medication Use in Older Adults.21,22


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30-day mortality, we created logistic regression models. Model 5 adjusted mortality for the total number of medi-cations prescribed before admission and all other covari-ables listed above, and model 6 adjusted for the number of Beers Criteria medications prescribed before admission, along with the previously described covariables. We calcu-lated standard errors using both asymptotic theory (oim) and bootstrapping, with minimal difference. We report oim results. We compared each model permutation using the Akaike information criterion and the Bayesian infor-mation criterion or area under the receiver operator curve and the Pearson χ2 test, as applicable. All analyzes were conducted with Stata version 13 (StataCorp).

RESULTS

During the 2-year study period, 319 patients met the inclusion criteria. Demographic and descriptive data are reported in Table 2. The median length of stay in the trauma centre was 8 days (range 1–155 d), and the mean was 16.8 days (95% confidence interval [CI] 14.37 to 19.31). The 30-day mortality rate was 1.88% (Table 2). The mean number of prescribed medications on admission was 4.36 (95% CI 3.93 to 4.78), with 204 patients (63.9%) taking at least 1 Beers Criteria medication and 100 (31.3%) taking 2 or more Beers Criteria medications. The mean change in total number of medications from admission to discharge was –0.52 (95% CI –0.81 to –0.23) (range –9 to 13). The mean change in number of Beers Criteria medi-cations was 0.03 (95% CI –0.08 to 0.13) (range –3 to 5). Of the patients who were taking Beers Criteria medications on admission, 92 (45.1%) were discharged taking the same number as on admission. There was no association

between the clinical service (e.g., general surgery, orthope-dic surgery) that discharged the patient and the number of prescribed medications or the change in number of medications.

Among the subcategories of Beers Criteria medications on admission, 69 patients (33.8%) were taking central ner-vous system drugs, 59 (28.9%) were taking gastrointestinal medications, 55 (27.0%) were taking cardiac medications, 9 (4.4%) were taking anticoagulants, 8 (3.9%) were taking antithrombotics, 7 (3.4%) were taking pain medications, 7 (3.4%) were taking endocrine medications, and 2 (1.0%) were taking antibiotics. These proportions were unaltered on discharge. Of note, only 3 patients who were not taking prescription pain medication before the injury were dis-charged home on new pain medication.

In multivariable regression analysis model 1, each addi-tional admission medication was associated with an inci-dence rate ratio (IRR) of 1.05 (95% CI 1.01 to 1.10). In model 2, each additional Beers Criteria medication was associated with an IRR of 1.15 (95% CI 1.03 to 1.30), rep-resenting an increased length of stay.

In model 3, only central nervous system medications had a statistically signi�cant impact on length of stay, with an IRR of 1.58 (95% CI 1.21 to 2.00). In model 4, Beers Criteria medications had an IRR of 1.46 (95% CI 1.13 to 1.88) for length of stay. Non-Beers medications were not statistically signi�cant predictors of length of stay.

With respect to 30-day mortality, model 5 showed that a 1-unit increase in any medication on admission had an adjusted odds ratio (OR) of 1.24 (95% CI 1.04 to 1.59). Model 6 showed that a 1-unit increase in the num-ber of Beers Criteria medications was associated with an adjusted OR of 2.02 (95% CI 1.16 to 3.51). The area under the receiver operator curve values were 0.81 and 0.77, respectively.

DISCUSSION

As the population ages, attention must be directed toward providing quality trauma care for older patients. Polyphar-macy is a known risk factor for hospital admission and injury in older adults and is a potential target for improve-ment in trauma outcomes. Moreover, a previous trauma admission is a risk factor for future traumatic injury, mak-ing polypharmacy reduction during a trauma admission a prospective injury prevention mechanism.25 The Beers Criteria can serve as both a clinical tool and a measure of quality and can be used as an admission screening tool in all older patients with trauma.

We found that the number of high-risk medications pre-scribed was not altered during trauma admission at our insti-tution. This represents a potential area for improvement via medication review and optimization, as a lack of reduction in high-risk medication prescriptions represents a lost opportunity to prevent future trauma and other adverse

Table 2. Demographic and descriptive characteristics of older patients with major trauma

CharacteristicMean (95% CI)*

n = 319

Age, yr 76.0 (75.06–76.88)

Male sex, no. (%) 207 (64.9)

Injury Severity Score 22.3 (21.43–23.10)

Length of stay, d 16.8 (14.37–19.31)

30-day mortality, no. (%) 6 (1.9)

Mechanism of injury, no. (%)

Fall from level 129 (40.4)

Fall from height 75 (23.5)

Motor vehicle related 69 (21.6)

Other 46 (14.4)

No. of visits to operating room 0.52 (0.43–0.62)

No. of admission medications 4.36 (3.93–4.78)

No. of discharge medications 4.87 (4.42–5.33)

No. of preadmission Beers Criteria medications

1.16 (1.01–1.31)

No. of discharge Beers Criteria medications 1.13 (0.99–1.28)

CI = confidence interval. *Except where noted otherwise.


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events, such as prolonged length of stay and death. It may be that, currently, effective medication reviews are simply not occurring or that there is a cultural preference to defer changes in long-term medications to primary care providers. Prescribers may also not be aware of high-risk medications on the Beers Criteria list and that suitable non-Beers medi-cation alternatives exist. In addition, there are many medica-tions included in the Beers Criteria list, several with stipula-tions such as “avoid in patients with CrCL [creatinine clearance] < 60 mL/min,” perhaps making it cumbersome as a clinical tool for medication review.22 Finally, some medica-tions may be necessary for certain patients and cannot be substituted or stopped. A medication review conducted by a dedicated trauma team member may mitigate these issues.

The importance of this failure to reduce high-risk med-ications is reinforced by the results of our multivariable analysis, which showed that each additional Beers Criteria medication prescribed before admission increased length of stay by 10% more than each additional medication of any type. This increase likely represents substantial resource use. Furthermore, Beers Criteria medications were statistically signi�cant predictors of 30-day mortality. These results support attention to medication review in this high-risk population.

We found that a third of older patients were prescribed a central nervous system medication. In addition, central nervous system medications were predictors of increased length of stay. Although this may be due to their ubiquity, other possible explanations include the underlying condi-tion warranting these medications or their anticholinergic and sedating effects. Moreover, many of these medications are known to induce hypotension, which may predispose patients to injury.22 On the other hand, several classes of medications were prescribed relatively infrequently in our study. The less frequent nature of medications such as antithrombotics, anticoagulants and endocrine medications may be a result of preinjury clinical indications, provider recognition of the risks of these medications or the sample size used in our study. Alternatively, these less frequently prescribed medications may increase the preadmission mortality rate, thereby excluding these patients from the hospital admission sample.

The Beers Criteria represent a guideline for medication review and a valuable process indicator for quality care in older people. Process indicators for quality care are a use-ful metric of health care system performance that allow for early improvement signal detection.28 The American Col-lege of Surgeons’ Trauma Quality Improvement Program informs trauma centres of their performance against national standards using quality indicators, and the use of the Beers Criteria is congruent with the college’s current approach of improving both trauma management and trauma prevention.

Results from this retrospective study can be used to design a medication review process based on the Beers

Criteria as a quality indicator for older patients with trauma at our institution. A dedicated trauma team member, such as a pharmacist or geriatrician, could facil-itate formal medication review and communication with primary care providers to promote a reduction in poly-pharmacy and high-risk medication use. Moreover, bringing attention to the need to scrutinize patient med-ications may strengthen the collective relationship between providers of trauma care and primary care net-works and thereby lead to further injury prevention mea-sures. The Beers Criteria is an applicable framework, as it differentiates between medications to avoid com-pletely, those to avoid in certain subpopulations and those to review and consider changing. It also provides a rationale and references for the recommendations. A future study prospectively assessing the changes in high-risk medications during the hospital stay and the effect on outcomes with implementation of the Beers Criteria could further illuminate the bene�t of including medica-tion review with the Beers Criteria as a process indicator for quality improvement.

Limitations

Several limitations of our work merit consideration. First, this study was a retrospective application of the Beers Cri-teria, and the pertaining issues surrounding bias and valid-ity apply. Further limitations include the lack of a compre-hensive analysis surrounding patient comorbidities, as these details were unavailable in the trauma registry. Comorbidities have been shown to be associated with worsened outcomes in geriatric trauma.29–33 It was not pos-sible to elucidate whether the increased length of stay and risk of mortality were due to the effect of the medications or whether Beers Criteria medications were a proxy mea-sure for comorbid conditions that resulted in poorer out-comes. To mitigate the effect of this omitted variable, we adjusted the impact of the independent variable of interest using comparable statistical models, including the effect of total medications compared to the effect of Beers Criteria medications alone. Further omitted variable bias may have been associated with over-the-counter medications, which were not included owing to data unavailability.

CONCLUSION

Providing high-quality care to older patients who have experienced trauma is important, and one way to attain this is by developing relevant and practical process indica-tors for quality improvement. Using the Beers Criteria as a quality process indicator in older patients with trauma may provide dedicated interdisciplinary trauma teams with an opportunity to audit preinjury medications and stop potentially harmful medications in this vulnerable population.


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Acknowledgements: The data analyzed in this study were provided by the Alberta Trauma Registry.

Af�liation: From the Department of Surgery, University of Alberta, Edmonton, Alta.

Competing interests: V. Fawcett and S. Widder are CJS associate edi-tors, but they were not involved in the review or the decision to publish this paper. None declared by E. Lester, M. Dykstra, C. Grant and B. Tsang.

Contributors: E. Lester, V. Fawcett, B. Tsang and S. Widder designed the study. M. Dykstra and C. Grant acquired the data, which E. Lester, M. Dykstra and V. Fawcett analyzed. E. Lester, M. Dykstra, C. Grant and S. Widder wrote the article, which all authors reviewed. All authors approved the �nal version to be published and can certify that no other individuals not listed as authors have made substantial contributions to the paper.

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24. Patterson SM, Hughes C, Kerse N, et al. Interventions to improve the appropriate use of polypharmacy in older people: a Cochrane sys-tematic review. Cochrane Database Syst Rev 2012;(5):CD008165.

25. McGwin G Jr, May AK, Melton SM, et al. Recurrent trauma in elderly patients. Arch Surg 2001;136:197-203.

26. Howard M, Dolovich L, Kaczorowski J, et al. Prescribing of potentially inappropriate medications to elderly people. Fam Pract 2004;21: 244-7.

27. Baker SP, O’Neill B, Haddon W Jr, et al. The Injury Severity Score: a method for describing patients with multiple injuries and evaluating emergency care. J Trauma 1974;14:187-96.

28. Smith PC, Mossialos E, Papanicolas I, et al. Performance measurement for health system improvement: experiences, challenges and prospects. Cam-bridge (UK): Cambridge University Press; 2009.

29. Labib N, Nouh T, Winocour S, et al. Severely injured geriatric pop-ulation: morbidity, mortality, and risk factors. J Trauma 2011;71: 1908-14.

30. Shoko T, Shiraishi A, Kaji M, et al. Effect of pre-existing medical conditions on in-hospital mortality: analysis of 20,257 trauma patients in Japan. J Am Coll Surg 2010;211:338-46.

31. van der Sluis CK, Timmer HW, Eisma WH, et al. Outcome in elderly injured patients: injury severity versus host factors. Injury 1997;28:588-92.

32. Milzman DP, Boulanger BR, Rodriguez A, et al. Pre-existing disease in trauma patients: a predictor of fate independent of age and injury severity score. J Trauma 1992;32:236-43, discussion 43-4.

33. Grossman MD, Miller D, Scaff DW, et al. When is an elder old? Effect of preexisting conditions on mortality in geriatric trauma. J Trauma 2002;52:242-6.




Prevalence of obstructive sleep apnea in male patients with surgically treated maxillary and zygomatic fractures

Background: Midface fractures can cause airway obstruction and breathing distur-bances. The purpose of the present study was to determine the prevalence of undiag-nosed obstructive sleep apnea (OSA) among patients with surgically treated maxillary and zygomatic fractures.

Methods: We retrospectively analyzed the medical records of 44 patients who had undergone surgical treatment of maxillary or zygomatic fractures between Jan. 1, 2003, and Dec. 31, 2013 at a single centre. All participants underwent polygraphy testing and were asked to complete the STOP (snoring, tiredness, observed apnea and high blood pressure) questionnaire, Nasal Obstruction Symptom Evaluation (NOSE) scale and Epworth Sleepiness Scale.

Results: There were 27 participants (61%) with maxillary fracture and 17 (39%) with zygomatic fracture. Obstructive sleep apnea was diagnosed in 24 (54%) of the 44 par-ticipants, of whom 15 (62%) had maxillary fractures and 9 (38%) had zygomatic frac-tures. Participants with OSA had a mean Apnea–Hypopnea Index (AHI) of 15.5 (stan-dard deviation [SD] 9.7) events/h, compared to 2.4 (SD 1.5) events/h for those without OSA (p < 0.001). Of the 30 participants with nose obstruction, 18 (60%) had an AHI of 5 or greater.

Conclusion: The results suggest that the prevalence of OSA was higher in surgical patients with midface fractures, independent of the type of fracture, than in the gen-eral population. The NOSE scale results showed signi�cant correlation with the pres-ence of OSA.

Contexte : Les fractures affectant la portion médiane du visage peuvent provoquer une obstruction des voies respiratoires et gêner la respiration. La présente étude avait pour but de déterminer la prévalence de l’apnée obstructive du sommeil (AOS) non diagnostiquée chez des patients ayant été traités chirurgicalement pour des fractures du maxillaire et de l’os zygomatique.

Méthodes : Nous avons analysé rétrospectivement les dossiers médicaux de 44 patients ayant subi un traitement chirurgical pour une fracture du maxillaire ou de l’os zygomatique entre le 1er janvier 2003 et le 31 décembre 2013 dans un seul éta-blissement. Tous les participants ont subi un test polygraphique et ont été invités à répondre aux questionnaires STOP (snoring, tiredness, observed apnea et high blood pressure), NOSE (Nasal Obstruction Symptom Evaluation), de même qu’à l’échelle de somnolence d’Epworth.

Résultats : Vingt-sept participants (61 %) avaient subi une fracture du maxillaire et 17 (39 %) de l’os zygomatique. L’AOS a été diagnostiquée chez 24 participants sur 44 (54 %), dont 15 (62 %) avaient subi une fracture du maxillaire et 9 (38 %) une frac-ture de l’os zygomatique. Les participants qui présentaient une AOS avaient un indice d’apnée-hypopnée (IAH) moyen de 15,5 (écart-type [É.-T.] 9,7) événements/h, contre 2,4 (É.-T. 1,5) événement/h pour les participants indemnes d’ASO (p < 0,001). Parmi les 30 participants qui avaient une obstruction nasale, 18 (60 %) avaient un IAH de 5 ou plus.

Conclusion : Ces résultats donnent à penser que la prévalence de l’AOS était plus élevée chez les patients opérés pour une fracture affectant la portion médiane du visage (indépendamment du type de fracture) que dans la population générale. Les résultats au questionnaire NOSE ont montré une corrélation signi�cative avec la pré-sence d’AOS.

Slaven Lupi-Ferandin, MD Tea Galic, DMD, PhD Natalija Ivkovic, MSc Renata Pecotic, MD, PhD Zoran Dogas, MD, PhD


Correspondence to: Z. Dogas Department of Neuroscience School of Medicine University of Split Soltanska 2, 21000 Split, Croatia [email protected]

DOI: 10.1503/cjs.002818

sleep-lupi.indd 105 2019-03-19 1:19 PM

RECHERCHE


M axillofacial fractures may result from a variety of injury types. The diagnosis and treatment of these injuries frequently require a multidisci-

plinary team approach. Zygomatic fractures are the second most common facial fractures, after nasal frac-tures.1 Mandibular and zygomatic bone fractures account for 80% of all facial bone fractures, and maxillary fractures account for 5%.1 The outcomes of maxillofacial trauma are of great signi�cance, both functionally and aesthet-ically.1,2 The best aesthetic and functional outcomes are obtained with early rigid �xation with screws and plates.2,3 Functionally, skeletal factors, which are usually not readily apparent on physical examination, may be important determinants of upper-airway patency during wakefulness but also during sleep.4,5

Among sleep-related breathing disorders, according to studies in the general population,6 obstructive sleep apnea (OSA) has emerged as a highly prevalent breathing disor-der, affecting 3%–7% of adult men and 2%–5% of adult women.7 Obstructive sleep apnea arises from a combina-tion of pathophysiological and anatomic factors, resulting in narrowing of the upper airway. The exact levels of obstruction may vary from one person to another. Invasive and noninvasive methods can be used to identify and evalu ate the level of obstruction, but some (e.g., magnetic resonance imaging and �broscopy) are too expensive and too invasive to be used in �eld surveys. Therefore, the clas-sical sleep questionnaires, anthropometric measurements and simple nose–throat examinations, which are readily accepted by patients, are useful for the identi�cation of those at increased risk for OSA.8

Patients who have undergone surgical treatment of maxillary or zygomatic fractures may have upper airway abnormalities related to the fracture and to the surgical procedure. Regular screening for OSA in these patients by means of standard questionnaires such as the STOP (snor-ing, tiredness, observed apnea and high blood pressure) questionnaire, the STOP-Bang questionnaire and the recently developed Nasal Obstruction Symptom Evalua-tion (NOSE) scale9–12 may be of additional value. The aim of the present study was to determine the prevalence of undiagnosed OSA among patients with surgically treated maxillary and zygomatic fractures. Also, we wanted to explore whether clinical data and NOSE scale scores can be used to identify those at increased risk for OSA in this population.

METHODS

Setting and participants

In this retrospective cohort study, we reviewed the medical records of male patients aged 18 years or more who had undergone surgical treatment of maxillary or zygomatic fractures between Jan. 1, 2003, and Dec. 31, 2013, at the

Department of Maxillofacial and Oral Surgery, University Hospital Center Split, Split, Croatia. Surgical treatment consisted of open reduction and rigid plate and screw �xa-tion. We considered the healing process to be complete 6 months postoperatively. Maxillary fractures were classi-�ed according to Le Fort patterns. Dentoalveolar fractures of the upper jaw were excluded. Zygomatic fractures were de�ned as zygomatic complex fractures (zygomaticomaxil-lary fractures). Patients with isolated fractures of the zygo-matic arch were excluded. Other exclusion criteria included postoperative malocclusion (less than ideal bone reposition achieved) and eventual rhinoseptoplasty after the initial surgical treatment. Patients who had concomi-tant neurosurgical trauma were also excluded.

Patients who met the diagnostic and treatment criteria were contacted by letter or telephone, and those willing to participate were recruited for the study.

This study was approved by the Ethics Committee of the University of Split School of Medicine and the Univer-sity Hospital Center Split and was undertaken in agree-ment with the principles of the Declaration of Helsinki. All participants signed written informed consent before participation.

Sleep assessment

All participants underwent unattended whole-night poly-graphy testing. The device used in this study to identify participants with OSA was the Embletta portable diagnos-tic system (Medcare). This system is highly sensitive and speci�c in quantifying the Apnea–Hypopnea Index (AHI) and has been validated as a screening and diagnostic tool for OSA.13 We defined apnea as complete cessation of respiratory airflow for a minimum of 10 seconds and hypopnea as a decrease in air�ow by more than 50% from baseline for at least 10 seconds, combined with a reduction in hemoglobin oxygen saturation of at least 3%.14–16 We defined the AHI as the average number of apneic and hypopneic events per hour of sleep, and we calculated the oxygen desaturation index as the number of decreases in arterial oxygen saturation of 3% or more per hour of sleep.14–16 We used the AHI to quantify the severity of OSA as follows: AHI 5–14.9: mild OSA; 15 ≤ AHI ≤ 30: moderate OSA; and AHI > 30: severe OSA.

All data were manually scored and evaluated in accor-dance with published guidelines of the American Academy of Sleep Medicine and the European Sleep Research Soci-ety14–16 by the same certi�ed sleep physician, who was blind to the participants’ involvement in the study. The data were analyzed if the total recorded time was 4 hours or longer.

Questionnaires

Demographic and anthropometric characteristics (e.g., age, body mass index, neck circumference) were


RESEARCH


collected at the Split Sleep Medicine Center before the whole-night sleep assessment. Participants were asked to complete the STOP questionnaire, NOSE scale and Epworth Sleepiness Scale (ESS). The STOP question-naire is 4-item self-reported questionnaire that has been shown to be a concise, easy-to-use screening tool for iden-tifying patients who are at increased risk for OSA (≥ 2 positive answers).9 The NOSE scale is a validated disease-specific instrument designed to measure nasal obstruction, commonly used in otolaryngology practice to provide an objective measure of nasal obstruction.10,12 It consists of 5 self-rated items, each scored from 0 to 4.10 The NOSE scale score represents the sum of the scores for the responses to the 5 individual items and ranges from 0 (no nasal obstruction) to 100 (worst possible prob-lems caused by nasal obstruction).10 The ESS is a self-administered questionnaire used to evaluate the level of daytime sleepiness.17 Patients are asked to rate their chance of falling asleep during 8 routine daytime situa-tions on a scale of 0–3. The �nal score is the total score for the 8 items and ranges from 0 to 24, with the cut-off value of 9 suggesting the presence of excessive daytime sleepiness.18

The Croatian versions of the STOP questionnaire and ESS have been validated.18


We performed statistical analyses using MedCalc for Win-dows version 11.5.1.0 (MedCalc Software). Continuous data were presented as mean and standard deviation (SD), and categorical variables were presented as whole numbers and proportions. We determined differences in variables between groups using the Mann–Whitney test for independent sam-ples and the χ2 test. We evaluated correlations between study variables and AHI using Spearman’s rank correlation coef�cients. Statistical signi�cance was set at p < 0.05.

