card minimum wages and employment

8/7/2019 Card Minimum Wages and Employment

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Minimum Wages and Employment: A Case Study of the Fast-Food Industry in New Jersey and Pennsylvania

On April 1, 1992, New Jersey's minimum wage rose from $4.25 to $5.05 perhour. To evaluate the impact of the law we surveyed 410 fast-food restaurants inNew Jersey and eastern Pennsylvania before and after the rise. Comparisons ofemployment growth at stores in New Jersey and Pennsylvania (where theminimum wage was constant) provide simple estimates of the effect of the higherminimum wage. We also compare employment changes at stores in New Jerseythat were initially paying high wages (above $5) to the changes at lower-wagestores. We find no indication that the rise in the minimum wage reducedemployment. (JEL 530, 523)

How do em ployers in a low-wage labor cent studies that rely on a similar com para-market respond to an increase in the mini- tive methodology have failed to detect amum wage? T he prediction f rom conven- negative employment effect of higher mini-tional econom ic theory is unam biguou s: a mum wages. Analyses of th e 1990-1991 in-rise in the minimum wage leads perfectly creases in the federal minimum wagecompetitive employers to cut employment (Lawrence F. Katz an d K rueger, 1992; Card,(George J. Stigler, 1946). Although studies 1992a) and of an earlier increase in thein the 1970's based on agg regate teenage minimum wage in California (Card, 1992b)employment rates usually confirmed this find no adverse employment impact. A studyprediction,' earlier studi es based o n com- of minimum-wage floors in Britain (Step henparisons of employment at affected and un- Machin and Alan Manning, 1994) reaches aaffected establishments often did not (e.g., similar conclusion.Richard A. Lester, 1960, 1964). Several re- This pap er presents new evidence on theeffect of minimum wages on establishment-level employment outcomes. W e analyze th eexperiences of 410 fast-food restaurants in

*Department of Economics, Princeton University, New Jersey and Pennsylvania following thePrinceton, NJ 08544. We are grateful to the Institute increase in New Jersey's minimum wagefor Research on Poverty, University of Wisconsin, for from $4.25 to $5.05 per hour. Comparisonspartial financial support. Thanks to Orley Ashenfelter, of employment, wages, and prices at storesCharles Brown, Richard Lester, Gary Solon, twoanonymous referees, and seminar participants at in New Jersey and Pennsylvania before andPrinceton, Michigan State, Texas A&M, University of after the rise offer a simple method forMichigan, university of Pennsylvania, ~ ni ve rs it J of evaluating th e effects of the-minimum wage.Chicago, and the NBER for comments and sugges- ~~~~~~i~~~~ within N~~ jerseybetweentions. We also acknowledge the expert research assis-tance of Susan Belden, Chris Burris, Geraldine Harris, high-wage payingand Jonathan Orszag. than the new minimum rate prior to its's ee Charles Brown et al. (1982,1983) for surveys of effective date) and other stores provide anthis literature. A recent update (Allison J. Wellington, alternative estimate of the impact of the1991) concludes that the employment effects of the new laweminimum wage are negative but small: a 10-percentincrease in the minimum is estimated to lower teenage In addition to the simplicity of our empir-employment rates by 0.06 percentage points. ical methodology, several other features of

772


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773V O L . 84 NO. 4 CARD AND KRUEGER: MINIMUM WAGE AND EMPLOYMENTthe New Jersey law and our data set arealso significant. First, the rise in the mini-mum wage occurred during a recession. Theincrease had been legislated two years ear-lier when the state economy was relativelyhealthy. By the time of the actual increase,the unemployment rate in New Jersey hadrisen substantially and last-minute politicalaction almost succeeded in reducing theminimum-wage increase. It is unlikely thatthe effects of the higher minimum wagewere obscured by a rising tide of generaleconomic conditions.Second, New Jersey is a relatively smallstate with an economy that is closely linkedto nearby states. We believe that a controlgroup of fast-food stores in eastern Pennsyl-vania forms a natural basis for comparisonwith the experiences of restaurants in NewJersey. Wage variation across stores in NewJersey, however, allows us to compare theexperiences of high-wage and low-wagestores within New Jersey and to test thevalidity of the Pennsylvania control group.Moreover, since seasonal patterns of em-ployment are similar in New Jersey andeastern Pennsylvania, as well as acrosshigh- and low-wage stores within New Jer-sey, our comparative methodology effec-tively "differences out" any. seasonal em-ployment effects.

Third, we successfully followed nearly 100percent of stores from a first wave of inter-views conducted just before the rise in theminimum wage (in February and March1992) to a second wave conducted 7-8months after (in November and December1992). We have complete information onstore closings and take account of employ-ment changes at the closed stores in ouranalyses. We therefore measure the overalleffect of the minimum wage on averageemployment, and not simply its effect onsurviving establishments.-Our analysis of employment trends atstores that were open for business beforethe increase in the minimum wage ignoresany potential effect of minimum wages onthe rate of new store openings. To assessthe likely magnitude of this effect we relatestate-specific growth rates in the number ofMcDonald's fast-food outlets between 1986

and 1991 to measures of the relative mini-mum wage in each state.