RESULTS

The study population consisted of 44 participants, 27 (61%) with maxillary fractures and 17 (39%) with zygo-matic complex fractures. None of the participants had mal-occlusion or rhinoseptoplasty in the postoperative period. Of the 27 participants with maxillary fractures, 8 had Le Fort I fractures, 8 had Le Fort II fractures, 4 had Le Fort III fractures, and 7 had a combination of 2 differ-ent patterns. The baseline demographic and anthropomet-ric characteristics of the participants are presented in Table 1. There were no statistically signi�cant differences between those with maxillary fractures and those with zygomatic complex fractures.

All participants had at least 4 hours of recorded time on polygraphy testing. The sleep characteristics of the partici-pants are presented in Table 2. There were no statistically

signi�cant differences between those with maxillary frac-tures and those with zygomatic fractures. In addition, the questionnaire scores did not differ signi�cantly between the 2 groups (Table 3). Therefore, we decided to unify the 2 groups of participants into 1 group for further analyses.

Obstructive sleep apnea was diagnosed in 24 partici-pants (54%). There were 14 participants (32%) with mild OSA, 7 (16%) with moderate OSA and 3 (7%) with severe OSA. Of the 24, 15 (62%) had maxillary fractures and 9 (38%) had zygomatic fractures. The participants with OSA differed signi�cantly from those without OSA in mean age (49.0 [SD 15.4] yr v. 33.3 [SD 15.1] yr, p = 0.001) and all sleep characteristics (Table 4). The par-ticipants with OSA had a mean AHI of 15.5 (SD 9.7) events/h, compared to 2.4 (SD 1.5) events/h for those without OSA (p < 0.001). The corresponding oxygen desaturation index values were 11.6 (SD 8.9) events/h and 1.9 (SD 1.2) events/h (p = 0.002).

Table 1. Baseline demographic and anthropometric characteristics of participants with maxillary or zygomatic complex fractures

Characteristic

Group; mean ± SD

p valueMaxillary n = 27

Zygomatic n = 17

Age, yr 41.8 ± 17.5 41.9 ± 17.1 0.95

Height, cm 184.0 ± 7.0 182.0 ± 6.0 0.3

Weight, kg 85.5 ± 8.9 85.0 ± 12.8 0.7

Body mass index 25.2 ± 3.9 25.9 ± 2.1 0.3

Neck circumference, cm 39.5 ± 3.3 39.4 ± 2.2 0.7

SD = standard deviation.

Table 2. Sleep characteristics of the 2 groups

Characteristic

Group; mean ± SD

p valueMaxillary Zygomatic

Apnea–Hypopnea Index (events/h)

10.8 ± 11.2 7.6 ± 6.6 0.8

Arterial oxygen saturation, % 95.3 ± 2.0 95.6 ± 1.3 0.9

Minimum arterial oxygen saturation, %

88.2 ± 6.1 89.1 ± 4.4 0.9

Oxygen desaturation index, % 9.2 ± 9.2 3.2 ± 1.8 0.2

Snoring time, min 261.2 ± 99.8 193.6 ± 58.9 0.2

SD = standard deviation.

Table 3. Questionnaire results for the 2 groups

Questionnaire Maxillary Zygomatic p value

Positive result† on STOP questionnaire, no. (%)

5 (18) 4 (24) 0.7*

Epworth Sleepiness Scale score, mean ± SD

4.3 ± 3.8 5.4 ± 3.2 0.2

NOSE scale score, mean ± SD

17.8 ± 21.9 29.1 ± 28.8 0.3

NOSE = Nasal Obstruction Symptom Evaluation; SD = standard deviation; STOP = snoring, tiredness, observed apnea, high blood pressure. *χ2 test (positive v. negative result). †Two or more positive answers.


RECHERCHE


Thirty-�ve participants (80%) had no risk for OSA, as estimated by the STOP questionnaire (Table 4). Of the 35, 19 did not have OSA according to the sleep study data. According to the STOP questionnaire, 9 participants (20%) had increased risk for OSA, and OSA was con�rmed in 8 (89%) of them during polygraphy testing.

The NOSE scale results showed that 30 participants (68%) had nose obstruction and 14 (32%) did not (Table 5). Of the 30 participants with nose obstruction, 18 (60%) had an AHI of 5 or greater. Only 6 (25%) of the 24 participants with OSA had no nose obstruction accord-ing to the NOSE scale. Signi�cant correlations were found between the AHI and the NOSE scale score (r = 0.323, p = 0.04) when we performed analysis on the overall study population (Table 6).

Five of the participants with OSA experienced excessive daytime sleepiness, with a mean ESS score of 11.6 (SD 1.8). Of the 5, 2 had a STOP score of 2 or greater, and 4 had nasal obstruction, estimated by the NOSE scale. Only 1 participant in the non-OSA group had excessive daytime sleepiness according to the ESS score.

DISCUSSION

In this retrospective analysis of the prevalence of OSA, nasal obstruction and daytime sleepiness in a sample of 44 participants with surgically treated maxillary or zygo-matic fractures, the prevalence of OSA was higher in our study population than in general population.7,19,20

The use of questionnaires as a screening tool for OSA proved to be a valuable, but special attention should be paid when using the STOP questionnaire in specific popu lations, since in our participants with midface injury it was unable to identify risk in a substantial proportion of

those with confirmed OSA. The NOSE scale showed good correlation with the presence of OSA: 75% of par-ticipants with OSA reported nose obstruction according to this scale. Nasal obstruction has been identi�ed as a risk factor for OSA and is a common sign in patients with the disorder.12,21,22

Obstructive sleep apnea is prevalent in surgical popula-tions and is considered to be an independent risk factor for perioperative complications in noncardiac opera-tions.23 Surgical patients with OSA are vulnerable to seda-tion, anesthesia and analgesia. The perioperative risk of patients with OSA may be reduced by appropriate screen-ing to detect undiagnosed OSA and to plan a speci�c peri-operative management plan for those with OSA.24 There-fore, it is important to identify patients at high risk for OSA preoperatively, which is not always easy. The short interval between the preoperative clinic visit and sched-uled surgery date, lack of patient willingness to undergo preoperative polysomnography or polygraphy testing and potentially long wait times for a sleep clinic appointment may hinder the diagnosis of OSA before surgery. By incorporating the STOP or STOP-Bang questionnaire and the NOSE scale into preoperative clinic practice, sur-gical patients at risk for OSA can be better detected, which can help during their perioperative and postopera-tive treatment.25

The STOP questionnaire has been validated in sur-gical patients at preoperative clinics.9 In our study popu-lation, the STOP questionnaire proved to have good pre-dictive value to detect participants with OSA: the disorder was confirmed during polygraphy testing in 89% of patients who had a STOP score of 2 or greater.

Table 4. Comparison between participants with and without obstructive sleep apnea

Variable

Group; mean ± SD*

p valueOSA

n = 24No OSA n = 20

Age, yr 49.0 ± 15.4 33.3 ± 15.1 0.001

Body mass index 26.2 ± 3.8 23.2 ± 2.9 0.09

Positive result on STOP questionnaire, no. (%) 8 (33) 1 (5) 0.05†

Epworth Sleepiness Scale score 5.4 ± 3.9 3.9 ± 3.1 0.2

NOSE scale score 29.2 ± 28.5 13.8 ± 17.6 0.04

Apnea–Hypopnea Index (events/h) 15.5 ± 9.7 2.4 ± 1.5 < 0.001

Oxygen desaturation index, % 11.6 ± 8.9 1.9 ± 1.2 0.002

Arterial oxygen saturation, % 94.9 ± 2.0 96.0 ± 1.2 0.04

Minimum arterial oxygen saturation, % 86.3 ± 6.0 91.1 ± 3.2 0.003

NOSE = Nasal Obstruction Symptom Evaluation; OSA = obstructive sleep apnea; SD = standard deviation; STOP = snoring, tiredness, observed apnea, high blood pressure. *Except where noted otherwise. †χ2 test (positive v. negative result).

Table 5. Nasal Obstruction Symptom Evaluation scale scores of participants with and without obstructive sleep apnea

Score

Group; no. (%) of participants

p valueOSA No OSA

0 (no obstruction) 6 (25) 8 (40) 0.3*

1–25 (mild obstruction) 8 (33) 8 (40)

26–50 (moderate obstruction) 3 (12) 2 (10)

> 50 (serious obstruction) 7 (29) 2 (10)

OSA = obstructive sleep apnea. *χ2 test (obstruction v. no obstruction).

Table 6. Correlations between study variables and the Apnea–Hypopnea Index in the overall study population

VariableSpearman’s coefficient ρ p value

Age 0.554 < 0.001

Body mass index 0.297 0.06

STOP questionnaire score 0.501 0.001

Epworth Sleepiness Scale score 0.146 0.4

NOSE scale score 0.323 0.04

NOSE = Nasal Obstruction Symptom Evaluation; STOP = snoring, tiredness, observed apnea, high blood pressure.


RESEARCH


In the study by Chung and colleagues,9 the STOP ques-tionnaire administered preoperatively to surgical patients had a sensitivity of 65.6%, 74.3% and 79.5% with AHI cut-off values of greater than 5 events/h, greater than 15 events/h and greater than 30 events/h, respectively. In addition, an earlier study from our group showed a high level of sensitivity and speci�city of the STOP question-naire in patients referred to the Split Sleep Medicine Center.18 The probability that the STOP questionnaire correctly predicted an AHI greater than 5 events/h was 84%, with a sensitivity of 96% and a speci�city of 83% at a cut-off point of 2 events/h for determining the risk of OSA. Our unpublished research on around 4000 partici-pants in the general population residing in the same region as those in the current study showed the risk of OSA to be 15.6% in participants aged 31–40 and 25.8% in those aged 41–50, substantially lower than the propor-tion of participants with con�rmed OSA in the present study, 54% (24/44).

The ESS scores of both participants with OSA and those without OSA in the current study were lower than those in the study by Pecotic and colleagues.18 This means that people with surgically treated midface injury either experienced less excessive daytime sleepiness or ignored it and reported fewer signs.

Limitations

We did not have any data on nose obstruction, snoring or risk of OSA in our study population before the midface trauma. However, we believe that our results support screening for OSA with the STOP questionnaire, STOP-Bang questionnaire and NOSE scale, as well as use of the ESS for assessing excessive daytime somnolence, in sur-gical populations to avoid perioperative and postoperative complications. However, caution must be used when interpreting our results owing to possibility of a large pro-portion of false-negative results in this speci�c population compared to the general population referred to sleep centres.

CONCLUSION

The NOSE scale showed signi�cant correlation with the presence of OSA. Therefore, we recommend its use in common otolaryngology and surgical practice. Specific questionnaires such as the STOP questionnaire used in this study may underestimate the risk of OSA in patients with surgically treated midface fractures, and the ESS may underestimate excessive daytime somnolence. In general, we recommend use of all those questionnaires, but special attention should be paid when they are used in speci�c sur-gical populations, such as patients with surgically treated midface injuries. Finally, the prevalence of OSA was higher in surgical patients with midface fractures regardless of

the type of fracture, maxillary or zygomatic, than in the general population.

Af�liations: From the Department of Maxillofacial and Oral Surgery, University Hospital Center Split, Split, Croatia (Lupi-Ferandin); Den-tal Medicine Program, University of Split School of Medicine, Split, Croatia (Galic); the Split Sleep Medicine Center, University of Split School of Medicine, Split, Croatia (Ivkovic, Pecotic, Dogas); and the Department of Neuroscience, University of Split School of Medicine, Split, Croatia (Pecotic, Dogas).


Contributors: S. Lupi-Ferandin and Z. Dogas designed the study. S. Lupi-Ferandin, N. Ivkovic and R. Pecotic acquired the data, which S. Lupi-Ferandin, T. Galic, R. Pecotic and Z. Dogas analyzed. S. Lupi-Ferandin, T. Galic, R. Pecotic and Z. Dogas wrote the article, which all authors reviewed and approved for publication.

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10. Pharmacological management of postoperative ileus Zeinali et al. Can J Surg 2009;52:153–7 *Based on page views on PubMed Central of research, reviews, commentaries and discussions in surgery. Updated Mar. 12, 2019.




Surgical site infection following abdominal surgery: a prospective cohort study

Background: Surgical site infection (SSI) is one of the most common complications of abdominal surgery and is associated with substantial discomfort, morbidity and cost. The goal of this study was to describe the incidence, bacteriology and risk factors associated with SSI in patients undergoing abdominal surgery.

Methods: In this prospective cohort study, all patients aged 14 years or more under-going abdominal surgery between Feb. 1 and July 31, 2016, at a single large academic hospital were included. Patients undergoing vascular, gynecological, urological or plas-tic procedures were excluded. Patients were followed prospectively for 30 days. Wound assessment was done with the Centers for Disease Control and Prevention de�nition of SSI. We performed multivariate analysis to identify factors associated with SSI.

Results: A total of 337 patients were included. The overall incidence of SSI was 16.3% (55/337); 5 patients (9%) had deep infections, and 25 (45%) had combined super�cial and deep infections. The incidence of SSI in open versus laparoscopic operations was 35% versus 4% (p < 0.001). The bacteria most commonly isolated were extended-spectrum β-lactamase-producing Escherichia coli, followed by Enterococ-cus species. Only 23% of cultured bacteria were sensitive to the prophylactic antibiotic given preoperatively. The independent predictors of SSI were open surgical approach, emergency operation, longed operation duration and male sex.

Conclusion: Potentially modi�able independent risk factors for SSI after abdominal sur-gery including open surgical approach, contaminated wound class and emergency surgery should be addressed systematically. We recommend tailoring the antibiotic prophylactic regimen to target the commonly isolated organisms in patients at higher risk for SSI.

Contexte : L’infection de plaie opératoire (IPO) est l’une des plus fréquentes compli-cations de la chirurgie abdominale et elle est associée à un inconfort, une morbidité et des coûts substantiels. L’objectif de cette étude était de décrire l’incidence, les don-nées bactériologiques et les facteurs de risque associés à l’IPO chez les patients soumis à une chirurgie abdominale.

Méthodes : Dans cette étude de cohorte, tous les patients de 14 ans ou plus soumis à une chirurgie abdominale entre le 1er février et le 31 juillet 2016 dans un seul grand établissement hospitalier universitaire ont été inclus. Les chirurgies vasculaires, gyné-cologiques, urologiques ou plastiques ont été exclues. Les patients ont été suivis de façon prospective pendant 30 jours. L’évaluation des plaies a été effectuée à partir de la dé�nition de l’IPO des Centers for Disease Control and Prevention. Nous avons procédé à une analyse multivariée a�n d’identi�er les facteurs associés à l’IPO.

Résultats : En tout, 337 patients ont été inclus. L’incidence globale des IPO a été de 16,3 % (55/337); 5 patients (9 %) ont présenté des infections profondes, et 25 (45 %) ont présenté des infections super�cielles et profondes. L’incidence des IPO lors d’interventions ouvertes c. laparoscopiques a été de 35 % c. 4 % (p < 0,001). Les bac-téries les plus souvent isolées étaient Escherichia coli productrices de β-lactamases à spectre élargi, suivies du genre Enterococcus. Seulement 23 % des bactéries cultivées se sont révélées sensibles à l’antibioprophylaxie administrée avant l’intervention. Les prédicteurs indépendants d’une IPO étaient l’approche chirurgicale ouverte, le carac-tère urgent de l’intervention, sa durée prolongée et le fait d’être de sexe masculin.

Conclusion : Dans le contexte de la chirurgie abdominale, les facteurs de risque d’IPO indépendants potentiellement modi�ables, incluant l’approche ouverte, la clas-sification de la contamination de la plaie et le caractère urgent de la chirurgie, méritent d’être systématiquement pris en compte et corrigés. Nous recommandons une antibioprophylaxie adaptée pour cibler les agents souvent isolés chez les patients exposés à un risque plus élevé d’IPO.

Aroub Alkaaki, MD* Osman O. Al-Radi, MD, MSc* Ahmad Khoja, MD Anfal Alnawawi, MD Abrar Alnawawi, MD Ashraf Maghrabi, MD Abdulmalik Altaf, MD Murad Aljiffry, MD, MSc

*Contributed equally as �rst authors

An earlier version of this work was presented at the Canadian Surgery Forum 2017, Sept. 14–16, 2017, Victoria, BC.


Correspondence to: M. Aljiffry Department of Surgery Faculty of Medicine King Abdulaziz University PO Box 80215 Jeddah 21589 Saudi Arabia [email protected]

DOI: 10.1503/cjs.004818

site-alkaaki.indd 111 2019-03-19 11:14 AM

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S urgical site infection (SSI) is de�ned by the Centers for Disease Control and Prevention as a wound infection that occurs within 30 days of an operative

procedure or within a year if an implant is left in place and the infection is thought to be secondary to surgery.1 It is one of the most common health-care–associated infections, occurring following 1%–3% of all surgical procedures.2 The rates of SSI are much higher with abdominal surgery than with other types of surgery, with several prospective studies indicating an incidence of 15%–25% depending on the level of contamination.2–5 Surgical site infection is pre-ventable and is associated with high morbidity and mortal-ity. In addition to the devastating impact on the patient’s course of treatment, it is associated with prolonged length of hospital stay and higher costs.6–8 Numerous risk factors may contribute to the development of SSI, with the most recognized factors being these incorporated in the Centers for Disease Control and Prevention and National Nosoco-mial Infections Surveillance System SSI risk index, includ-ing wound classi�cation, American Society of Anesthesiol-ogists (ASA) score and duration of the operation.9

The primary objectives of this study were to describe the incidence and risk factors associated with SSI in patients undergoing abdominal surgery. The secondary objectives were to study the microbiological pattern of SSI in our population and their antibiotic sensitivity, and to study the effect of SSI on postoperative length of stay.

METHODS

Study design and participants

We conducted a prospective cohort study at King Abdulaziz University Hospital in Jeddah, Saudi Arabia. The target population consisted of all patients aged 14 years or more undergoing abdominal surgery from Feb. 1 to July 31, 2016. Patients were identi�ed with the use of the operating room electronic registry. We excluded patients undergoing abdominal surgery for vascular, gynecological, urological or plastic indications. Patients who left the operating theatre with an open packed wound or with a vacuum-assisted dressing were also excluded. Ethics approval was obtained from the institutional review board.

Data collection

Variables included in the analysis were patient demo-graphic characteristics, preoperative risk factors (diabetes mellitus, immunosuppression, HIV infection, use of che-motherapy and steroid use), smoking status, body mass index, ASA classi�cation, and preoperative hemoglobin and albumin levels. Operative variables included operation performed, duration of surgery, use of prophylactic anti-biotics, wound contamination class, surgical approach (open v. laparoscopic), urgency of surgery and drain use.

Outcome measures studied in addition to wound infection included admission to the intensive care unit, length of stay, postoperative complications and death.

Patients were followed prospectively for 30 days in the ward, outpatient clinic or dressing clinic or through tele-phone interview. Wound assessment was done with the Centers for Disease Control and Prevention and National Healthcare Safety Network de�nition of SSI.1 Surgical site infection was classi�ed as super�cial (involving the skin and subcutaneous tissue only), deep (involving deeper soft tissues such as fascia and muscle layers) or organ space (involving any part of the anatomy that was opened or manipulated during surgery) (Table 1). The results of cul-ture and antimicrobial sensitivity were included.


Discrete variables were described as frequency and propor-tion, and continuous variables were described as mean and standard deviation or median and interquartile range. We used the Pearson and Wilcoxon univariable tests to guide the multivariable models for discrete and continuous vari-ables, respectively. We used multivariable logistic regres-sion models to identify preoperative and operative variables independently associated with SSI. Model �t was assured with bootstrap validation and calibration. Normality and linearity were tested, and appropriate transformation was incorporated as needed. We used the R statistical package (R Foundation for Statistical Computing) for the analyses.

RESULTS

In total, 337 patients were enrolled in the study, 193 females (57.3%) and 144 males (42.7%) with a mean age of 43.6 years. All patients completed the 30-day

Table 1. Surgical site infection classification according to the Centers for Disease Control and Prevention and National Healthcare Safety Network1

Superficial incisional SSI Deep incisional SSI

Occurs within 30 d Occurs within 30 d

Only skin and subcutaneous tissue

Deep soft tissues (fascial and muscle layers)

Patient has at least 1 of the following: a. Purulent drainage from incision b. Organisms identified from wound c. Superficial incision that is deliberately opened by surgeon d. Diagnosis of SSI by surgeon

Patient has at least 1 of the following: a. Purulent drainage from deep incision b. An incision that spontaneously dehisces or is deliberately opened or aspirated by surgeon, with or without culture c. Abscess or other evidence of infection that is detected on gross anatomic or histopathologic examination, or imaging

And patient has at least 1 of the following: pain or tenderness, localized swelling, erythema or heat

And patient has at least 1 of the following: fever (temperature > 38°C), localized pain or tenderness

SSI = surgical site infection.


RESEARCH


follow-up apart from those who died before 30 days (n = 4). The mean body mass index was 31. Seventy patients (20.8%) were diabetic, and 41 (12.2%) were smokers. Chemotherapy, systemic steroid therapy and other immunosuppressive medications were not common in this cohort (Table 2). Most patients (279 [82.8%]) had an ASA score less than 3. The mean operative time was 145.2 minutes. A total of 199 cases (59.0%) were per-formed laparoscopically. Most patients (257 [76.3%]) underwent elective surgery. The most frequent type of surgery was laparoscopic cholecystectomy (111 proced-ures [32.9%]), followed by hernia repair (67 [19.9%]) and

bariatric surgery (56 [16.6%]). Most operations (257 [76.3%]) were classi�ed as clean-contaminated.

The overall rate of SSI was 16.3% (55/337). The wound classi�cation is detailed in Table 2 for patients with and without SSI. The demographic and preoperative variables for the 2 groups are presented in Table 3. The median post-operative hospital stay was 2 days for patients without SSI, compared to 13 days for those with SSI (p < 0.001). Details of postoperative variables by SSI group are shown in Table 4. Of the 55 patients with SSI, 25 (45%) had super�-cial infections, 5 (9%) had deep infections, and 25 (45%) had combined super�cial and deep infections. Surgical site infection rates per procedure are illustrated in Figure 1. The highest SSI rate was detected in laparotomy wounds: SSI developed in 24 (69%) of the 35 patients in this group. None of the patients who underwent bariatric surgery expe-rienced SSI. The infection rate was 34.8% following open

Table 2. Incidence of surgical site infection by wound class following open and laparoscopic surgery

Wound class

Surgical approach; no. of patients (% with SSI)

Open Laparoscopic All

Clean 51 (5.9) 5 (0.0) 56 (5.4)

Clean-contaminated 69 (46.4) 188 (3.2) 257 (14.8)

Contaminated/dirty 18 (72.0) 6 (17.0) 24 (58.3)

All 138 (34.8) 199 (3.5) 337 (16.3)

SSI = surgical site infection.

Table 4. Demographic, operative and postoperative risk factor variables of patients with and without surgical site infection

Variable

Group; no. (%) of patients*

p valueNo SSI SSI

Operation < 0.001†

Laparoscopic cholecystectomy 106 (37.6) 5 (9)

Hernia repair 61 (21.6) 6 (11)

Gastric sleeve/bypass 56 (19.8) 0 (0)

Appendectomy 14 (5.0) 4 (7)

Laparotomy 11 (3.9) 24 (44)

Colectomy/abdominoperineal resection/low anterior resection

6 (2.1) 5 (9)

Hepatectomy/Whipple procedure

4 (1.4) 8 (14)

Other 24 (8.5) 3 (5)

Wound class < 0.001†

Clean 53 (18.8) 3 (5)

Clean-contaminated 219 (77.6) 38 (69)

Contaminated/dirty 10 (3.5) 14 (25)

Urgency < 0.001†

Emergent 50 (17.7) 30 (54)

Elective 232 (82.3) 25 (45)

Approach < 0.001†

Laparoscopic 192 (68.1) 7 (13)

Laparoscopic converted to open

3 (1.1) 3 (5)

Open midline 35 (12.4) 35 (64)

Open nonmidline 52 (18.4) 10 (18)

Length of operation, min, median (IQR)

83 (110–155) 184 (113–292) < 0.001†

Drains 78 (27.6) 27 (49) 0.007‡

Blood transfusion 14 (5.0) 22 (40) < 0.001†

Postoperative stay, d, median (IQR)

2 (1–3) 13 (6–25) < 0.001†

Intensive care unit admission 22 (7.8) 28 (51) < 0.001‡

Death within 30 d 2 (0.7) 2 (4) 0.07

IQR = interquartile range; SSI = surgical site infection. *Except where noted otherwise. †Pearson test. ‡Wilcoxon test.

Table 3. Demographic and preoperative risk factor variables of patients with and without surgical site infection

Variable

Group; no. (%) of patients*

p valueNo SSI n = 282

SSI n = 55

Age, yr, median (IQR) 40 (31–52) 54 (43–61) < 0.001†

Sex

Male 104 (36.9) 40 (73) < 0.001‡

Female 178 (63.1) 15 (27)

Body mass index, median (IQR)

29 (25–36) 27 (25–31) 0.02†

ASA score < 0.001‡

1 143 (50.7) 16 (29)

2 107 (37.9) 13 (24)

3 31 (11.0) 18 (33)

4 1 (0.4) 6 (11)

5 0 (0.0) 2 (4)

Diabetic 52 (18.4) 18 (33) 0.02‡

Smoking status 0.004‡

Smoker 36 (12.8) 5 (9)

Former smoker 8 (2.8) 5 (9)

Nonsmoker 216 (76.6) 34 (62)

Unknown 22 (7.8) 11 (20)

Steroid use 11 (3.9) 4 (7) 0.3‡

Chemotherapy 6 (2.1) 4 (7) 0.04‡

Preoperative hemoglobin level, g/L, median (IQR)

13 (12–14) 12 (11–14) 0.1†

Preoperative albumin level, g/L, median (IQR)

36 (33–38) 31 (26–35) < 0.001†

Benign disease 267 (94.7) 33 (60) < 0.001‡

ASA = American Society of Anesthesiologists; IQR = interquartile range; SSI = surgical site infection. *Except where noted otherwise. †Wilcoxon test. ‡Pearson test.


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operations, compared to 3.5% following laparoscopic pro-cedures (p < 0.001). Surgical site infection developed in 22 (61%) of the 36 patients with malignant disease, com-pared to 33 (11.0%) of the 300 patients with benign disease (p < 0.001). The 30-day mortality rate was 0.7% among noninfected patients and 3.6% among infected patients.

Generally, infected patients were older and more com-monly male (40 [73%] v. 104 [36.9%], p < 0.001) than non-infected patients. Almost half (26 [47%]) of infected patients had an ASA score greater than 2. Patients with a low albumin level were more prone to SSI (p < 0.001). Notably, more than half (30 [54%]) of infected patients underwent emergency surgery, and most (48 [87%]) had an open surgical approach. Just over half (28 [51%]) were admitted to the intensive care unit. Use of steroids, preop-erative hemoglobin level, prophylactic antibiotic therapy and death within 30 days were not associated with SSI on univariable analysis. Multivariable analysis revealed that patients who underwent open laparotomy were 6.5 times (95% con�dence interval [CI] 2.16–19.6) more likely to experience SSI than those who had laparoscopic procedures (Table 5). Surgical site infection was 4.8 times (95% CI 1.58–14.4) more likely to develop in patients who under-went emergent operations than those who underwent elec-tive procedures. Operation duration and male sex were also independent predictors of SSI, with odds ratios of 2.1 (95% CI 1.23–3.6) and 2.6 (95% CI 1.02–6.6), respectively.

The cost of surgery is mainly related to the duration of the hospital stay. We found that length of hospital stay was predicted to increase with open versus laparoscopic proce-dures, long operative times (a surrogate for complexity)

and the development of SSI (Fig. 2). A patient who under-went a laparoscopic operation less than 150 minutes in length and did not experience SSI was predicted to stay less than 2 days in hospital, whereas a patient who under-went an open operation that lasted 350 minutes in whom SSI developed was predicted to stay in hospital for 12 days.

Microbiological cultures were available for 50 patients with SSI. The most commonly implicated pathogens were gram-negative bacteria, namely Escherichia coli (26 patients [52%], of whom 16 had extended-spectrum β-lactamase-producing E. coli ), followed by gram-positive bacteria

Fig. 1. Surgical site infection (SSI) rates per operation.

Laparotomy

Laparoscopic cholecystectomy

Hernia repair

Other

Hepatectomy/Whipple procedure

Gastric sleeve/bypass

Colectomy/abdominoperineal resection/low anterior resection

Appendectomy

0 20 40 60

Frequency (proportion with SSI in black)

80 100

Table 5. Multivariate logistic regression analysis of factors associated with surgical site infection

Risk factor OR (95% CI)

Male sex 2.6 (1.02–6.6)

Length of operation (86 min v. 181 min [25th v. 75th percentile])

2.1 (1.23–3.6)

Urgency (emergent v. elective) 4.7 (1.58–14.4)

Approach (open v. laparoscopic) 6.5 (2.16–19.6)

Age 1.09 (0.57–2.1)

Body mass index 1.20 (0.68–2.1)

Smoking 0.55 (0.15–2.1)

Diabetic 1.52 (0.52–4.5)

ASA score 1.27 (0.75–2.2)

Blood transfusion 0.93 (0.28–3.2 )

Preoperative albumin level 0.91 (0.63–1.3)

Malignant disease 2.35 (0.65–8.5)

Preoperative antibiotic 2.32 (0.71–7.6)

Wound type (contaminated/dirty v. clean-contaminated)

1.59 (0.48–5.3)

ASA = American Society of Anesthesiologists; CI = confidence interval; OR = odds ratio.


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Fig. 3. Culture results for 50 patients.

% o

f cul

ture

s

Escherichia coli Gram-positive bacteria Klebsiella Other

Acinetobacter baumannii Pseudomonas Candida albicans None

60

50

40

30

20

10

0

Fig. 2. Effect of surgical site infection (SSI) and open v. laparoscopic approach on length of stay.

LaparoscopicOpen

Length of operation, min

10

50 100 150 200 250 300 350 50 100 150 200 250 300 350

SSISSI

No SSINo SSI

5

0

Pred

icte

d po

stop

erat

ive

stay

, d


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(19 patients [38%]) and a considerable number of Acineto-bacter baumannii and Pseudomonas (Fig. 3). Interestingly, only 23% of cultured bacteria were sensitive to the pro-phylactic antibiotic given preoperatively.

All of our patients were given the antibiotic prophylaxis according to our standard hospital protocols (Table 6). All patients undergoing elective procedures take an antiseptic bath with chlorhexidine gluconate solution the night before the surgery. When hair removal is considered necessary, clipping is carried out in the operating theatre. Topically administered chlorhexidine gluconate or iodine is used for skin preparation.

DISCUSSION

In this cohort, SSI developed in 55 patients (16.3%) under-going abdominal surgery, compatible with reported rates in the literature.2,10 However, our rate is slightly higher that those reported in studies done in Saudi Arabia, 12%11 and 10.5%.12 The variation is partially attributable to the higher number of complex oncological and emergency procedures performed in our tertiary teaching hospital.

Multivariable analysis identi�ed open surgical approach, emergency operation, length of the operation and male sex as independent predictors of SSI. Open surgical approach and emergency surgery were documented as risk factors for SSI in previous reports.13–15 We found that patients who had open surgery were 6.5 times more likely to get SSI than those who had laparoscopic surgery. Emergency surgery increased the risk of SSI �vefold compared to elec-tive surgery. The rate of SSI was signi�cantly higher in male patients. This �nding is not novel.16 Although there is no consensus regarding why male patients are predisposed to SSI, studies have shown that, in laparoscopic chole-cystectomy, male sex is a predictor of longer and more dif-�cult operations and has a higher rate of conversion.17–19 In addition, it is known that there are sex differences in skin colonization that may be associated with differences in skin thickness, sebum production and skin pH.20,21

Among the 3 components of the National Nosocomial Infections Surveillance System risk index, only the dura-tion of surgery was an independent predictor for SSI in our study. A patient who had an operation lasting longer than the 75th percentile (> 3 h in our cohort) had double the risk of SSI in contrast to an operation lasting less than the 25th percentile (86 min). This �nding is in keeping with previous reports in the literature.5,11 Longer operative time reflects the complexity of the surgery. It would also increase the wound susceptibility to infection by increasing the exposure to potential contamination and decreasing the tissue concentration of antibiotic.22 To overcome the decreased concentration of antibiotic that occurs with pro-longed operations, readministration of the antibiotics is recommended.23

The ASA score and wound class were not signi�cant predictors of SSI in the multivariable model in our study. Certain other known risk factors for SSI such as body mass index, diabetes and smoking were also not found to be sta-tistically signi�cant. Obesity was not found to be a risk fac-tor for SSI in our patients. This may be explained by the fact that obese patients had more laparoscopic bariatric operations, which have a low risk of infection.24 Patients with diabetes and smokers had higher rates of infection; however, the differences were not statistically signi�cant. This may have been due to the sample size and the speci�c case-mix at our institution.

In this study, the commonest organisms isolated from patients with SSI were gram-negative bacteria, namely extended-spectrum β-lactamase-producing E. coli. This �nding is contrary to those in studies that revealed more gram-positive bacteria such as Staphylococcus aureus and coagulase-negative staphylococci.2,25 However, other authors reported �ndings similar to ours, with more com-mon gram-negative bacteria isolated from the infected abdominal wounds in Al-Ahsa,11 Saudi Arabia2 and Tanza-nia.5 We also found that most of the pathogens were mul-tiresistant to the commonly prescribed prophylactic anti-

Table 6. King Abdulaziz University Hospital surgical prophylaxis guidelines

Operation Antibiotic prophylaxis Antibiotic

Upper gastrointestinal (esophagus, stomach and small bowel)

Recommended Cefazolin,* 1–2 g or Clindamycin,† 600 mg + gentamicin, 120 mg intravenously

Hepatobiliary

Bile duct, pancreatic, liver and open gallbladder

Recommended Cefazolin,* 1–2 g or Clindamycin,† 600 mg + gentamicin, 120 mg intravenously

Laparoscopic cholecystectomy

Not recommended; should be considered in patients at high risk

Cefazolin,* 1–2 g or Clindamycin,† 600 mg + gentamicin, 120 mg intravenously

Lower gastrointestinal (appendix and colorectum)

Highly recommended Cefoxitin, 1–2 g intravenously every 6 h preoperatively for 3 doses or Cefazolin,* 1–2 g intravenously + metronidazole, 500 mg intravenously

Hernia repair, groin (inguinal/femoral with or without mesh, laparoscopic or incisional

Not recommended —

*A dose of 2 g of cefazolin is recommended for patients weighing more than 80 kg. Dosing cefazolin for renal impairment: creatine clearance 35–54 mL/min: administer full dose in intervals of ≥ 8 hours; creatine clearance 11–34 mL/min: administer half of usual dose every 12 hours; creatine clearance ≤ 10 mL/min: administer half of usual dose every 18-–24 hours. †No dosage adjustment required for renal impairment.


RESEARCH


biotics. This might explain why we found a high rate of deep SSI. Further consideration regarding the selection of appropriate prophylactic antibiotics will be needed, espe-cially in patients at high risk.

Surgical site infection was associated with increased length of postoperative stay. Moreover, an increased operative time was associated with both higher SSI rate and prolonged post-operative stay. Although we did not evaluate the economic impact of SSI in our study, it is likely that longer postopera-tive stay due to SSI entails a higher cost of patient care.

CONCLUSION

The present study identi�ed several independent risk fac-tors for SSI following abdominal surgery that should be addressed systematically. We believe that these results will be helpful in updating the guidelines for preventing SSI in the region. Furthermore, we recommend tailoring the pro-phylactic antibiotic regimens to target the commonly iso-lated organisms, especially in the presence of independent risk factors.

Acknowledgements: The authors acknowledge the medical students who helped with data collection: Mohmd Alhamed, Faisal Idrees and Sundos Turostani.

Af�liations: From the Department of Surgery, Faculty of Medicine, King Abdulaziz University, Jeddah, Saudi Arabia.


Contributors: A. Alkaaki, O. Al-Radi, A. Khoja, Abrar Alnawawi, A. Maghrabi, A. Altaf and M. Aljiffry designed the study. A. Alkaaki, A. Khoja, Anfal Alnawawi, Abrar Alnawawi and M. Aljiffry acquired the data, which A. Alkaaki, O. Al-Radi, Anfal Alnawawi, Abrar Alnawawi, A. Altaf and M. Aljiffry analyzed. A. Alkaaki, O. Al-Radi, Anfal Alnawawi, A. Maghrabi and M. Aljiffry wrote the article, which all authors reviewed and approved for publication.

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24. Shabanzadeh DM, Sørensen LT. Laparoscopic surgery compared with open surgery decreases surgical site infection in obese patients: a systematic review and meta-analysis. Ann Surg 2012; 256:934-45.

25. Sugiura T, Uesaka K, Ohmagari N, et al. Risk factor of surgical site infection after pancreaticoduodenectomy. World J Surg 2012;36: 2888-94.




Does oxidized zirconium make a difference? Midterm cohort survivorship of symmetric posterior condyle posterior-stabilized total knee arthroplasty

Background: A novel symmetric posterior condylar design and option of a femoral component with an outer zirconium oxide layer were introduced to a successful asym-metric condylar total knee arthroplasty system in 2005. Given the paucity of informa-tion on this modi�ed design, we performed a study to determine its early to midterm survivorship and performance.

Methods: Patients who received the Genesis II Symmetric Posterior Condyle or Legion Primary total knee (cobalt–chromium [CoCr] or oxidized zirconium–niobium [OxZr]) (Smith & Nephew) implanted at the study centre between March 2007 and December 2013 were enrolled into a prospective database. We retrospectively reviewed the database and performed survival analysis using Kaplan–Meier techniques.

Results: There were 2178 patients (1359 women [62.4%]; mean age 64.6 yr; mean body mass index 35.0) with 2815 knee replacements available for analysis. Survival rates were 98.2% (95% con�dence interval [CI] 97.6%–98.7%) for failure for any rea-son at 2 years and 96.8% (95% CI 96.0%–97.7%) at 5 years. Age (hazard ratio [HR] 0.97, 95% CI 0.94–0.997) and female sex (HR 0.45, 95% CI 0.27–0.75) were protec-tive, whereas body mass index (HR 1.02, 95% CI 0.99–1.05) and OxZr implant (HR 1.11, 95% CI 0.57–2.18) did not in�uence survivorship. Oxford Knee Score values improved from a mean of 40.2 to 21.8 by 2 years (p < 0.001), with no difference between the CoCr and OxZr groups.

Conclusion: The symmetric posterior condylar posterior-stabilized knee offers excellent midterm survivorship. Implant bearing surface did not have an in�uence on survivorship to 5 years, and, thus, use of OxZr implants may not be justi�ed.

Contexte : Un nouveau modèle à condyles postérieurs symétriques et l’option d’un élément fémoral pourvu d’une couche d’oxyde de zirconium ont été introduits avec succès dans un système de prothèse totale du genou à condyles asymétriques en 2005. Compte tenu du manque de renseignements concernant ce modèle modi�é, nous avons procédé à une étude pour en déterminer la survie et le rendement à court et à moyen terme.

Méthodes : Les patients ayant reçu la prothèse à condyles postérieurs symétriques Genesis II ou une prothèse totale primaire Legion (au cobalt–chrome [CoCr] ou au zirconium–niobium oxydé [OxZr]) (Smith & Nephew) au centre de recherche entre mars 2007 et décembre 2013 ont été inscrits dans une base de données prospective. Nous avons procédé à une revue rétrospective de la base de données et effectué une analyse de survie à l’aide de techniques de Kaplan–Meier.

Résultats : On a dénombré 2178 patients (dont 1359 femmes [62,4 %]; âge moyen 64,6 ans; indice de masse corporelle moyen 35,0) totalisant 2815 arthroplasties du genou pour analyse. Les taux de survie ont été de 98,2 % (intervalle de con�ance [IC]à 95 % 97,6 %–98,7 %) pour le paramètre défaillance de toutes causes à 2 ans et de 96,8 % (IC à 95 % 96,0 %–97,7 %) à 5 ans. L’âge (risque relatif [RR] 0,97, IC à 95 % 0,94–0,997) et le fait d’être de sexe féminin (RR 0,45, IC à 95 % 0,27–0,75) ont con-féré un effet protecteur, tandis que l’indice de masse corporelle (RR 1,02, IC à 95 % 0,99–1,05) et l’implant OxZr (RR 1,11, IC à 95 % 0,57–2,18) n’ont eu aucune in�u-ence sur la survie. Les scores Oxford d’évaluation du genou se sont améliorés, passant d’une moyenne de 40,2 à 21,8 en l’espace de 2 ans (p < 0,001), sans différence entre les groupes CoCr et OxZr.

A. Ross Demcoe, MD Eric R. Bohm, MD, MSc David R. Hedden, MD Colin D. Burnell, MD Thomas R. Turgeon, MD, MPH

Accepted July 10, 2018

Correspondence to: T. Turgeon 310-1155 Concordia Ave Winnipeg MB R2K 2M9 [email protected]

DOI: 10.1503/cjs.007518

oxidize-turgeon.indd 118 2019-03-19 8:37 AM

RESEARCH


T he Genesis II (Smith & Nephew) total knee implant system was released to the general market in 1996 and since its release has experienced good

overall performance. Case series from the design sur-geons1,2 as well as registry data3,4 have shown excellent sur-vivorship and functional results in patients undergoing primary total knee replacement with this system.5 The original system uses built-in external rotation of the femo-ral component in an attempt to optimize patellofemoral tracking, minimize the risk of femoral notching and pre-vent tibiofemoral shear forces in extension and �exion.1 Design variations were introduced into the system in 2005 to accommodate a condylar-constrained polyethylene insert requiring a symmetric posterior condyle design. This variant was initially branded as the Symmetric Poste-rior Condyle and later as the Legion Primary. In an effort to reduce wear, the components were made available in both standard cobalt–chromium (CoCr) bearing as well as an oxidized zirconium–niobium (OxZr) gradient ceramic bearing surface. The OxZr implants are predominantly zirconium–niobium metal with a 4- to 6-mm gradient layer of zirconium oxide on the articulating surface.6 Pro-posed bene�ts of the more expensive oxidized zirconium surface are increased wettability,7 decreased peak rough-ness,8 decreased adhesive and abrasive wear,7 and its appli-cability in patients with known metal sensitivities given the absence of nickel in the base alloy.9–11 The tibial com-ponent of this device is a titanium–aluminum–vanadium alloy that does not have the issue of nickel as seen in CoCr alloy femoral components.

There have been reports of seemingly minor changes in implant design having effects on overall survivorship and revision rates in both hip and knee arthroplasty.12,13 These occurrences show the need to critically evaluate the effect of minor design changes on implant function and survivorship.

Given the paucity of information regarding the survi-vorship of this iteration of these modi�cations to a success-ful design, we performed a study to determine the early to midterm survivorship of the Symmetric Posterior Condyle total knee arthroplasty components at a single institution. Additional goals were to assess for differences in revision risk between the OxZr and CoCr implants and to evaluate the impact of age, sex and body mass index on revision risk and assess the functional scores in patients in whom the prosthesis has been used.

METHODS

The institution’s ethics review board approved the study. All patients received the Genesis II Symmetric Posterior

Condyle or Legion Primary total knee implanted at the study centre between March 2007 and December 2013. All implants in the study were cemented and used a poste-rior stabilized design. Selection of OxZr versus CoCr was left to surgeon preference in discussion with the patient; however, selection of OxZr was restricted to patients aged less than 60 years with expected higher activity levels and a history of metal sensitivity, based on regional guidelines. Preoperatively, patients provided informed consent and were enrolled into a prospective research database. The database documents patient demographic characteristics, diagnosis, procedure details and outcome scores.

All procedures performed by the 4 participating fellowship-trained arthroplasty surgeons were eligible for inclusion in the study. Patients were followed at 6 weeks, 6 months, 1 year, 2 years and every 2 years thereafter for routine monitoring and surveillance. Follow-up data were captured from March 2007 until August 2014. Age at time of surgery, height and weight were documented, and body mass index was calculated. The bearing surface of the fem-oral component was also documented. In addition, we ana-lyzed the preoperative, 1-year postoperative and 2-year postoperative Oxford Knee Score values (best possible score 12, worst possible score 60). This instrument has proven to be reliable, valid and responsive to clinical change.14–16


We performed survival analysis using Kaplan–Meier tech-niques for the entire population and then separately for the CoCr and OxZr implants. We compared survivorship of the CoCr and OxZr version using the Cox proportional hazards method to control for the possible confounding effects of age,17,18 sex17–19 and body mass index19,20 on revi-sion risk. We compared demographic characteristics and functional score using t tests for continuous data and the χ2 test for dichotomous data.

RESULTS

A total of 2178 patients with 2815 primary total knee replacements were identi�ed in the database. The demo-graphic data can be found in Table 1. The mean age of the patients at the time of surgery was 64.7 (standard deviation [SD] 10.7) years, and the patients were predominantly female (1359 [62.4%]). The average body mass index of all eligible patients from the database was 35.0 (SD 8.5).

There were 2286 CoCr components and 529 OxZr components implanted. Patients who received the OxZr

Conclusion : La prothèse du genou à condyles postérieurs symétriques postéro- stabilisée possède une excellente survie à moyen terme. La surface portant l’implant n’a pas eu d’in�uence sur la survie à 5 ans et, donc, l’utilisation d’implants OxZr ne sera peut-être pas justi�ée.


RECHERCHE


implants were younger than those who received the CoCr implants and were predominantly male (Table 2). There was no difference in body mass index between the 2 implant groups.

Survivorship

Kaplan–Meier analysis of the pooled implants showed a 2-year survival rate of 98.2% (95% CI 97.6%–98.7%) for failure for any reason and a 5-year survival rate of 96.8% (95% CI 96.0%–97.7%) (Fig. 1). There were 63 cases that required revision, with infection being the most common reason (Table 3). Cox proportional hazards analysis showed that both female sex and increased age lowered revision risk and that there was a trend toward increased revision risk with increased body mass index, but this did not reach statistical signi�cance (Table 4). The individual CoCr and OxZr Kaplan–Meier survival curves are pre-sented in Figure 2. After adjustment for age, sex and body mass index, we could not detect any difference in survival between the 2 materials.

Outcome scores

There were 1851 total knee replacements with preopera-tive Oxford Knee Score values available for analysis. The mean preoperative score was 40.2 (SD 8.3). The mean score improved to 22.6 (SD 8.6) at 1 year postoperatively and 21.8 (SD 8.9) at 2 years. The improvement in the score at 1 year was statistically signi�cant (95% CI 17.06–18.09), as was the improvement at 2 years (95% CI 17.06–

18.09). No differences were found between patients who received OxZr implants and those who received CoCr implants.

DISCUSSION

We found that the overall (combined CoCr and OxZr) revision rate of the Legion Primary knee system was 1.8% at 2 years and 3.2% at 5 years. The Australian National Joint Replacement Registry shows a cumulative revision rate at 3 years of 1.8% for the posterior-stabilized CoCr bearing models and 3.2% for the posterior-stabilized OxZr models.3 Although this may re�ect a difference in the per-formance of these two materials, it may also result from selection bias from the OxZr products’ being used in patients at increased risk for failure, such as those less than age 55. At 5 years, the registry shows cumulative revision rates of 2.1% for CoCr models and 4.1% for OxZr models with posterior-stabilized designs.3 The �ndings in the cur-rent study are consistent with the results from the Austra-lian registry.

Our results also suggest that both age at time of surgery and sex have an influence on overall survivorship. Increased age at time of surgery decreased the revision rate, with a hazard ratio of 0.968, which is in keeping with registry data. Data from the National Joint Registry of England, Wales and Northern Ireland and the Isle of Man demonstrate decreasing cumulative revision rates at each incremental increase in age of 5 years.4 Similarly, the New Zealand Joint Registry illustrates a statistically signi�cant decrease in revision rate with each successive incremental increase in age of 10 years.21 In our series, female sex had a positive in�uence on implant survivorship. Again, this is in keeping with reports from several national joint regis-tries.3,4,21 The 2016 report from National Joint Registry of England, Wales and Northern Ireland and the Isle of Man

Table 1. Patient demographic and clinical characteristics

Characteristic No. (%) of patients*

Female sex (n = 2178) 1359 (62.4)

Age, yr, mean ± SD 64.7 ± 10.7

Body mass index, mean ± SD 35.0 ± 8.5

Cobalt–chromium implant (n = 2815) 2286 (81.2)

Length of follow-up, yr, mean ± SD 3.2 ±1.9


Table 2. Patient demographic characteristics by femoral component material

Characteristic

Material; no. (%) of implants*

p valueCobalt–chromium

n = 2286

Oxidized zirconium n = 529

Female sex 1483 (83.4) 296 (16.6) < 0.001

Male sex 803 (77.5) 233 (22.5)

Age, yr, mean ± SD 67.4 ± 9.4 52.3 ± 7.5 < 0.001

Body mass index, mean ± SD

35.0 ± 9.0 34.3 ± 8.3 0.1


Fig. 1. Pooled Kaplan–Meier product-limit survival estimate of the Genesis II Symmetric Posterior Condyle/Legion Primary total knee implant (Smith & Nephew). Grey shading represents 95% confidence intervals.

1.00

0.95

0.90

0.85

0.80

Sur

viva

l pro

babi

lity

Time, yr0 2 4 6 8


RESEARCH


indicates that males are more likely to undergo revision than females of comparable age if they were less than 75 years old at the time of primary surgery.4 Given that the predominant indication for revision in our study was infec-tion, this may be related to the �nding of Rasouli and col-leagues22 that male sex was associated with an increased risk of surgical site infection, with an odds ratio of 1.79.

Femoral bearing surface did not have a statistically sig-ni�cant in�uence on overall survivorship of the Legion Primary implant in our series, even when we controlled for

the possible confounding effects of age, sex and body mass index. The OxZr implant is a premium implant with an associated increase in cost. The differential varies between centres based on regional contracting; however, the cost of the OxZr femoral component may be as much as 4 times that of the otherwise identical CoCr component. Tribo-logical studies have suggested superior characteristics of OxZr femoral implants in comparison to the standard CoCr femoral implants.7,23–25 Interestingly, these superior characteristics have not resulted in consistent published reports of improvement in in vivo wear or superior clinical results in knee arthroplasty. This is in contrast to total hip arthroplasty, where data from the Australian joint registry suggest superiority of this bearing surface in terms of sur-vival.3 Matched comparison retrieval studies have shown a decrease in femoral component damage in OxZr compo-nents8,26,27 and a corresponding decrease in polyethylene damage.26 Synovial �uid analysis in patients with bilateral total knee replacement failed to show differences in poly-ethylene particle number, weight, size and shape between the CoCr knee and the OxZr knee,28 and no clinical differ-ences were found between groups.29 A small randomized controlled trial evaluating OxZr and CoCr implants in patients undergoing simultaneous total knee arthroplasty procedures failed to show signi�cant differences in clinical, subjective and radiographic outcomes at 5 years.30 Vertullo and colleagues31 reviewed 17 577 cases from the Australian joint registry involving a cruciate-retaining asymmetric-�exion knee and found no difference in implant survival in patients less than 75 years of age, with poorer survival of OxZr implants at or beyond this age. Our study supports these �ndings in a posterior-stabilized symmetric �exion knee by the same single manufacturer.

CONCLUSION

Although there are theoretical advantages to using the OxZr implant bearing surface, we did not detect an improvement in early to midterm implant survivorship. With no apparent survival bene�t to the OxZr surface in knees, the use of these more expensive implants may not be warranted to extend implant longevity until longer-term evidence of superiority is presented. The symmetric posterior condylar posterior-stabilized design as a whole offers excellent early to midterm survivorship, with a cumulative revision rate of 1.8% at 2 years and 3.2% at 5 years in the current study. The �ndings support the con-tinued use of this design iteration, which is based on a pre-vious implant with an excellent record. Both increased age and female sex had a positive in�uence on implant survi-vorship; however, we could not detect any effect of body mass index on survivorship.

Acknowledgements: This study was supported by funds granted from the Thorlakson Chair at the Department of Surgery, University of Manitoba.

Fig. 2. Kaplan–Meier product-limit survival estimates for cobalt–chromium (CoCr) and oxidized zirconium (OxZr) femoral implants. Grey shading represents 95% confidence intervals.

0 2 4 6

1.00

0.95

0.90

0.85

0.80

Sur

viva

l pro

babi

lity

Duration of implant survival, yr

CoCrOxZr

Table 3. Reasons for revision

DiagnosisNo. of cases

n = 63

Infection 23

Instability 9

Stiffness 6

Extensor mechanism injury 5

Aseptic loosening 3

Patellar maltracking 7

Locking mechanism dissociation 2

Hematoma 1

Recurrent hemarthrosis 1

Pain 2

Fracture 1

Patellar clunk 1

Undocumented 2

Table 4. Risk of revision by age, sex, body mass index and femoral implant material

Predictor HR (95% CI)

Oxidized zirconium femur 1.110 (0.567–2.179)

Age 0.968 (0.939–0.997)

Female sex 0.450 (0.269–0.754)

Body mass index 1.023 (0.994–1.054)



RECHERCHE


Af�liations: From Kelowna Orthopedics, Kelowna, BC (Demcoe); and the Concordia Hip and Knee Institute, Division of Orthopaedic Sur-gery, University of Manitoba, Winnipeg, Man. (Bohm, Hedden, Burnell, Turgeon).

Competing interests: A.R. Demcoe has no conflicts to declare. E.R. Bohm, D.R. Hedden, C.D. Burnell and T.R. Turgeon all declare institutional research support (unrelated to this project) from DePuy Synthes, Hip Innovation Technology, Smith & Nephew and Zimmer, and institutional fellowship educational funding from DePuy Synthes and Smith & Nephew. E.R. Bohm declares speakers bureau/paid presen-tations for Zimmer, DePuy Synthes and Stryker and unpaid consultancy for Smith & Nephew. D.R. Hedden declares a paid consultancy for Smith & Nephew. T.R. Turgeon declares speakers bureau/paid presen-tations for DePuy Synthes and unpaid consultancy for Smith & Nephew.

Contributors: A.R. Demcoe, E.R. Bohm and T.R. Turgeon designed the study. D.R. Hedden and C.D. Burnell acquired the data, which A.R. Demcoe, E.R. Bohm and T.R. Turgeon analyzed. A.R. Demcoe and T.R. Turgeon wrote the article, which all authors reviewed. All authors approved the �nal version to be published and can certify that no other individuals not listed as authors have made substantial contri-butions to the paper.

References

1. Bourne RB, Laskin RS, Guerin JS. Ten-year results of the �rst 100 Genesis II total knee replacement procedures. Orthopedics 2007;30 (Suppl):83-5.

2. Laskin RS, Davis J. Total knee replacement using the Genesis II prosthesis: a 5-year follow up study of the �rst 100 consecutive cases. Knee 2005;12:163-7.

3. Annual report 2015. Adelaide (Australia): Australian Orthopaedic Association National Joint Replacement Registry; 2015.

4. 13th annual report. Hemel Hempstead (UK): National Joint Registry for England, Wales, Northern Ireland and the Isle of Man; 2016.

5. Bhandari M, Pascale W, Sprague S, et al. The Genesis II in primary total knee replacement: a systematic literature review of clinical out-comes. Knee 2012;19:8-13.

6. Spector BM, Ries MD, Bourne RB, et al. Wear performance of ultra-high molecular weight polyethylene on oxidized zirconium total knee femoral components. J Bone Joint Surg Am 2001;83-A (Suppl 2 Pt 2):80-6.

7. White SE, Whiteside LA, McCarthy DS, et al. Simulated knee wear with cobalt chromium and oxidized zirconium knee femoral compo-nents. Clin Orthop Relat Res 1994;309:176-84.

8. Brandt JM, Guenther L, O’Brien S, et al. Performance assessment of femoral knee components made from cobalt–chromium alloy and oxidized zirconium. Knee 2013;20:388-96.

9. Laskin RS. An oxidized Zr ceramic surfaced femoral component for total knee arthroplasty. Clin Orthop Relat Res 2003;416:191-6.

10. Innocenti M, Carulli C, Matassi F, et al. Total knee arthroplasty in patients with hypersensitivity to metals. Int Orthop 2014;38:329-33.

11. Hofer JK, Ezzet KA. A minimum 5-year follow-up of an oxidized zirconium femoral prosthesis used for total knee arthroplasty. Knee 2014;21:168-71.

12. Han HS, Kang SB, Yoon KS. High incidence of loosening of the femoral component in Legacy posterior stabilised-�ex total knee replacement. J Bone Joint Surg Br 2007;89:1457-61.

13. Middleton RG, Howie DW, Costi K, et al. Effects of design changes on cemented tapered femoral stem �xation. Clin Orthop Relat Res 1998;355:47-56.

14. Dawson J, Fitzpatrick R, Murray D, et al. Questionnaire on the per-ceptions of patients about total knee replacement. J Bone Joint Surg Br 1998;80:63-9.

15. Garratt AM, Brealey S, Gillespie WJ, et al. Patient-assessed health instruments for the knee: a structured review. Rheumatology (Oxford) 2004;43:1414-23.

16. Murray DW, Fitzpatrick R, Rogers K, et al. The use of the Oxford hip and knee scores. J Bone Joint Surg Br 2007;89:1010-4.

17. Dy CJ, Marx RG, Bozic KJ, et al. Risk factors for revision within 10 years of total knee arthroplasty. Clin Orthop Relat Res 2014;472: 1198-207.

18. Koh IJ, Cho WS, Choi NY, et al.; Kleos Korea Research Group. Causes, risk factors, and trends in failures after TKA in Korea over the past 5 years: a multicenter study. Clin Orthop Relat Res 2014;472: 316-26.

19. Pfefferle KJ, Gil KM, Fening SD, et al. Validation study of a pooled electronic healthcare database: the effect of obesity on the revision rate of total knee arthroplasty. Eur J Orthop Surg Traumatol 2014;24: 1625-8.

20. Si HB, Zeng Y, Shen B, et al. The in�uence of body mass index on the outcomes of primary total knee arthroplasty. Knee Surg Sports Traumatol Arthros 2015;23:1824-32.

21. The New Zealand Joint Registry: sixteen year report. January 1999 to December 2014. Wellington (NZ): New Zealand Orthopaedic Associ-ation; 2015.

22. Rasouli MR, Restrepo C, Maltenfort MG, et al. Risk factors for sur-gical site infection following total joint arthroplasty. J Bone Joint Surg Am 2014;96:e158.

23. Tsukamoto R, Chen S, Asano T, et al. Improved wear performance with crosslinked UHMWPE and zirconia implants in knee simula-tion. Acta Orthop 2006;77:505-11.

24. Ezzet KA, Hermida JC, Colwell CW Jr, et al. Oxidized zirconium femoral components reduce polyethylene wear in a knee wear simu-lator. Clin Orthop Relat Res 2004;428:120-4.

25. Ezzet KA, Hermida JC, Steklov N, et al. Wear of polyethylene against oxidized zirconium femoral components effect of aggressive kinematic conditions and malalignment in total knee arthroplasty. J Arthroplasty 2012;27:116-21.

26. Heyse TJ, Chen DX, Kelly N, et al. Matched-pair total knee arthro-plasty retrieval analysis: oxidized zirconium vs. CoCrMo. Knee 2011;18:448-52.

27. Heyse TJ, Elpers ME, Nawabi DH, et al. Oxidized zirconium versus cobalt–chromium in TKA: pro�lometry of retrieved femoral compo-nents. Clin Orthop Relat Res 2014;472:277-83.

28. Kim YH, Kim JS, Huh W, et al. Weight of polyethylene wear parti-cles is similar in TKAs with oxidized zirconium and cobalt–chrome prostheses. Clin Orthop Relat Res 2010;468:1296-304.

29. Kim YH, Park JW, Kim JS. Comparison of the Genesis II total knee replacement with oxidised zirconium and cobalt–chromium femoral components in the same patients: a prospective, double-blind, ran-domised controlled study. J Bone Joint Surg Br 2012;94:1221-7.

30. Hui C, Salmon L, Maeno S, et al. Five-year comparison of oxidized zirconium and cobalt–chromium femoral components in total knee arthroplasty: a randomized controlled trial. J Bone Joint Surg Am 2011;93:624-30.

31. Vertullo CJ, Lewis PL, Graves S, et al. Twelve-year outcomes of an Oxinium total knee replacement compared with the same cobalt–chromium design: an analysis of 17,577 prostheses from the Austra-lian Orthopaedic Association National Joint Replacement Registry. J Bone Joint Surg Am 2017;99:275-83.




Effect of predicted travel time to trauma care on mortality in major trauma patients in Nova Scotia

Background: Trauma is a leading contributor to the burden of disease in Canada, accounting for more than 15 000 deaths annually. Although caring for injured patients at designated trauma centres (TCs) is consistently associated with survival bene�ts, it is unclear how travel time to de�nitive care in�uences outcomes. Using a population-based sample of trauma patients, we studied the association between predicted travel time (PTT) to TCs and mortality for patients assigned to ground transport.

Methods: Victims of penetrating trauma or motor vehicle collisions (MVCs) in Nova Scotia between 2005 and 2014 were identi�ed from a provincial trauma regis-try. We conducted cost distance analyses to quantify PTT for each injury location to the nearest TC. Adjusted associations between TC access and injury-related mortality were then estimated using logistic regression.

Results: Greater than 30 minutes of PTT to a TC was associated with a 66% increased risk of death for MVC victims (p = 0.045). This association was lost when scene deaths were excluded from the analysis. Sustaining a penetrating trauma greater than 30 minutes from a TC was associated with a 3.4-fold increase in risk of death. Following the exclusion of scene deaths, this association remained and approached signi�cance (odds ratio 3.48, 95% con�dence interval 0.98–14.5, p = 0.053).

Conclusion: Predicted travel times greater than 30 minutes were associated with worse outcomes for victims of MVCs and penetrating injuries. Improving communi-cation across the trauma system and reducing prehospital times may help optimize outcomes for rural trauma patients.

Contexte : Les traumatismes contribuent pour une bonne part au fardeau de la mala-die au Canada; on leur attribue plus de 15 000 décès annuellement. Même si les soins prodigués aux patients victimes de traumatismes dans les centres de traumatologie désignés (CTD) sont toujours associés à des gains au plan de la survie, on ignore quelle est l’in�uence du temps de transfert vers le CTD sur l’issue. À partir d’un échantillon de patients polytraumatisés basé dans la population, nous avons analysé le lien entre le temps de transfert prévu (TTP) vers le CTD et la mortalité des patients transportés par voie terrestre.

Méthodes : On a identi�é les victimes de traumatismes pénétrants ou d’accidents de la route en Nouvelle-Écosse entre 2005 et 2014 à partir d’un registre provincial de traumatologie. Nous avons analysé la distance de coût pour quanti�er le TTP à partir de chaque scène vers le CTD le plus proche. Les liens ajustés entre l’accès au CTD et la mortalité liée au traumatisme ont ensuite été estimés par régression logistique.

Résultats : Un délai de TTP de plus de 30 minutes pour arriver au CTD a été asso-cié à un accroissement de 66 % du risque de décès chez les patients polytraumatisés (p = 0,045). Ce lien s’annulait si on excluait de l’analyse les décès survenus sur la scène de l’accident. Subir un traumatisme ouvert à plus de 30 minutes de distance d’un CTD a été associé à une augmentation par un facteur de 3,4 du risque de décès. Une fois les décès sur la scène de l’accident exclus, ce lien a persisté et s’est rapproché du seuil de signi�cation (rapport des cotes 3,48, intervalle de con�ance de 95 % 0,98–14,5, p = 0,053).

Conclusion : Des temps de transfert prévus supérieurs à 30 minutes ont été associés une issue plus défavorable pour les victimes d’accidents de la route et de traumatismes pénétrants. L’amélioration de la communication entre les divers éléments du système de traumatologie et la réduction du temps préhospitalier pourrait optimiser l’issue pour les patients victimes de traumatismes en région rurale.

Gavin Tansley, MD, MPH Nadine Schuurman, PhD Matthew Bowes, MD Mete Erdogan, PhD, MHI Robert Green, MD, DABEM Mark Asbridge, PhD Natalie Yanchar, MD, MSc

Accepted Aug. 28, 2018

Correspondence to: G. Tansley QEII Health Sciences Centre 1276 South Park St Halifax NS B3H 2Y9 [email protected]

DOI: 10.1503/cjs.004218

effect-green.indd 123 2019-03-19 1:07 PM

RECHERCHE


I njury is one of the largest public health concerns in Can-ada, accounting for 15 000 deaths and more than $20 bil-lion in direct and indirect costs annually.1 In response to

this substantial disease burden, trauma systems have been established to match patient needs with appropriate health facility resources.2 Preferentially triaging severely injured patients directly to specialized trauma centres (TCs) has been shown to reduce mortality.3 Although regionalization of trauma care services reduces trauma-related mortality,4 it has signi�cant access implications because of the concentra-tion of specialized resources in few discrete geographic loca-tions. Distance to a TC correlates with time to de�nitive care, and there is evidence that shorter prehospital times are associated with increased survival.5–9

Health care access, de�ned as the degree of �t between patient needs and the health care system, is a relevant con-cept to policy-makers, as poor access may negatively impact health care utilization.10 Access has classically been distilled into 5 dimensions: availability, accessibility, accommodation, affordability and acceptability.10–12 Furthermore, access can be categorized by spatial (accessibility, availability) and non-spatial factors (affordability, acceptability, accommoda-tion).13 The importance of the spatial relationship between patients and health care services becomes clear when access is framed, in part, as a geographic construct. Although patient transport time can be used to analyze access to care, these times may be biased by various factors, including transport urgency and local triage practices. An alternative approach to determine the influence of geography on trauma patient outcomes is to estimate transport time from injury location to a TC using geospatial analysis.

Predicted travel time (PTT) is a measure of spatial access to de�nitive trauma care. Given that signi�cant resources are invested to maintain TC readiness, understanding how PTTs in�uence patient outcomes may help improve the equity and ef�ciency of trauma care delivery by informing changes to trauma system organization.14 One Canadian study has previously shown that greater PTTs to trauma care result in lower TC utilization rates following major injury.15 Variability in PTTs to Canadian TCs has also been reported.16,17 However, the relationship between PTTs to TCs and mortality following major trauma remains unclear. The objective of our study was to evaluate the association between mortality and PTT to TCs for patients assigned to ground transport by the Emergency Health Services (EHS) system using a population-based sample of trauma patients in Nova Scotia and linking injury location with data from a provincial trauma registry.

METHODS

Setting

Nova Scotia is the second most densely populated prov-ince in Canada with an average population density of

18 persons/km2.18,19 Notably, most of the population resides outside the only census metropolitan area; thus, rural traumas are commonly seen across the province. Trauma care in Nova Scotia is available at 8 level-III TCs, 1 adult level-I TC, and 1 pediatric level-I TC.17 All level-III TCs in Nova Scotia are capable of performing emer-gency, life-saving surgical services (e.g., trauma laparot-omy). Emergency Health Services provide ground-based prehospital transport using a comprehensive network of ground ambulances. Although aeromedical transport is also available, these assets are used in fewer than 5% of scene responses and are therefore not the focus of this study. If paramedics are within 30 minutes of a level-I or level-III TC (including estimated extrication time), the patient is taken to the highest-level TC. In cases where paramedics are not within 30 minutes of a level-I or level-III TC, the destination and launch decision is made on a case-by-case basis by paramedics and air medical transport (AMT) personnel based on patient condition, response time of AMT (i.e., EHS LifeFlight), transport time by ground ambulance to the nearest emergency department (ED) compared with the nearest TC, capability of the nearest ED, and the level of care available by paramedics (present and from a possible intercepting crew).20

Study design and data collection

This observational geospatial analysis used data obtained from the Nova Scotia Trauma Registry (NSTR). The NSTR is a provincial population-based trauma registry under the Nova Scotia Department of Health & Wellness that captures demographic and clinical data on all major traumas occurring in the province.21 Emergency Health Services (or the police in cases of scene deaths) record the coordinates of the pickup location of all victims using global positioning systems (GPS). These data are abstracted into the NSTR.

We limited our study to trauma patients who were injured in a motor vehicle collision (MVC) or by a penetrating mechanism, as these are known to be time-sensitive cases requiring rapid transport to the nearest TC. All traumas with an Injury Severity Score (ISS) greater than 11 related to MVCs (ICD-10 V01 to V99) or penetrating mechanisms (ICD-10 W25, W26, W32–34, W45, X72–74, X78, X93–95, X99, Y22–24) between Jan. 1, 2005, and Dec. 31, 2013, were eligible for inclusion. Individuals who were missing GPS coordinates or whose pickup location was inconsistent with the injury location were excluded. We excluded any trauma cases that were solely air transport cases or that involved a combination of ground and air transport. All duplicate entries were removed before analysis.

The locations of TCs and the provincial road network used in the spatial analyses were obtained from a commer-cially available data set (CanMap, DMTI Spatial). We used commercially available geographic information system (GIS) software (ArcMap, Esri) for all geospatial analyses.


RESEARCH


Cost distance analysis

Cost distance analyses were performed to model travel times from all points in Nova Scotia to the nearest TC. This validated method calculates the accumulated travel cost in minutes associated with travelling across a surface from any point in the study area to speci�ed destinations (i.e., TCs). For use in these analyses, we constructed a 100 m2 gridded cost surface using the provincial road net-work and each road segment’s corresponding speed limit. Cells without a road were assigned a value corresponding to a travel speed of 5 km/h–1 (i.e., the average speed of walking). As prehospital transport is expected to use estab-lished road networks predominantly, other barriers, such as hydrologic features, were not incorporated into the cost surface. The �nal output was a continuous surface where each pixel corresponded to the time required to travel from that geographic location to the nearest TC. Further details on the model development and validation for emer-gency transport in Nova Scotia are discussed elsewhere.22

Predicted travel time to trauma care

Overlaying the point locations of major traumas related to MVCs or penetrating mechanisms over the cost distance out-puts allowed us to estimate PTT to trauma care for a cohort of patients from the NSTR. The PTT corresponding to the point location of each injury was extracted from the cost dis-tance outputs and incorporated into the statistical models.

Statistical model building

Logistic regression models of mortality risk among pene-trating trauma victims and MVC-related trauma victims were estimated using time to TC care (< 10 minutes [refer-ence], 10–20 minutes, 20–30 minutes, > 30 minutes), age (continuous), sex (female [reference], male), socioeconomic status (SES; high SES [reference], low SES), and ISS (con-tinuous) as covariates. All covariates were de�ned a priori based on previously identified associations. We defined SES using the Vancouver Area Neighbourhood Derivation Index (VANDIX) of the patient’s residential location.23 For cases where residential postal codes were unavailable, injury location was used as a proxy for place of residence. The most deprived quintile was de�ned as low SES. The out-come of interest was death, either in-hospital or before arrival. Bivariate comparisons were performed using the Student t test or χ2 test, where appropriate. Adjusted analy-ses were performed, including and excluding scene deaths, to better delineate potential impacts of postinjury care on observed associations. Scene deaths included patients who died from their injuries before the arrival of police or EHS as well as patients who died at the scene while being cared for by EHS. Spatial autocorrelation of model residuals was excluded using the Global Moran’s I statistic. We per-formed a sensitivity analysis of PTTs only to level-I TCs in Nova Scotia (i.e., excluding level-III TCs) to evaluate the consistency of observed results. Statistical analyses were performed using Stata software version 14 (StataCorp).

RESULTS

Selection and characteristics of the study population

Between Jan. 1, 2005, and Dec. 13, 2013, a total of 1568 MVC-related traumas and 243 penetrating traumas were eligible for inclusion. Excluded from this study were 1675 major traumas from all other mechanisms. Following the exclusion of duplicates and entries with missing or inconsis-tent injury locations, 1304 cases of MVC-related trauma and 231 cases of penetrating trauma remained for spatial analysis (Fig. 1). All TCs were successfully geolocated. Most MVC-related traumas (64.0%) and pene-trating injury cases (57.0%) occurred in locations outside of Halifax Regional Municipality, which is the sole urban area in the province.

The general characteristics of the study population are described

Fig. 1. Selection of study participants. MVC = motor vehicle collision; NSTR = Nova Scotia Trauma Registry.

Traumas with relevant mechanisms of injury in the NSTR, 2005-2014

n = 1811

Mechanism: MVC

n = 1568

Mechanism: penetrating injury

n = 243

MVC-related traumas suitable for spatial analysis

n = 1304

Penetrating traumas suitable for spatial analysis

n = 231

264 cases excluded: Inconsistent injury location n = 110 Missing injury location n = 67 Duplicate entry n = 87

12 cases excluded: Inconsistent injury location n = 0 Missing injury location n = 5 Duplicate entry n = 7


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in Table 1. Victims of both injury mechanisms were predominantly young and male. Among victims of penetrating injuries, 67.0% of trauma cases were related to firearms, while the rest were caused by sharp objects, most commonly knives. Mortality among patients injured in MVCs was 25.0%, com-pared with 56.3% in patients with penetrating trauma. In both groups, patients who died had a higher mean ISS, and a greater proportion of these patients had low SES.

Predicted travel time to trauma care in Nova Scotia via ground transport

The cost distance analysis of ground-based travel time to trauma care in the province of Nova Scotia is illustrated in Figure 2. This analysis showed regional variation in TC accessibility, but most points in the province were found to have access to a TC within 30 minutes of driv-ing time. Overall, 29.7% of MVC-related traumas were located more than 30 minutes away from a TC (median

Fig. 2. Results of cost distance analysis showing predicted travel times to trauma care in Nova Scotia.

Level-III trauma centres

Level-I trauma centres

Travel time

120 mins

15 minskm0 0 50 100 150 200

Potential spatial accessto trauma centres

Table 1. Characteristics of patients with MVC-related or penetrating traumas

Characteristic

MVC-related traumas Penetrating traumas

All patientsn = 1304

Patients who diedn = 326

All patientsn = 231

Patients who diedn = 130

Age, yr, mean ± SD 39.4 ± 20.7 43.3 ± 22.7 44.6 ± 20.3 54.3 ± 17.5

Male sex, no. (%) 896 (68.7) 241 (73.9) 217 (93.9) 126 (96.9)

Incident location < 30 min from TC* care, no. (%)

Yes 917 (70.3) 209 (64.1) 169 (73.2) 83 (63.9)

No 387 (29.7) 117 (35.9) 62 (26.8) 47 (36.2)

ISS, mean ± SD 27.4 ± 14.3 40.2 ± 18.6 28.9 ± 17.8 35.0 ± 19.6

Low SES, no. (%) 329 (21.9) 96 (26.0) 78 (33.8) 41 (52.3)

LOS, d, mean ± SD 15.7 ± 25.6 4.0 ± 10.7 14.6 ± 19.4 3.0 ± 5.1

Surgical procedure, no. (%) 524 (40.2) 30 (9.2) 81 (35.1) 9 (6.9)

ISS = Injury Severity Score; LOS = length of hospital stay; MVC = motor vehicle collision; SD = standard deviation; SES = socioeconomic status; TC = trauma centre.

*Includes both level-I and level-III TCs


RESEARCH


time to TC 19.3 minutes), and 26.8% of penetrating traumas were located more than 30 minutes away from a TC (median time to TC 15.3 minutes).

Comparison of trauma cases by predicted driving time from a trauma centre

Using bivariate analyses, we compared the characteristics of patients with MVC-related and penetrating traumas who were located 30 minutes or less of PTT from a TC with those of patients located more than 30 minutes away (Table 2). For MVC-related traumas, the unad-justed mortality was higher for individuals injured more than 30 minutes away from a TC (30.2/100 persons v. 22.8/100 persons, p = 0.005). Pre-, post- and on-scene intervals were all higher in the group injured more than 30 minutes away from a TC (all p < 0.001). The prob-ability of scene death was also higher for individuals injured more than 30 minutes away from a TC (20.9% v. 12.2%, p < 0.001). Finally, individuals injured more than 30 minutes away from a TC were 52% more likely to be ejected from the vehicle (p < 0.001).

For penetrating traumas, the unadjusted mortality was significantly higher for individuals injured more than 30 minutes away from a TC (75.8/100 persons v. 49.1/100 persons, p < 0.001). Individuals injured more than 30 minutes away from TCs were also more likely to be older, to experience longer prehospital intervals, and to have low SES (all p < 0.01).

Influence of predicted travel time to trauma care on mortality

We conducted logistic regression analyses to determine the in�uence of PTT to trauma care on mortality (Table 3).

Among patients with MVC-related traumas, lack of PTT to TC care within 30 minutes was found to be associated with a 66% increased risk of death after adjustment for the confounding variables of age, sex, ISS and SES identi�ed a priori (odds ratio [OR] 1.66, 95% con�dence interval [CI] 1.09–2.52, p = 0.018). Importantly, this association was lost when scene deaths were excluded from the analysis (OR 0.93, 95% CI 0.58–1.46, p = 0.78,). In addition, the likeli-hood of death in these patients was associated with male sex (OR 1.45, p = 0.034), increasing age (OR 1.02, p < 0.001), and increasing ISS (OR 1.11, p < 0.001).

In cases of penetrating trauma, a PTT greater than 30 minutes from a TC was associated with increased likeli-hood of death (OR 3.43, 95% CI 1.37–8.59, p = 0.039). Notably, after scene deaths were excluded from the analy-sis, this association remained and approached signi�cance (OR 3.48, 95% CI 0.98–14.5, p = 0.053). Risk of death was also associated with increasing age (OR 1.06, p < 0.001) and increasing ISS (OR 1.07, p < 0.001) in these patients. Results were similar following a sensitivity analysis of PTTs only to level-I TCs in the province. There was no difference in the likelihood of death between patients taken to a level-I TC and those transported to a level-III TC.

DISCUSSION

Owing to the high prevalence of rural trauma in Canada, it is important to understand the impact of trauma care accessibility on the outcomes of injured patients. By com-bining spatial analyses with more traditional statistical models, our study shows that trauma patients injured in areas with higher PTTs to TCs have increased risk of death. The magnitude of this association and the poten-tial explanations underlying it are modi�ed by the mech-anism of injury. Greater likelihood of ejection following

Table 2. Patient characteristics and predicted travel time to a trauma centre

Characteristic


TC > 30 minn = 387

TC ≤ 30 minn = 917 p value

TC > 30 minn = 62

TC ≤ 30 minn = 169 p value

Unadjusted mortality* 30.2 22.8 0.0051 75.8 49.1 < 0.001

Age, yr, mean ± SD 39.3 ± 20.8 39.4 ± 20.6 0.90 52.0 ± 20.5 41.9 ± 19.6 < 0.001

Male sex, no. (%) 281 (72.6) 615 (67.1) 0.047 60 (96.8) 157 (92.9) 0.24

ISS, mean ± SD 27.7 ± 14.0 27.2 ± 14.4 0.60 30.9 ± 17.8 28.2 ± 17.8 0.32

Prehospital time, min, mean ± SD

Prescene time 18.6 ± 12.8 12.7 ± 8.6 < 0.001 25.9 ± 18.8 13.4 ± 35.5 0.0037

Scene time 35.5 ± 22.4 28.8 ± 19.3 < 0.001 34.3 ± 23.8 18.6 ± 21.3 < 0.001

Postscene time 35.4 ± 25.5 26.8 ± 31.3 < 0.001 44.7 ± 31.3 13.8 ± 8.7 < 0.001

Extrication time 35.8 ± 20.9 33.6 ± 20.5 0.60 — — —

On-scene death, no. (%)† 81 (20.9) 112 (12.2) < 0.001 39 (62.9) 74 (43.8) < 0.001

Ejection, no. (%) 122 (31.5) 189 (20.6) < 0.001 — — —

Low SES, no. (%) 117 (30.3) 154 (17.1) < 0.001 32 (51.6) 48 (28.4) 0.004

EHS = Emergency Health Services; ISS = Injury Severity Score; MVC = motor vehicle collision; SD = standard deviation; SES = socioeconomic status; TC = trauma centre.

*Per 100 persons.

†Included deaths at the scene before arrival of police/EHS or while being cared for by EHS.


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MVCs in more remote areas suggests that behavioural patterns, such as seatbelt use, may be at least partly responsible for this association. In cases of penetrating trauma, higher PTTs to trauma care also increased the risk of death, and this association remained following exclusion of scene deaths, suggesting that differences in postinjury care or prolonged postscene transport times may explain this result. Patients injured in areas with poorer access to trauma care were more likely to experi-ence longer prehospital intervals; thus, reducing prehos-pital times for patients injured in remote locations may improve their outcomes.

In Nova Scotia, prehospital time depends on triage practices as well as transport urgency and is therefore an inherently biased variable. In our data set there was an inverse relationship between recorded prehospital time and probability of survival for MVC victims, providing evidence that analyzing prehospital time alone is inade-quate for complex trauma systems. Access is an independ-ent variable determined exclusively by injury location and the location of TCs. As such, geospatial analysis is a superior way to study the influence of geography on patient outcomes than using actual transport times. To serve system decision-makers better, our method of analy-sis could be adapted to focus on subpopulations with spe-ci�c needs and timelines to care, such as patients with traumatic brain injuries.24

It has been demonstrated consistently that caring for injured patients in a designated TC is associated with higher survival rates.5,25 Accordingly, �eld triage guide-lines have been developed that stipulate that an injured patient should be transported directly to a designated TC irrespective of its proximity to the injury location.26 Sev-eral studies have identi�ed signi�cant survival bene�ts in trauma patients with shorter prehospital times.5–9 Longer prehospital times are typical for rural trauma patients,27–29 suggesting a potential disadvantage for those injured in remote areas, which is consistent with our results. In con-trast to our �ndings, however, some have found that pre-

hospital time has no effect on patient mortality,30 while others report decreased overall mortality as the prehospi-tal time interval increased.31 In their comparison of rural and urban trauma patients in Oregon and Washington, Newgard and colleagues32 found that mortality was not signi�cantly different for rural and urban patients; how-ever, a greater proportion of rural deaths occurred shortly after injury. Prehospital time may be more impor-tant for certain subgroups of trauma patients. In a sys-tematic review of 20 studies, Harmsen and colleagues9 found that shorter prehospital intervals conferred survival benefits only for patients with central nervous system injuries and hemodynamically unstable patients injured by penetrating mechanisms, presumably because of the high prevalence of life-threatening surgical lesions in these populations. Importantly, Brown and colleagues33 showed that not all prehospital time is equal. The authors reported increased risk of death in patients with pro-longed scene time, while prolonged response time and transport time were not associated with mortality. Fur-thermore, Roisilen and colleagues34 showed that the esti-mated effect of transport time to hospital care varies depending on the choice of statistical model.

Few Canadian studies have examined the impact of PTTs to trauma care on patient outcomes. Lawson and colleagues17 reported increased unadjusted mortality for patients injured more than 60 minutes of driving time from level-I or level-II trauma care. Although the authors used residential postal codes as a surrogate for injury location, this surrogate could be inaccurate in trauma systems dominated by blunt mechanisms.35 In addition, the unadjusted nature of the statistical analysis limited their ability to elucidate any potential explana-tory or confounding factors underlying their �ndings. This is important, given the potential for individuals injured in rural locations to have additional risk factors, such as lower SES or more severe injuries, for adverse outcomes.29,36 Some of the first trauma-related spatial analyses that incorporated multivariable statistical models

Table 3. Logistic regression analysis on influence of predicted travel time on mortality

Variable


Adjusted OR (95% CI) p value Adjusted OR (95% CI) p value

Time to TC care — 0.045 — 0.039

< 10 min Reference — Reference —

10–20 min 1.02 (0.65–1.60) — 1.53 (0.65–3.62) —

20–30 min 1.13 (0.70–1.82) — 2.96 (0.86–10.21) —

> 30 min 1.66 (1.09–2.52) — 3.43 (1.37–8.59) —

Male sex 1.45 (1.03–2.04) 0.034 1.47 (0.36–6.08) 0.59

Age 1.02 (1.02–1.03) < 0.001 1.06 (1.04–1.08) < 0.001

ISS 1.11 (1.09–1.12) < 0.001 1.07 (1.03–1.10) < 0.001

Low SES 1.19 (0.83–1.71) 0.35 0.88 (0.41–1.87) 0.74

CI = confidence interval; ISS = Injury Severity Score; MVC = motor vehicle collision; OR = odds ratio; SES = socioeconomic status; TC = trauma centre.


RESEARCH


were conducted by Crandall and colleagues37 in urban Chicago. After adjusting for several relevant confound-ing variables, the authors reported that victims of �rearm-related penetrating trauma had 23% increased odds of death if they were injured farther than 5 minutes of PTT from a TC. However, the urban setting and spe-ci�c population of this study provides little generalizabil-ity to a rural trauma system.

Minimizing the time interval between injury and pro-vision of definitive care, a concept popularized as the “golden hour,” has been a central tenet of postinjury care for decades.38 Several studies have reported improved sur-vival for rural trauma patients transported by air com-pared with ground transport,39,40 including 2 previous studies performed in Nova Scotia.41,42 While EHS Life-Flight is able to land at or near the scene of an emer-gency, this occurs relatively infrequently, which is why we limited our analysis to ground transport. At the time of the study, EHS LifeFlight did not autolaunch; these services had to be requested by ground paramedics. As early activation of EHS LifeFlight is critical to getting the patient to definitive care, better communication across the trauma system may improve patient outcomes. Outcomes could also be improved by having advanced paramedics on rural EHS services. Finally, increased edu-cational efforts may improve outcomes in rural trauma cases. The Rural Trauma Team Development Course (RTTDC) is based on the concept that in most situations, rural facilities can form a trauma team consisting of at least 3 core members,43 and there is evidence that partici-pation in RTTDC can signi�cantly improve prehospital times.44,45 In Nova Scotia, the provincial trauma program (Trauma Nova Scotia) began offering RTTDC in 2012. Further research is warranted to determine whether implementation of RTTDC training in Nova Scotia has had an effect on outcomes in rural trauma patients.

Limitations

Our study has several limitations. First, it is subject to the known limitations of retrospective data collection. The accuracy of injury location data collected by EHS has not been externally validated and relies on personnel manually indicating when they arrive on scene. There is, however, an available �eld that denotes the reliability of the coordinates, which helped to avoid the inclusion of inaccurate data in the analysis. Second, the analysis was limited to data from a single provincial trauma system. Although the epidemiology of injuries in Nova Scotia is comparable to that in other provinces, systemic differ-ences in postinjury care remain possible; thus, our results will need to be replicated in a geographically and polit-ically distinct region. Third, patients who required trans-port from the scene by EHS LifeFlight were excluded from the analysis. These patients tend to be more

severely injured and would likely bene�t the most from rapid transport times. It should be noted that the TC-level designations did not change over the study period, but it is likely that there were a number of major trauma care advances (both pre-hospital and in-hospital) during this time. Finally, residual confounding by unidenti�ed factors is possible, though unlikely, as there was no spa-tial autocorrelation of the model residuals.

CONCLUSION

In our study of cases of MVC and penetrating injury assigned to ground transport by the EHS system, we found that injury in a location with a PTT greater than 30 minutes to a TC was associated with poor patient outcomes. Efforts to improve communication within the trauma system and reduce prehospital time may improve outcomes for patients injured in remote locations.

Acknowledgements: Data used in this research were made available by Trauma Nova Scotia at the Nova Scotia Department of Health and Wellness. Any opinions expressed by the authors do not necessarily reflect the opinions of the Nova Scotia Department of Health and Wellness or Trauma Nova Scotia.

Af�liations: From the Department of Surgery, Dalhousie University, Halifax, NS (Tansley); the Department of Geography, Simon Fraser University, Burnaby, BC (Schuurman); the Nova Scotia Medical Examiner Service, Dartmouth, NS (Bowes); Trauma Nova Scotia, Nova Scotia Department of Health and Wellness, Halifax, NS (Erdogan, Green); the Department of Critical Care, Dalhousie University, Halifax, NS (Green); the Department of Community Health and Epidemiology, Dalhousie University, Halifax, NS (Asbridge); and the Department of Surgery, University of Calgary, Calgary, Alta. (Yanchar).


Contributors: G. Tansley, N. Schuurman, M. Bowes, M. Asbridge and N. Yanchar designed the study. G. Tansley acquired and analyzed the data, which N. Schuurman, M. Erdogan, R. Green and M. Asbridge also analyzed. G. Tansley, M. Erdogan and M. Asbridge wrote the article, which all authors reviewed and approved for publication.

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33. Brown JB, Rosengart MR, Forsythe RM, et al. Not all prehospital time is equal: in�uence of scene time on mortality. J Trauma Acute Care Surg 2016;81:93-100.

34. Røislien J, Lossius HM, Kristiansen T. Does transport time help explain the high trauma mortality rates in rural areas? New and tradi-tional predictors assessed by new and traditional statistical methods. Inj Prev 2015;21:367-73.

35. Boyle JM, Lampkin C. 2007 Motor Vehicle Occupant Safety Survey. Volume 4: Crash Injury and Emergency Medical Services Report. 2008. Available: https://www.ems.gov/pdf/research/Studies-and -Reports/MVO_Safety_Survey.pdf (accessed 2017 Dec. 8).

36. Schuurman N, Bell N, Hameed MS, et al. A model for identifying and ranking need for trauma service in nonmetropolitan regions based on injury risk and access to services. J Trauma 2008;65:54-62.

37. Crandall M, Sharp D, Unger E, et al. Trauma deserts: distance from a trauma center, transport times, and mortality from gunshot wounds in Chicago. Am J Public Health 2013;103:1103-9.

38. Rogers FB, Rittenhouse KJ, Gross BW. The golden hour in trauma: dogma or medical folklore? Injury 2015;46:525-7.

39. Buchanan IM, Coates A, Sne N. Does mode of transport confer a mortality bene�t in trauma patients? Characteristics and outcomes at an Ontario lead trauma hospital. CJEM 2016;18:363-9.

40. Zhu TH, Hollister L, Opoku D, et al. Improved survival for rural trauma patients transported by helicopter to a veri�ed trauma center: a propensity score analysis. Acad Emerg Med 2018;25:44-53.

41. Mitchell AD, Tallon JM, Sealy B. Air versus ground transport of major trauma patients to a tertiary trauma centre: a province-wide comparison using TRISS analysis. Can J Surg 2007;50:129-33.

42. McVey J, Petrie DA, Tallon JM. Air versus ground transport of the major trauma patient: a natural experiment. Prehosp Emerg Care 2010;14:45-50.

43. American College of Surgeons. Rural Trauma Team Development Course. Available at: https://www.facs.org/quality-programs/trauma/education/rttdc (accessed 2017 June 13).

44. Dennis BM, Vella MA, Gunter OL, et al. Rural Trauma Team Development Course decreases time to transfer for trauma patients. J Trauma Acute Care Surg 2016;81:632-7.

45. Malekpour M, Neuhaus N, Martin D, et al. Changes in rural trauma prehospital times following the Rural Trauma Team Development Course training. Am J Surg 2017;213:399-404.



REVIEW • REVUE

Limberg flap versus Karydakis flap for treating pilonidal sinus disease: a systematic review and meta-analysis

Background: The Limberg �ap reconstruction and the Karydakis �ap reconstruction are the 2 most used off-midline closure techniques in pilonidal sinus surgery. The current evidence is inconclusive as to which is the optimal technique. The aim of this systematic review and meta-analysis was to compare differences in outcomes between these 2 �ap-based techniques.

Methods: We identi�ed studies by a systematic literature search of the Embase, MEDLINE (PubMed), Cochrane Library and Google Scholar databases and studies selected as per the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) checklist. Only randomized controlled trials (RCTs) that com-pared the Limberg �ap (standard or modi�ed) and the Karydakis �ap were included in this review.

Results: Operative time was shorter by 7 minutes in the Karydakis group than in the Limberg group (mean difference 7.00 min, 95% con�dence interval [CI] 0.53 to 13.48). The seroma formation rate was signi�cantly higher in the Karydakis cohort (odds ratio [OR] 0.36, 95% CI 0.24 to 0.56); however, after excluding studies with a high risk of bias, the sensitivity analysis showed no signi�cant differences in seroma formation rate between the 2 techniques (OR 0.76, 95% CI 0.31 to 1.85). Other out-comes of interest showed no significant differences between the Limberg and Karydakis techniques.

Conclusion: There were no significant differences between the Limberg and Karydakis techniques. Future RCTs with strict adherence to CONSORT guidelines will further elucidate the ef�cacy of these surgical procedures.

Contexte : Les reconstructions à l’aide de lambeaux de Limberg et de Karydakis sont 2 des techniques de fermeture décalées de la ligne médiane les plus utilisées pour la chirurgie du sinus pilonidal. Les preuves actuelles ne permettent pas de conclure à la supériorité de l’une par rapport à l’autre. Le but de la présente revue systématique/méta-analyse était de comparer les différences de résultats entre ces 2 techniques de lambeaux.

Méthodes : Nous avons recensé des études au moyen d’une interrogation systéma-tique des bases de données Embase, MEDLINE (PubMed), bibliothèque Cochrane et Google Scholar et les études sélectionnées à l’aide de la liste de véri�cation PRISMA (Preferred Reporting Items for Systematic Reviews and Meta-Analyses). Seuls les essais randomisés et contrôlés (ERC) qui comparaient les lambeaux de Limberg (stan-dard ou modi�é) et de Karydakis ont été inclus dans cette revue.

Résultats : Les interventions ont duré 7 minutes de moins dans le groupe Karydakis que dans le groupe Limberg (différence moyenne 7,00 min, intervalle de con�ance [IC] de 95 % 0,53 à 13,48). Le taux de formation de séromes a été signi�cativement plus élevé dans la cohorte Karydakis (rapport ces cotes [RC] 0,36, IC de 95 % 0,24 à 0,56); par contre, après avoir exclu les études comportant un important risque de biais, l’analyse de sensibilité n’a montré aucune différence signi�cative quant au taux de for-mation de séromes entre les 2 techniques (RC 0,76, IC de 95 % 0,31 à 1,85). Les autres paramètres d’intérêt n’ont montré aucune différence signi�cative entre les techniques de Limberg et de Karydakis.

Conclusion : On n’a noté aucune différence signi�cative entre les techniques de Limberg et de Karydakis. De prochains ERC strictement conformes aux lignes direc-trices CONSORT permettront de préciser davantage l’ef�cacité de ces interventions chirurgicales.

Paschalis Gavriilidis, PhD Emil Bota, MD

Accepted Apr. 27, 2018; Published online Feb. 1, 2019

Correspondence to: P. Gavriilidis Diana, Princess of Wales Hospital Northern Lincolnshire and Goole Scartho Rd Grimsby, North East Lincolnshire DN33 2BA UK [email protected]

DOI: 10.1503/cjs.003018

flap-gavriilidis.indd 131 2019-03-18 6:47 PM

REVUE


T he reported incidence rate of pilonidal disease is 25 per 100 000 people.1 Initially, the pathogenesis was considered to be congenital.2 Today, however,

the theory that it is acquired is more widely accepted. Work by Georgios Karydakis,3,4 who highlighted 3 main factors contributing to pilonidal disease (loose hair, an external force that facilitates insertion of hair into the skin and an underlying vulnerability of natal cleft skin), had a pivotal role in this paradigm shift. The �rst 2 fac-tors are related to personal hygiene and lifestyle, and their modulation can in�uence the initiation, develop-ment and recurrence of pilonidal disease. Armstrong and Barcia5 reported that improved hygiene, an active lifestyle and hair control in the natal cleft area decreased the need for surgical procedures and resulted in faster return to work. However, the third factor can be modi�ed only surgically. The rhomboid, well-vascularized transposition Limberg �ap was �rst proposed by the Russian maxillofa-cial surgeon Alexander Limberg (1894–1974), author of the fundamental work in plastic surgery “Mathematical principles of local plastic procedures on the surface of the human body.”6,7

Patients with chronic pilonidal sinus disease are usually candidates for �ap procedures. A chronic disease can be de�ned as any condition with constant or recurrent symp-

toms observed over several months, without any indica-tion of spontaneous healing.8,9 Flap-based Limberg and Karydakis techniques both aim to achieve off-midline clo-sure of the surgical defect to �atten the natal cleft. In both techniques, methylene blue is injected in the pilonidal sinuses to guide the excision. The Karydakis technique consists of an asymmetric elliptical excision of the affected area (Fig. 1). The upper and lower poles of the ellipse are placed about 2 cm to the side of the midline. Subse-quently, after full-thickness mobilization of the contralat-eral surgical margin and �xation of the base of the �ap to the sacral fascia, the skin edges are sutured off the midline. In the modi�ed Karydakis technique, �xation to the sacral fascia is not required. The Limberg technique consists of a rhomboid-shaped excision of the affected area down to the sacral fascia (Fig. 1). The cephalic and caudal apex of the rhombus are placed about 2 cm to the side of the mid-line. A fasciocutaneous rhomboid �ap is then created and transposed so the defect can be covered without any ten-sion. In contrast, the classic Limberg �ap consists of a symmetric rhomboid excision with its apices placed on the midline.3,6–8 Karydakis reported a wound complication rate of 8% and a recurrence rate of 2% for 7471 procedures.3,4 The reported recurrence rate with the Limberg technique is 5%.8

Fig. 1. (A) Karydakis technique. (B) Limberg technique with rhomboid transposition flap.

A

B


REVIEW


The existing evidence is inconclusive as to which tech-nique is the optimal treatment for pilonidal sinus disease. To help clarify this issue, we performed a systematic review to compare the Limberg and Karydakis �ap tech-niques through meta-analysis of randomized controlled trials (RCTs). The primary outcomes investigated were wound infection, wound dehiscence and time to return to work. Hematoma formation, length of hospital stay, recurrence and patient satisfaction rate were secondary outcomes.

METHODS

This systematic review and meta-analysis was carried out in accordance with the guidelines set out in the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) checklist.

Literature search

We performed a systematic literature search of articles published in the last 20 years using the Embase, MEDLINE (PubMed), Cochrane Library and Google Scholar databases and free text and MeSH search terms (standard or modi�ed Limberg �ap; standard or modi�ed Karydakis �ap; pilonidal sinus surgery and disease; ran-domized or nonrandomized controlled trial). We also per-formed a grey literature search on the ClinicalTrials.gov Web site. We manually checked references of the retrieved articles for further analysis. Disagreements between the authors were resolved by consensus.

Study selection and inclusion and exclusion criteria

Only RCTs that compared standard or modi�ed Limberg �ap and standard or modi�ed Karydakis �ap in pilonidal sinus surgery were included in the study. All nonrandom-ized studies, reviews and narrative articles were excluded.

Data extraction and outcomes

Two reviewers (P.G. and E.B.) independently extracted the following summary data for the included studies: name of authors, patient age, operative time, wound infec-tions, wound dehiscence, hematoma, seroma, recurrences, length of hospital stay, time to return to work and patient satisfaction.

Risk of bias assessment of included studies

The 2 authors independently assessed the risk of selection bias, attrition bias, detection bias, performance bias and reporting bias for each included study. We categorized the risk of bias according to the Cochrane Handbook for System-atic Reviews of Interventions10 as high, low or unclear.

Definitions

We de�ned operative time as time from scalpel to the skin until the last skin stitch. Wound infections and seromas were reported based on clinical �ndings. Culture results and wound dehiscence (partial or complete) were de�ned by the individual study authors. We de�ned length of hos-pital stay as the number of days from the operation day until the day of discharge. Patient satisfaction was reported as described by each of the included studies’ authors.


We conducted statistical analysis using Review Manager 5.3 software (Cochrane Collaboration). Heterogeneity was assessed through the I2 test, and cut-off values of 25%, 50% and 75% were considered low, moderate and high, respectively.11 For I2 values above 25%, both �xed- and random-effects models were used and the results compared between them. In cases in which the I2 value was 25% or less, we used �xed-effects models throughout.

We analyzed dichotomous variables based on odds ratios (ORs) with 95% con�dence intervals (CIs). For the analyzed outcomes, the reference categories were selected so that OR < 1 favoured the Limberg technique. We com-bined continuous variable based on both the mean differ-ence and standardized mean difference. For studies that did not report the means and variances for the 2 groups, we estimated these values from the median, range and sample size where possible, using the technique described by Hozo and colleagues.12 In all analyses, the signi�cance level was set at p < 0.05.

Sensitivity analysis

We conducted analyses of primary and secondary out-comes using both random-effects and �xed-effect models in order to assess the impact of heterogeneity on the results. In addition, we performed subgroup analysis of the studies in which the modi�ed �ap technique was used. We did not estimate publication bias because fewer than 10 studies were included in the review.13

RESULTS

Characteristics of included studies

The search strategy yielded 93 papers. A further 3 papers were obtained via manual reference searching, for a total of 96 studies. After removal of duplicates, 87 studies remained. Their abstracts were screened and 41 studies were removed as they did not ful�ll the inclusion criteria. The full articles of the remaining 46 studies were read. Twenty-six 1-arm studies and 11 retrospective analyses were excluded, leaving 9 studies14–22 that met the criteria


REVUE


for inclusion in this systematic review and meta-analysis (Fig. 2).

A total of 1421 patients were involved in the selected studies, of whom 773 (54.4%) underwent Limberg and 648 (45.6%) underwent Karydakis �ap reconstruction. Demo-graphic characteristics were similar between the 2 groups (Table 1). All studies included patients with chronic piloni-dal disease. Five studies compared the modi�ed Limberg

�ap to the Karydakis �ap;15,17–19,21 those results were similar to those for the total sample (Table 2). Bali and col-leagues19 enrolled only patients with recurrent disease. The most commonly used excluding criteria of the RCTs were recurrent disease, American Society of Anesthesiologists physical status class 3 or greater, diabetes, drug or alcohol addiction, renal failure and immunosuppression. Patients with laterally extended sinuses unsuitable to be included in

Fig. 2. Flow diagram of literature search strategy.

Records identified through database searching

n = 93

Additional records identified through other sources

n = 3

Records after duplicates removedn = 87

Records screenedn = 87

Excluded: did not fulfill inclusion criteria n = 41

Full-text articles assessed for eligibility

n = 46

Excluded n = 37• 1-arm study n = 26• Retrospective analysis

n = 11

Studies included in qualitative synthesis

n = 9

Studies included in quantitative synthesis

(meta-analysis)n = 9


REVIEW


Tab

le 1

. Stu

dy

char

acte

rist

ics*

Stud

y/ye

ar/

coun

try

No.

of

patie

nts

Age

, yr,

mea

n

± S

D

Ope

rativ

e tim

e, m

in,

mea

n ±

SD

No.

(%) w

ith

wou

nd

infe

ctio

n

No.

(%) w

ith

wou

nd

dehi

scen

ce

No.

(%) w

ith

hem

atom

a fo

rmat

ion

No.

(%) w

ith

sero

ma

form

atio

nTy

pe o

f dra

in

Leng

th o

f ho

spita

l sta

y, d

, m

ean

± S

D

Tim

e to

retu

rn

to w

ork,

d,

mea

n ±

SD

Patie

nt

satis

fact

ion

rate

, mea

n sc

ore

± S

D

or n

o. (%

)N

o. (%

) with

re

curr

ence

Ahm

ed e

t al.,

22

2017

, Pak

ista

n75

75

32.2

± 9

.8

33.6

± 9

.7

p =

0.2

NR

NR

NR

NR

NR

Red

ivac

3.97

± 0

.71

2.

93 ±

0.6

6

p <

0.0

01

15.5

3 ±

1.2

2

13.1

3 ±

1.1

5

p <

0.0

01

NR

NR

Khan

et a

l.,21

20

16, P

akis

tan†

90

9042

.7 ±

14

43

.4 ±

16

p

= 0

.2

NR

11 (1

2)

30 (3

3)

p =

0.0

01

NR

NR

6 (7

) 20

(22)

p

= 0

.003

NR

NR

NR

NR

NR

Bal

i et a

l.,20

20

15, T

urke

y37

34

25 ±

5

24 ±

5

p =

0.2

54 ±

5

48 ±

5

p =

0.0

01

4 (1

1)

8 (2

4)

p =

0.0

01

1 (3

) 2

(6)

p =

0.6

8 (2

2)

3 (9

) p

= 0

.004

3 (8

) 4

(12)

p

= 0

.8

B-V

ak1.

44 ±

0.5

3

± 0

.75

p

= 0

.001

8 ±

1.5

17

± 1

.5

p <

0.0

01

2.88

± 0

.16

1.

43 ±

0.4

p

< 0

.001

0 (0

) 0

(0)

Toka

c et

al.,

19

2015

, Tur

key†

46

4629

.28

± 8

5

28.3

5 ±

85

p

> 0

.05

44.5

± 6

.6

42.9

± 6

.2

p >

0.0

5

3 (7

) 3

(7)

p >

0.0

5

NR

NR

NR

Plac

ed, t

ype

NR

1.06

± 0

.3

1.03

± 0

.17

p

> 0

.05

20.6

1 ±

7.8

9

23.2

9 ±

6.4

2

p <

0.0

5

NR

3 (7

) 2

(4)

p >

0.0

5

Ars

lan

et a

l.,18

20

14, T

urke

y‡96

91

26.5

± 5

.9

24.7

± 5

.1

p =

0.0

8

51.1

± 6

.8

50.9

± 7

.3

p =

0.0

8

2 (2

) 6

(7)

p <

0.0

5

2 (2

) 14

(15)

p

< 0

.001

4 (4

) 3

(3)

p =

0.9

5 (5

) 18

(20)

p

< 0

.05

Jack

son-

Prat

t1.

3 ±

0.5

1.

3 ±

0.4

p

= 0

.5

20.8

± 6

.5

19.1

± 3

.4

p =

0.7

71 (7

4)

64 (7

0)

p =

0.8

6 (6

) 10

(11)

p

= 0

.003

Ars

lan

et a

l.,18

20

14, T

urke

y†10

8

9124

.7 ±

5.1

24

.7 ±

5.1

p

= 1

.0

52.9

± 7

.3

50.9

± 7

.3

p =

0.0

8

5 (5

) 6

(7)

p =

0.3

4 (4

) 14

(15)

p

< 0

.002

3 (3

) 3

(3)

p =

1.0

8 (7

) 18

(20)

p

= 0

.002

Jack

son-

Prat

t1.

3 ±

0.4

1.

3 ±

0.4

p

= 1

.0

19.8

± 4

.6

19.1

± 3

.4

p =

1.0

84 (7

8)

64 (7

0)

p =

0.7

2 (2

) 10

(11)

p

= 0

.002

Bes

sa,17

201

3,

Egyp

t†60

60

23 ±

2.2

5

23 ±

6.2

5

p =

0.4

52 ±

3.5

33

± 3

p

= 0

.001

3 (5

) 2

(3)

p =

1.0

11 (1

8)

6 (1

0)

p =

0.0

03

NR

0 (0

) 3

(5)

p =

0.2

18 F

renc

hN

RN

R43

(72)

58

(97)

2 (3

) 1

(2)

p =

0.6

Ate

s et

al.,

16

2011

, Tur

key

134

13

4N

R50

.14

± 6

.964

2.

32 ±

8.6

4

p =

0.0

01

8 (6

) 4

(3)

14 (1

0)

8 (6

)3

(2)

1 (1

) p

= 0

.9

3 (2

) 1

(1)

p =

0.9

Plac

ed, t

ype

NR

3.80

± 1

.19

3.

40 ±

0.9

4

p =

0.0

3

8.55

± 1

.18

8.

45 ±

1.1

6

p =

0.5

3.16

± 1

.40

7.

08 ±

1.7

p

= 0

.001

9 (7

) 4

(3)

Can

et a

l.,15

20

10, T

urke

y†77

68

22 ±

5.2

5

22 ±

2.7

5

p =

0.6

52.8

± 1

6 4

0.

4 ±

2.5

p

= 0

.01

3 (4

) 3

(4)

p =

1.0

1 (1

) 1

(1)

p =

1.0

1 (1

) 1

(1)

p =

1.0

3 (4

) 3

(4)

p =

1.0

On

surg

eon’

s pr

efer

ence

4.8

± 2

.6

5.5

± 2

p

= 0

.1

21.5

± 8

.5

18.8

± 7

.7

p =

0.0

9

70 (9

1)

60 (8

8)

p =

1.0

4 (5

) 3

(4)

p =

1.0

Erso

y et

al.,

14

2009

, Tur

key

50

5027

.8 ±

6.5

25

.8 ±

6.4

NR

4 (8

) 13

(26)

p

= 0

.02

NR

NR

NR

Plac

ed, t

ype

NR

NR

14 ±

10.

25

15 ±

10

p

= 0

.3

NR

NR

Pool

ed

diff

eren

ce

(n =

142

1)

773

64

8M

D 0

.47

(9

5% C

I 0.

18 to

1.1

1)

MD

7.0

0 (9

5% C

I 0.

53 to

13.

48)

OR

0.6

(9

5% C

I 0.

41 to

1.4

)

OR

0.6

0

(95%

CI

0.20

to 1

.81)

OR

1.6

(9

5% C

I 0.

76 to

3.5

1

OR

0.3

6

(95%

CI

0.24

to 0

.56)

—M

D –

0.07

(9

5% C

I 0.

45 to

0.3

0)

MD

0.3

3

(95%

CI

0.99

to 1

.66)

OR

0.8

3

(95%

CI

0.34

to 2

.01)

OR

0.8

8

(95%

CI

0.38

to 2

.04)

CI =

con

fiden

ce in

terv

al; M

D =

mea

n di

ffer

ence

; NR

= n

ot re

port

ed; O

R =

odd

s ra

tio; S

D =

sta

ndar

d de

viat

ion.

*T

hrou

ghou

t tab

le, fi

rst v

alue

is fo

r Lim

berg

flap

, sec

ond

valu

e is

for K

aryd

akis

flap

. †M

odifi

ed L

imbe

rg fl

ap.

‡Sta

ndar

d Li

mbe

rg fl

ap.


REVUE


the fusiform excision of the Karydakis procedure were transferred to the Limberg procedure.15

Study quality

Generally, RCT quality was poor (Table 3). There was a widespread high risk of inadequate blinding and incom-plete outcome data. Allocation was adequately concealed in 516–19 of the 9 studies. Only 1 study17 adequately addressed the sequence generation and blinding of outcome asses-sors, and none of the studies performed an intention-to-treat analysis for the missing data.

Outcomes of interest

Operative time was 7 minutes shorter in the Karydakis group than in the Limberg group (mean difference

7.00 min, 95% CI 0.53 to 13.48), although heterogeneity was extremely high (I2 = 99%). The seroma formation rate was signi�cantly higher in the Karydakis group than in the Limberg group (OR 0.36, 95% CI 0.24 to 0.56, I2 = 19%) (Fig. 3). However, subgroup analysis including only studies with a low risk of bias showed no statistically signi�cant difference in seroma formation rate between the 2 tech-niques (OR 0.76, 95% CI 0.31 to 1.85, I2 = 25%) (Fig. 3). For all other outcomes (wound infection/dehiscence, hematoma, recurrence, length of hospital stay and patient satisfaction), no signi�cant differences were found between the 2 procedures (Table 2).

Sensitivity analysis

Subgroup analysis of the studies including the modi�ed Limberg �ap did not show any differences compared to the total sample of studies. The differences found in seroma formation rates disappeared when studies with high risk of bias were excluded (Fig. 3, Table 2). No dis-crepancies were found between the results produced with �xed-effects and random-effects models.

DISCUSSION

This systematic review and meta-analysis showed no remarkable differences between the Limberg and Karydakis procedures in rates of wound infection/dehiscence, hema-toma, recurrence and patient satisfaction, or length of hos-pital stay. There were only 2 differences between the 2 techniques: operative time and seroma formation rate. In the analysis, a modest difference in operative time was

Table 3. Risk of bias

Study

Type of bias; risk

Sequence generation

Allocation concealment

Incomplete outcome data Blinding

Ahmed et al.22 Unclear High High High

Khan et al.21 Unclear High High High

Bali et al.20 High High High High

Tokac et al.19 Unclear Low High High

Arslan et al.18 Unclear Low High High

Bessa17 Low Low Unclear Low

Ates et al.16 Unclear Low High High

Can et al.15 Unclear Low High High

Ersoy et al.14 High High High High

Pooled data Low 1 Low 5 Low 0 Low 1

Table 2. Outcomes of interest

OutcomeNo. of studies/

patients Estimated effect (95% CI) I2, %

Total sample

Age13–21 9/1244 MD 0.47 (–0.18 to 1.11) 0

Operative time14–19 7/1082 MD 7.00 (0.53 to 13.48) 99

Wound infection13–21 8/1182 OR 0.65 (0.41 to 1.04) 18

Wound dehiscence14–19 6/990 OR 0.60 (0.20 to 1.81) 72

Hematoma14–19 5/870 OR 1.63 (0.76 to 3.51) 0

Seroma14–20 7/1170 Peto OR 0.36 (0.24 to 0.56) 19

High-quality seroma studies14–16,19 4/604 Peto OR 0.76 (0.31 to 1.85) 25

Recurrence14–19 6/1011 OR 0.88 (0.38 to 2.04) 94

Length of hospital stay14–19 7/1112 MD –0.07 (–0.45 to 0.30) 97

Time to return to work14–19 8/1212 MD 0.33 (–0.99 to 1.66) 94

Patient satisfaction rate14,19 6/990 OR 0.83 (0.34 to 2.01) 75

Subgroup analysis, modified Limberg flap

Wound infection14,17,18,20 4/556 OR 0.91 (0.42 to 1.94) 0

Wound dehiscence14,17,18 3/464 OR 0.70 (0.13 to 3.84) 75

Seroma14,17,18,20 4/644 OR 0.33 (0.99 to 0.57) 0

Recurrence14,17,18,20 4/556 OR 0.77 (0.22 to 2.63) 48

Length of hospital stay14,17,18 3/436 MD –0.00 (–0.12 to 0.12) 45

Time to return to work14,17,18 3/436 MD 0.37 (–2.04 to 2.77) 72

CI = confidence interval; MD = mean difference; OR = odds ratio.


REVIEW


noted: the procedure time was 7 minutes shorter for the Karydakis group than for the Limberg group. In a clinical setting, however, this is unlikely to be of any importance. Previous authors have attributed this difference in operative time to 2 factors: �rst, the classic Karydakis �ap does not require extensive mobilization, and second, it does not require �xation at the sacral fascia,15,17 thereby reducing the amount of operative time required. One potential con-founding factor with this �nding may be the type of anes-thesia used in each study. However, it was not possible to retrieve this information adequately for every study, which made it impossible to determine whether anesthetic type did indeed in�uence operative time.

Seroma formation rates were slightly lower in the Limberg group than in the Karydakis group. However, a subgroup analysis including only higher-quality studies did not show differences in seroma formation rates between the 2 procedures. Arslan and colleagues18 reported the highest difference in seroma formation rate between the Limberg and Karydakis procedures (5% and 20%, respec-

tively). However, their study contained incomplete data regarding dislodgement of drains, which made it dif�cult to control for confounders. In the same study, there is con-tradictory information regarding the length of hospital stay and the duration of drainage and follow-up. Overall, the patients had a mean length of hospital stay of 1.3 (standard deviation 0.4) days and a duration of drainage of 3.1 (stan-dard deviation 1.3) days. The authors reported that some patients were discharged with their drain in situ. However, this proportion of patients was not re�ected in the study’s mean difference and standard deviation of the duration of drainage. Milone and colleagues’23 meta-analysis of the role of drainage after excision and primary closure revealed that the routine use of drains does not have a statistically signi�cant advantage regarding rates of seroma formation, infection or recurrence.

All other outcomes (wound infection, wound dehis-cence, hematoma, recurrence, length of hospital stay and patient satisfaction) in the current study were similar between the Limberg and Karydakis groups.

Fig. 3. Rate of seroma formation. Top: total sample; bottom: subgroup analysis of higher-quality studies. CI = confidence interval; SLF = standard Limberg flap; MLF = modified Limberg flap.

Can et al.,15 2010Ates et al.,16 2011Bessa,17 2013Arslan et al.,18 SLF, 2014Arslan et al.,18 MLF, 2014Bali et al.,20 2015Khan et al.,21 2016

3305836

77134

6096

1083790

313

1818

420

68134

6091913490

6.84.73.5

24.126.8

7.626.6

0.88 (0.17–4.49)2.75 (0.38–19.74)

0.13 (0.01–1.28)0.26 (0.11–0.62)0.34 (0.15–0.77)0.67 (0.14–3.14)0.29 (0.12–0.66)

0.01 0.1 1 10 100

Study/year

Total sampleLimberg

Favours (Limberg) Favours (Karydakis)

Events TotalKarydakis

Events TotalRisk of bias

Weight, % A B C D E F

–

++++++

– ––– –

– –– –– –– –

– –– –

– –– –

– –

Peto odds ratioFixed-effect Peto (95% CI)



Can et al.,15 2010Ates et al.,16 2011Bessa,17 2013Arslan et al.,18 SLF, 2014Arslan et al.,18 MLF, 2014Bali et al.,20 2015Khan et al.,21 2016

Total (95% CI)Total eventsHeterogeneity: χ2 = 3.98, df = 3 (p = 0.3); I2 = 25%Test for overall effect: Z = 0.61 (p = 0.5)

3305836

9

77134

6096

1083790

313

1818

420

11

28 67

68134

6091913490

30.320.715.5

0.00.0

33.50.0

100.0296308 0.76 (0.31–1.85)

100.0568602 0.36 (0.24–0.56)

0.88 (0.17–4.49)2.75 (0.38–19.74)

0.13 (0.01–1.28)0.26 (0.11–0.62)0.34 (0.15–0.77)0.67 (0.14–3.14)0.29 (0.12–0.66)

0.01 0.1 1 10 100

Study/year

Subgroup analysis

Limberg

Favours (Limberg) Favours (Karydakis)

Events TotalKarydakis

Events TotalRisk of bias

Weight, % A B C D E FPeto odds ratio

Fixed-effect Peto (95% CI)

–

++++++

– ––– –

– –– –– –– –

– –– –

– –– –

– –

Total (95% CI)Total eventsHeterogeneity: χ2 = 7.44, df = 6 (p = 0.3); I2 = 19%Test for overall effect: Z = 4.63 (p < 0.001)

A Random sequence generation (selection bias)B Allocation concealment (selection bias)C Blinding of outcome assessment (detection bias)

D Blinding of participants and personnel (performance bias)E Incomplete outcome data (attrition bias)F Selective reporting (reporting bias)

Risk of bias legend


REVUE


Enriquez-Navascues and colleagues24 compared differ-ent surgical techniques for chronic pilonidal disease using traditional meta-analysis. By default, traditional meta-analyses can be used only to compare 2 treatment approaches. When 3 or more treatments are to be com-pared, network meta-analysis using either Bayesian or fre-quentist statistics must be used.25

Limitations

The main limitation of this review is that, although the included studies were RCTs, most were of relatively low quality and had an unclear or high risk of bias: of the 9 studies, only 1 was graded as having a low risk of bias.17 Also, the studies’ populations were likely heterogeneous, as shown by the high I2 value in the analysis of outcome of interests. The protocols, inclusion criteria, surgical tech-nique and follow-up periods also differed, which make the �ndings not only statistically but also clinically heterogen-eous. In this context, the results of this meta-analysis must be interpreted with caution.

CONCLUSION

Even taking the study limitations into account, the Limberg and Karydakis flap operations in adequately selected patients with pilonidal disease do not show any significant differences between them. Future RCTs should have strict selection criteria, prede�ned surgical outcome measures, blind outcome assessors, common methods of outcome assessment and follow-up periods of 2–3 years, and must show strict adherence to standards recommended in the CONSORT guidelines.26 These fea-tures will help further elucidate the ef�cacy of these 2 sur-gical procedures.

Af�liations: From the Department of General and Colorectal Surgery, Diana, Princess of Wales Hospital, Northern Lincolnshire and Goole, Grimsby, UK.


Contributors: P. Gavriilidis designed the study and wrote the article. Both authors acquired and analyzed the data, reviewed the article and approved the �nal version for publication.

References

1. Lee PJ, Raniga S, Biyani DK, et al. Sacrococcygeal pilonidal disease. Colorectal Dis 2008;10:639-50, discussion 651-2.

2. Hodges RM. Pilonidal sinus. Boston Med Surg J 1880;103:485-6. 3. Karydakis GE. New approach to the problem of pilonidal disease.

Lancet 1973;2:1414-5. 4. Karydakis GE. Easy and successful treatment of pilonidal sinus after

explanation of its causative process. Aust N Z J Surg 1992;62:385-9. 5. Armstrong JH, Barcia PJ. Pilonidal sinus disease. The conservative

approach. Arch Surg 1994;129:914-7. 6. Triana RJ. Alexander A. Limberg MD. Arch Facial Plast Surg 1999;1:

226.

7. de Parades V, Bouchard D, Janier M, et al. Pilonidal sinus disease. J Visc Surg 2013;150:237-47.

8. Petersen S, Koch R, Stelzner S, et al. Primary closure techniques in chronic pilonidal sinus: a survey of the results of different surgical approaches. Dis Colon Rectum 2002;45:1458-67.

9. Søndenaa K, Andersen E, Nesrik I, et al. Patient characteristics and symptoms in chronic pilonidal sinus disease. Int J Colorectal Dis 1995; 10:39-42.

10. Higgins JPT, Greens S, editors. Cochrane handbook for systematic reviews of interventions. Version 5.1 [update March 2011]. London (UK): The Cochrane Collaboration; 2011.

11. Higgins JP, Thompson SG, Deeks JJ, et al. Measuring inconsistency in meta-analyses. BMJ 2003;327:557-60.

12. Hozo SP, Diulbegovic B, Hozo I. Estimating the mean and variance from the median, range, and the size of a sample. BMC Med Res Methodol 2005;5:13.

13. Harbord RM, Harris RJ, Sterne JA. Updated tests for small-study effects in meta-analyses. Stata J 2009;9:197-210.

14. Ersoy E, Devay AO, Aktimur R, et al. Comparison of the short-term results after Limberg and Karydakis procedures for pilonidal disease: randomized prospective analysis of 100 patients. Colorectal Dis 2009; 11:705-10.

15. Can MF, Sevinc MM, Hancerliogullari O, et al. Multicentre pro-spective randomized trial comparing modi�ed Limberg �ap transpo-sition and Karydakis �ap reconstruction in patients with sacrococcy-geal pilonidal disease. Am J Surg 2010;200:318-27.

16. Ates M, Dirican A, Savac M, et al. Short and long-term results of the Karydakis �ap versus the Limberg �ap for treating pilonidal sinus disease: a prospective randomized study. Am J Surg 2011;202: 568-73.

17. Bessa SS. Comparison of short-term results between the modi�ed Karydakis �ap and the modi�ed Limberg �ap in the management of pilonidal sinus disease: a randomised controlled trial. Dis Colon Rec-tum 2013;56:491-8.

18. Arslan K, Said Kokcam S, Koksal H, et al. Which �ap method should be preferred for the treatment of pilonidal sinus? A prospective ran-domized study. Tech Coloproctol 2014;18:29-37.

19. Tokac M, Dumlu EG, Aydin MS, et al. Comparison of modi�ed Limberg �ap and Karydakis �ap operations in pilonidal sinus sur-gery: prospective randomized study. Int Surg 2015;100:870-7.

20. Bali I, Aziret M, Sözen S, et al. Effectiveness of Limberg and Kary-dakis �ap in recurrent pilonidal sinus disease. Clinics (Sao Paulo) 2015; 70:350-5.

21. Khan KJ, Ghaffar A, Choudhry S, et al. Comparison of early out-come between modi�ed Limberg and Karydakis �ap procedures in patients with sacrococcygeal pilonidal sinus. Pak J Med Health Sci 2016;10:631-4.

22. Ahmed Z, Shahid M, Malik MS, et al. Comparison of Karydakis technique with Limberg �ap procedure for sacrococcygeal pilonidal sinus disease in terms of hospital stay and work loss. Pak Armed Forces Med J 2017;67:141-4.

23. Milone M, Di Minno MND, Musella M, et al. The role of drainage after excision and primary closure of pilonidal sinus; a meta-analysis. Tech Coloproctol 2013;17:625-30.

24. Enriquez-Navascues JM, Emparanza Jl, Alkorta M, et al. Meta-analysis of randomized controlled trials comparing different tech-niques with primary closures for chronic pilonidal sinus. Tech Colo-proctol 2014;18:863-72.

25. Palmer TM, Sterne AC, editors. Meta-analysis in Stata: an updated col-lection from the Stata Journal. 2nd ed. College Station (TX): Stata Press; 2016.

26. CONSORT 2010. Ottawa: The Consort Group. Available: www.consort-statement.org/consort-2010 (accessed 2018 Nov. 14).



DISCUSSIONS IN SURGERY • DISCUSSIONS EN CHIRURGIE

Laparoscopic colectomy: trends in implementation in Canada and globally

T he �rst study of laparoscopic colectomy (LC) was published in 1991.1 Throughout the 1990s, adoption of the technique was hampered by con-cerns regarding oncologic safety and effectiveness.2,3 This concern led to

further studies and, by 2004, randomized controlled trial evidence clearly showed that LC accelerated patients’ postoperative recovery and reduced length of hospital stay (LOS) while providing equivalent oncologic outcomes compared with open colectomy (OC).4 Fewer postoperative complications and shorter LOS imply cost savings to the health care system, which is of considerable importance in an era of rising health care expenditures across all of Canada.5

Comparisons with other high-income countries suggest a lower adoption rate of LC in Canada.6–9 The Canadian Association of General Surgeons (CAGS) formed a task force to evaluate the barriers to adoption of laparoscopic colon surgery and to propose potential intervention strategies to enhance the use of the procedure. As an initial step, the task force was asked to complete a literature review of existing publications on this topic to identify patterns of implementation and barriers to adoption in Canada and other countries.

UPTAKE OF LAPAROSCOPIC COLECTOMY IN CANADA

Population-based data from Ontario show that between 2002 and 2009, the proportion of elective LC increased from 13% to 37%.10 In British Columbia, the proportion of patients with colon cancer undergoing LC increased from 2% to 25% between 2003 and 2008.11 Using population-level data, Hoogerboord and colleagues12 showed that from 2004 to 2014, the pan-Canadian rate of LC for elective surgeries increased from 9% to 52%. Provincial utilization rates ranged from 11% in Newfoundland to 60% in British Columbia.12 The rate of LC in Canada was lower than in South Korea, the Netherlands and the United States, but higher than in the United Kingdom, Norway and Sweden (Fig. 1).6–9

PREDICTORS OF UPTAKE OF LAPAROSCOPIC COLECTOMY

Compared with OC, the laparoscopic approach is technically more dif�cult because of the absence of tactile feedback, operating with 2-dimensional vision, and limited degrees of manoeuvrability of instruments. The number of cases required to complete the learning curve for LC varies between 30

Marius Hoogerboord, MB ChB, MSc James Ellsmere, MD, MS Antonio Caycedo-Marulanda, MD, MSc Carl Brown, MD, MSc Shiva Jayaraman, MD, MESc David Urbach, MD, MSc Sean Cleary, MD, MSc, MPH

Accepted May 18, 2018

Correspondence to: S. Cleary Department of Surgery, Mayo 12-11D Mayo Clinic 200 First St SW Rochester MN 55905 [email protected]

DOI: 10.1503.cjs.003118

Comparisons with other high-income countries suggest that Canada has been slower to adopt laparoscopic colectomy (LC). The Canadian Association of General Surgeons sought to evaluate the barriers to adoption of laparoscopic colon surgery and to propose potential intervention strategies to enhance the use of the procedure. Given the clinical bene�ts of laparoscopic surgery for patients, the increasing needs for surgical care and the desire of Canadian gen-eral surgeons to advance their specialty and enhance the care of their patients, it is an important priority to improve the utilization of LC.

SUMMARY

trends-cleary.indd 139 2019-03-19 12:57 PM

DISCUSSIONS EN CHIRURGIE


and 70 and depends on the extent of formal training in laparoscopic surgery, exposure to other advanced laparo-scopic techniques, dexterity and ability to perform delicate manoeuvres with 2-dimensional vision.13

The Canadian health care system can present unique challenges to the adoption of new technology and min-imally invasive surgical techniques. In the survey by Moloo and colleagues,14 limitations in operating room time alloca-tion were cited by 55% of surgeons as a barrier to LC.Since most hospitals in Canada function with a global operating budget and volume-based and/or quality-based hospital reimbursement is limited in extent and impact, the introduction of new technology and equipment into Can-adian operating rooms has been much slower than in other countries. Restrictions in capital and operating budgets have limited investments in permanent and single-use lapa-roscopic equipment. Approximately 25% of Canadian sur-geons felt that they did not have adequate laparoscopic equipment to perform LC at their hospitals.14

POTENTIAL STRATEGIES TO INCREASE THE USE OF LAPAROSCOPIC COLECTOMY

Effective strategies to increase use of LC may include greater emphasis on advanced laparoscopic training during resi-

dency, more minimally invasive surgery fellowships, and intensive hands-on training courses and mentorship pro-grams for established surgeons (Box 1).14–16 Also, dissemina-tion of practice guidelines and consensus statements, practice audits and feedback, and the development of care algorithms and clinical care guidelines have been shown to positively affect change in surgical practice and institutional culture.17

Several methods for teaching advanced laparoscopic techniques to established surgeons exist, including short (weekend) courses, comprehensive programs consisting of didactic sessions, cadaver laboratories, and observation of live surgeries and simulation platforms (e.g., minimally invasive surgery trainer).16 The optimal model for training and dissemination of advanced laparoscopic surgery into community practice has not been established. However,

Fig. 1. Proportion of laparoscopic colectomy in Canada compared with other high-income countries.

0

20

40

60

2004 2005 2006 2007 2008 2009 2010 2011 2012 2013 2014

Pro

port

ion

lapa

rosc

opic

col

ecto

my

(%)

Year

South Korea

Canada

United Kingdom

NetherlandsUnited States

Norway

Sweden

Sweden

United States

South Korea

Norway

Canada United Kingdom

Netherlands

Box 1: Strategies to increase use of laparoscopic colectomy in CanadaResident training

• Increase exposure to advanced laparoscopic proceduresFellowship training

• Increase number of minimally invasive surgery fellowship programsPractising surgeons

• Intensive hands-on courses• Postcourse mentorship programs


DISCUSSIONS IN SURGERY


adoption is enhanced when mentorship by laparoscopic experts forms part of a training program.18 A team-based approach, where training also includes operating room nurses, has been shown to be more effective in establishing successful laparoscopic programs.19

CONCLUSION

The rate of LC in Canada appears to be lower than in other comparable high-income countries, including South Korea, the Netherlands and the United States, with sig-ni� cant variability across the country. This variability in use of LC indicates a variety of barriers, some of which may be speci�c to the Canadian health care environment; a review of the literature includes surgeon-related factors (e.g., completing a steep learning curve, access to training and mentorship) and regional health system factors (e.g., �nancial constraints, operating room time and resource limitations, hospital and physician reimbursement models, and access to technology). Given the clinical bene�ts of laparoscopic surgery for patients, the increasing needs for surgical care and the desire of Canadian general surgeons to advance their specialty and enhance the care of their patients, it is an important priority to improve the utiliza-tion of LC. Given the complexities of these issues and the importance to patients and the profession, the Canadian Association of General Surgeons has assembled a task force to study the barriers to LC and propose priorities and initiatives to address the identified barriers and increase the use of LC across the country.

Af�liations: From the Division of General Surgery, Department of Surgery, Dalhousie University, Halifax, NS (Hoogerboord, Ellsmere); the Division of General Surgery, Department of Surgery, Northern Ontario School of Medicine, Sudbury, Ont. (Caycedo-Marulanda); the Division of General Surgery, Department of Surgery, University of British Columbia, Vancouver, BC (Brown); the Division of General Surgery, Department of Surgery, University of Toronto, Toronto, Ont. (Jayaraman, Urbach); the Department of Surgery, St. Joseph’s Health Centre, Toronto, Ont. (Jayaraman); the Department of Surgery, Women’s College Hospital, Toronto, Ont. (Urbach); and the Department of Sur-gery, Mayo Clinic, Rochester, MN (Cleary).

Competing interests: S. Jayaraman declares consulting and speaker fees from Ethicon, Olympus and Baxter; consulting fees from IPSEN, Pendopharm and P�zer; and speaker fees from Celgene, outside the submitted work. S. Cleary declares honoraria for consulting activities from Ethicon, Erbe and Olympus as well as steering committee activ-ities for AstraZeneca, outside the submitted work. No other competing interests declared.

Contributors: All authors contributed substantially to the conception, writing and revision of this article and approved the �nal version for publication.

References

1. Jacobs M, Verdeja JC, Goldstein HS. Minimally invasive colon resection (laparoscopic colectomy). Surg Laparosc Endosc 1991; 1:144-50.

2. Alexander RJ, Jaques BC, Mitchell KG. Laparoscopically assisted colectomy and wound recurrence. Lancet 1993;341:249-50.

3. O’Rourke N, Price PM, Kelly S, et al. Tumour inoculation during laparoscopy. Lancet 1993;342:368.

4. Nelson H, Sargent D, Wieand HS, et al.; Clinical Outcomes of Sur-gical Therapy Study Group. A comparison of laparoscopically assisted and open colectomy for colon cancer. N Engl J Med 2004;350:2050-9.

5. National Health Expenditure Trends. 1975-2015. Ottawa, Ontario: Canadian Institute for Health Information; 2015.

6. Moghadamyeghaneh Z, Carmichael JC, Mills S, et al. Variations in laparoscopic colectomy utilization in the United States. Dis Colon Rectum 2015;58:950-6.

7. Taylor EF, Thomas JD, Whitehouse LE, et al. Population-based study of laparoscopic colorectal cancer surgery 2006-2008. Br J Surg 2013;100:553-60.

8. Kolfschoten NE, van Leersum NJ, Gooiker GA, et al. Successful and safe introduction of laparoscopic colorectal cancer surgery in Dutch hospitals. Ann Surg 2013;257:916-21.

9. Park SJ, Lee KY, Lee S. Laparoscopic surgery for colorectal cancer in Korea: nationwide data from 2008-2013. J Minim Invasive Surg 2015;18:39-43.

10. Chan BP, Gomes T, Musselman RP, et al. Trends in colon cancer surgery in Ontario: 2002-2009. Colorectal Dis 2012;14:e708-12.

11. Aslani N, Lobo-Prabhu K, Heidary B, et al. Outcomes of laparo-scopic colon cancer surgery in a population-based cohort in British Columbia: Are they as good as the clinical trials? Am J Surg 2012;204:411-5.

12. Hoogerboord CM, Levy AR, Hu M, et al. Uptake of elective laparo-scopic colectomy for colon cancer in Canada from 2004/05 to 2014/15: a descriptive analysis. CMAJ Open 2018;6:E384-90.

13. Schlachta CM, Mamazza J, Seshadri PA, et al. De�ning a learning curve for laparoscopic colorectal resections. Dis Colon Rectum 2001; 44:217-22.

14. Moloo H, Haggar F, Martel G, et al. The adoption of laparoscopic colorectal surgery: a national survey of general surgeons. Can J Surg 2009;52:455-62.

15. Dominguez EP, Barrat C, Shaffer L, et al. Minimally invasive surgery adoption into an established surgical practice: impact of a fellowship-trained colleague. Surg Endosc 2013;27:1267-72.

16. Birch DW, Sample C, Gupta R. The impact of a comprehensive course in advanced minimal access surgery on surgeon practice. Can J Surg 2007;50:9-12.

17. Urbach DR. Closing in on surgical practice variations. Ann Surg 2014;259:628-9.

18. Birch DW, Bonjer HJ, Crossley C, et al. Canadian consensus confer-ence on the development of training and practice standards in advanced minimally invasive surgery: Edmonton, Alta., Jun. 1, 2007. Can J Surg 2009;52:321-7.

19. Birch DW, Misra M, Farrokhyar F. The feasibility of introducing advanced minimally invasive surgery into surgical practice. Can J Surg 2007;50:256-60.



DISCUSSIONS IN SURGERY • DISCUSSIONS EN CHIRURGIE

Deployment of second-generation resuscitative endovascular balloon occlusion of the aorta for unresponsive hypotension in a polytrauma patient

R esuscitative endovascular balloon occlusion of the aorta (REBOA) is a less invasive method of aortic occlusion as a temporary measure to manage noncompressible hemorrhagic control. This technique con-

sists of a multitude of steps, including arterial access, sheath placement, catheter insertion, balloon introduction, position selection, balloon in�ation, balloon de�ation, and removal. Although the technique has remained rela-tively similar, the manufacturing of the device has changed over time. The �rst generation REBOA used a 12-French sheath to deliver the balloon within the common femoral artery (CFA) requiring surgical repair. The introduction of the new second-generation ER-REBOA catheter allows for the use of a 7-French sheath and a built-in-pressure monitoring lumen; it is also guidewire-free and can be used in a �uoroscopy-free setting.1 These improvements allow for a further reduction in time to occlusion. We describe the use of a second-generation ER-REBOA in an unresponsive, hypotensive polytrauma patient.

CASE

An otherwise healthy 61-year-old man presented to the emergency depart-ment at our level-I trauma centre after a 2-storey fall. The patient presented with a Glasgow Coma Scale score of 13, which deteriorated to 9, resulting in immediate intubation for airway protection. The patient was tachycardic with a heart rate of 112, and was initially normotensive with a systolic blood pressure (SBP) of 144 mm Hg, which rapidly declined to 85 mm Hg. The patient was treated according to Advanced Trauma Life Support principles. A chest radiograph showed right subcutaneous emphysema with no evidence of thoracic bleeding. As a result, a right-sided chest tube was inserted. A pel-vic radiograph showed an unstable pelvis fracture with a left vertical shear

Tiffany Paradis, BSc Omar Bekdache, MD David Bracco, MD Jeremy Grushka, MD, MSc Tarek Razek, MD David Lasry, MD Andrew Beckett, MD, MSc

Accepted July 27, 2018

Correspondence to: O. Bekdache Montreal General Hospital 1650 Cedar Ave. Montreal QC H3G 1A4 [email protected]

DOI: 10.1503/cjs.007618

Noncompressible hemorrhagic control remains one of the most challenging areas in damage control medicine and continues to be a leading cause of pre-ventable death. For decades, emergency thoracotomy or laparotomy and aortic cross clamping have remained the gold standard intervention. Recently, there has been a movement toward less invasive techniques for noncompressible hemorrhagic control, such as resuscitative endovascular balloon occlusion of the aorta (REBOA). The REBOA technique involves in�ation of an endovascu-lar balloon within the abdominal aorta proximal to the vascular injury to tem-porarily inhibit bleeding. Although the literature is robust on this new tech-nique, skepticism remains about whether REBOA is superior to aortic cross clamping, as it has been associated with complications including organ and limb ischemia, limb amputation, femoral aneurysm, and thrombosis.

SUMMARY

generation-paradis.indd 142 2019-03-19 10:51 AM

DISCUSSIONS IN SURGERY


pattern. Therefore, a pelvic binder was placed. As the focused abdominal for sonography in trauma (FAST) was positive, our massive transfusion protocol was initiated. The patient was immediately given 2 units of packed red blood cells (PRBCs), and an emergency trauma laparot-omy was scheduled.

Despite aggressive resuscitation the patient was a transient responder with an elevated lactate of 5 mmol/L. Owing to deteriorating hemodynamics, the patient was transported to the operating room; during the preparation, we decided to obtain noncompressible hemorrhage control using REBOA at zone 3 (bifurcation of the abdominal aorta). Zone 3 was selected as the patient had severely displaced pelvic fractures, which we believed to be the primary source of hemorrhage. Zone 1 was not selected, as we wanted to avoid complications of visceral and spinal ischemia. During preparation for emergency laparotomy, the trauma team leader percuta-neously accessed the left CFA by ultrasound guidance using an 18-G needle.

To ensure accessibility of the left CFA, a notch was placed in the pelvic binder. External landmarking was used to determine the placement site for balloon occlusion. A 7-French arterial inducer (AVANTI) was inserted into the left CFA and we advanced the ER-REBOA (Prytime Med-ical) for zone 3 aortic deployment.1

The balloon was deployed at a catheter length of 27 cm. External landmarking and catheter length were used to estimate zone 3 balloon location. As this was our �rst REBOA case, we also veri�ed location using �uoros-copy before in�ation with 4 mL of saline. Time from

ultrasound-guided percutaneous access to balloon in�a-tion in zone 3 was 43 seconds. There was an immediate increase in systolic blood pressure from 90 mm Hg to 125 mm Hg after balloon deployment (Fig. 1). Total bal-loon occlusion time was 36 minutes while surgical hem-orrhage was controlled. The balloon was safely de�ated by the trauma surgeon in conjunction with the anesthesia team and con�rmed through �uoroscopy before removal (Fig. 2). During the trauma laparotomy, the following injuries were identi�ed: grade III splenic laceration, grade I liver injury, blunt hollow viscus injury, and a zone 3 ret-roperitoneal hematoma. Surgical management consisted of splenectomy, preperitoneal pelvic packing, small bowel repair and liver packing. Intraoperative angioemboliza-tion was performed through access via the right groin to embolize bleeding branches of the iliac artery. The patient received an additional 4 units of PRBCs, 5 units of platelets and 4 units of fresh frozen plasma. Estimated blood loss was 2 L. As the patient was initially unrespon-sive to hemodynamic management, it was our clinical judgment that if the ER-REBOA was not placed, the patient would have required additional blood products and may have required resuscitative thoracotomy.

Distal Doppler pulses were obtained for dorsalis pedis in both limbs to con�rm patency of lower extremity arter-ies. We placed a temporary abdominal closure, and the patient was then transferred to the intensive care unit (ICU) for ongoing resuscitation.

The femoral sheath was removed in the ICU 17 hours after the procedure, and CFA imaging was per-formed using CT angiogram to con�rm adequate limb

Fig. 1. Intraoperative blood pressure measurement in a 61-year-old man; systolic blood pressure (mm Hg) is plotted on the y axis and time on the x axis. Preoperative measurements were those taken in the trauma bay upon arrival. Arrow 1 is time of catheter insertion (16:08:15), arrow 2 is time of balloon inflation (16:09), and arrow 3 is time of balloon deflation (16:45), at which point the catheter was also removed. The sheath was left in place until postoperative day 1, at which point it was removed in the intensive care unit.

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15:20 15:26 15:28 15:29 16:05 16:10 16:15 16:20 16:25 16:30 16:35 16:40 16:45 16:50 16:55 17:00 17:05 17:10 17:15 17:20 17:25

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Intraoperative blood pressure measurementPreoperative measurements


DISCUSSIONS EN CHIRURGIE


perfusion and absence of complications. The patient’s postoperative course of 48 hours proved to be unevent-ful; he experienced no limb ischemia or reperfusion injuries. At 30 days postadmission, the patient was on the trauma service ward for ongoing management of an open abdomen.

DISCUSSION

Noncompressible hemorrhage is a leading cause of pre-ventable death in patients with traumatic injury.2 The concept of resuscitative endovascular aortic occlusion as a method of hemorrhage control is not novel.3 This technique was originally reported by Hughes during the Korean war, where the procedure was performed on 2 critically wounded soldiers.4 Although the first patients succumbed to their wounds, the potential for use as a resuscitative method was con�rmed and is now showing promise to improve survival through tempo-rary occlusion of the aorta, thereby restoring perfusion to the brain and heart.5 The technique of REBOA in trauma with off-label endovascular catheters has been used, but large arterial sheaths of up to 14-French has limited their use.5

To our knowledge, this is the �rst report of use of the ER-REBOA catheter in Canada since it obtained Health Canada approval in October 2017, and we have since placed a second device for a junctional hemor-rhage due to a gunshot wound to the proximal super�-cial femoral artery. The ER-REBOA system offers a compliant aortic balloon on 5-French catheters with an

integral arterial line. The device also does not require a guidewire, which allows for very rapid deployment.1 Although frequently used in the United States and Japan, REBOA has yet to be widely adopted in Canad-ian trauma centres.

Multidisciplinary training of the trauma program, emergency, and operating room staff allowed for rapid mobilization for REBOA placement. This training included practitioners undergoing external REBOA courses, hands-on practice in gaining CFA access with interventional radiology, didactic presentations and trauma team simulations.

CONCLUSION

The endovascular revolution has now placed an easily and safely deployable REBOA device in the hands of non-interventional radiologists and nonvascular surgeons. Our case report shows that with multidisciplinary training, ER-REBOA can be performed successfully in Canadian trauma centres; however, further studies are needed to fully de�ne its role in Canadian Trauma Systems.

Acknowledgements: The authors acknowledge the generous sup-port of the Montreal General Hospital Foundation, and the McGill University Health Centre Emergency Medicine and Adult Trauma Programs.

Af�liations: From the Faculty of Medicine McGill University, Mon-treal, Que. (Paradis); the Trauma Program, McGill University Health Centre, Montreal, Que, (Bekdache, Bracco, Grushka, Razek, Lasry, Beckett); and the Royal Canadian Medical Services, Montreal, Que. (Beckett).


Contributors: All authors contributed substantially to the conception, writing and revision of this article and approved the �nal version for publication.

References

1. ER REBOA. Prytime Medical — The REBOA Company; 2017. Available: http://prytimemedical.com/contact-us/ (accessed 2018 May 12).

2. Moore LJ, Brenner M, Kozar RA, et al. Implementation of resus-citative endovascular balloon occlusion of the aorta as an alterna-tive to resuscitative thoracotomy for noncompressible truncal hemorrhage. J Trauma Acute Care Surg 2015;79:523-530; discus-sion 530-2.

3. Brenner M, Inaba K, Aiol� A, et al. Resuscitative endovascular bal-loon occlusion of the aorta and resuscitative thoracotomy in select patients with hemorrhagic shock: early results from the American Association for the Surgery of Trauma’s Aortic Occlusion in Resusci-tation for Trauma and Acute Care Surgery Registry. J Am Coll Surg 2018;226:730-40.

4. Hughes CW. Use of an intra-aortic balloon catheter tamponade for controlling intra-abdominal hemorrhage in man. Surgery 1954; 36:65-8.

5. Smith SA, Hilsden R, Beckett A, et al. The future of resuscitative endovascular balloon occlusion in combat operations. J R Army Med Corps 2017; doi: 10.1136/jramc-2017-000788.

Fig. 2. Fluoroscopy image showing balloon deflation (white arrow) in zone 3. The inferior vena cava filter can be seen on the right (black arrow), in addition to pelvic packing material (dotted arrow).


The Division of General Surgery at the Vernon Jubilee Hospital in Vernon, BC, seeks a full-time general surgeon to replace one member of a group of 6 FTE in 2020. Applicants must be FRCSC certi�ed or eligible in 2020. Proof of ACLS, CMPA, and BC licensure mandatory. Post-Fellowship training is not required. Applicants must be competent and willing to perform upper/lower endoscopy and surgeries including but not limited to: breast, colorectal, skin malignancies, gallbladder disease, hernias, acute GI conditions, and minor procedures.

VJH is a close-knit, 160-bed facility with a patient care tower opened in 2011 with 5 operating rooms, a 6-bed ICU, FRCP anesthesia, pediatrics, internal medicine, medical oncology, CT, US, nuclear medicine, and MRI (opening 2019). VJH is af�liated with the University of British Columbia and is a training site for the UBC General Surgery Postgraduate program and the UBC undergraduate medical program. Vernon is a jewel of a small city, dotted by lakes, cradled by mountains in the North Okanagan Valley, surrounded by orchards, vineyards, farms, and a world-class ski hill: Silver Star/Sovereign Lakes. Other surgical specialties at VJH include ENT, obstetrics-gynecology, ophthalmology, orthopedics, and urology.

Applications should consist of a current curriculum vitae and three letters of reference for receipt by July 31, 2019. Incomplete applications will not be considered. Successful short-listed candidates will be contacted in August 2019 for interviews. Please submit applications to:

Dr. H. HwangVernon Jubilee Hospital

2101 32 St., Vernon, BC V1T 5L2Email [email protected]

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Otolaryngologist OpportunityTrail, BC

An opportunity is available for an Otolaryngologist to practice at Kootenay Boundary Regional Hospital in Trail.

Come and enjoy what it feels like to be appreciated and valued while providing service to the wonderful community.

*Also inquire if interested in practicing in an urban community.*

[email protected]

Rural Incentives ~ Relocation Support ~ Diverse Practice Options

www.betterhere.caFor more information & to apply email:

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