I. The New Jersey LawA bill signed into law in November 1989raised the federal minimum wage from $3.35per hour to $3.80 effective April 1, 1990,with a further increase to $4.25 per hour onApril 1, 1991. In early 1990 the New Jerseylegislature went one step further, enactingparallel increases in the state minimum wagefor 1990 and 1991 and an increase to $5.05per hour effective April 1, 1992. The sched-uled 1992 increase gave New Jersey the

highest state minimum wage in the countryand was strongly opposed by business lead-ers in the state (see Bureau of NationalAffairs, Daily Labor Report, 5 May 1990).In the two years between passage of the$5.05 minimum wage and its effective date,New Jersey's economy slipped into reces-sion. Concerned with the potentially ad-verse impact of a higher minimum wage, thestate legislature voted in March 1992 tophase in the 80-cent increase over two years.The vote fell just short of the margin re-quired to override a gubernatorial veto, andthe Governor allowed the $5.05 rate to gointo effect on April 1 before vetoing thetwo-step legislation. Faced with the prospectof having to roll back wages for minimum-wage earners, the legislature dropped theissue. Despite a strong last-minute chal-lenge, the $5.05 minimum rate took effectas originally planned.

11. Sample Design and EvaluationEarly in 1992 we decided to evaluate theimpending increase in the New Jersey mini-mum wage by surveying fast-food restau-rants in New Jersey and eastern Pennsylva-niae2 Our choice of the fast-food industrywas driven by several factors. First, fast-foodstores are a leading employer of low-wageworkers: in 1987, franchised restaurants em-

2At the time we were uncertain whether the $5.05rate would go into effect or be overridden.


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THE AMERICAN ECONOMIC REVIEW

Waue I , February 15-March 4, 1992:Number of stores in sample frame:a Number of refusals: Number interviewed: Response rate (percentage):

Wace 2, Nocember 5 - D ecember 31, 1992:Number of stores in sample frame: Number closed: Number under rennovation: Number temporarily closed:' Number of refusals: Number intervie~ed:~

A1l

4736341086.7

4106221399

SEPTEMBER 1994

Stores in:NJ PA

364 10933 30331 7990.9 72.5

331 795 12 02 01 0321 78

aStores with working phone numbers only; 29 stores in original sample frame haddisconnected phone numbers.'~ncludes one store closed because of highway construction and one store closedbecause of a fire.'Includes 371 phone interviews and 28 personal interviews of stores that refused aninitial request for a phone interview.

ployed 25 percent of all workers in therestaurant industry (see U.S. Department ofCommerce, 1990 table 13). Second, fast-foodrestaurants comply with minimum-wage reg-ulations and would be expected to raisewages in response to a rise in the minimumwage. Third, the job requirements andproducts of fast-food restaurants are rela-tively homogeneous, making it easier to ob-tain reliable measures of employment,wages, and product prices. The absence oftips greatly simplifies the measurement ofwages in the industry. Fourth, it is relativelyeasy to construct a sample frame of fran-chised restaurants. Finally, past experience(Katz and Krueger, 1992) suggested thatfast-food restaurants have high responserates to telephone survey^.^Based on these considerations we con-structed a sample frame of fast-food restau-

3 ~ na pilot survey Katz and Krueger (1992) obtainedvery low response rates from McDonald's restaurants.For this reason, McDonald's restaurants were excludedfrom Katz and Krueger's and our sample frames.

rants in New Jersey and eastern Pennsylva-nia from the Burger King, KFC, Wendy's,and Roy Rogers chain^.^ The first wave ofthe survey was conducted by telephone inlate February and early March 1992, a littleover a month before the scheduled increasein New Jersey's minimum wage. The surveyincluded questions on employment, startingwages, prices, and other store characteris-t i c ~ . ~Table 1 shows that 473 stores in our sam-ple frame had working telephone numberswhen we tried to reach them in February-March 1992. Restaurants were called asmany as nine times to elicit a response. Weobtained completed interviews (with someitem nonresponse) from 410 of the restau-rants, for an overall response rate of 87percent. The response rate was higher inNew Jersey (91 percent) than in Pennsylva-

4 ~ h esample was derived from white-pages tele-phone listings for New Jersey and Pennsylvania as ofFebruary 1992.'copies of the questionnaires used in both waves ofthe survey are available from the authors upon request.


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775VOL.84 NO. 4 C A mAND KRUEGER: MINIiiMUM WAGE AND EMPLOYMENT

nia (72.5 percent) because our interviewermade fewer call-backs to nonrespondents inPenn~ylvania.~In the analysis below we in-vestigate possible biases associated with thedegree of difficulty in obtaining the first-wave interview.The second wave of the survey was con-ducted in November and December 1992,about eight months after the minimum-wageincrease. Only the 410 stores that re-sponded in the first wave were contacted inthe second round of interviews. We success-fully interviewed 371 (90 percent) of thesestores by phone in November 1992. Becauseof a concern that nonresponding restaurantsmight have closed, we hired an interviewerto drive to each of the 39 nonrespondentsand determine whether the store was stillopen, and to conduct a personal interview ifpossible. The interviewer discovered that sixrestaurants were permanently closed, twowere temporarily closed (one because of afire, one because of road construction), andtwo were under renovation.' Of the 29 storesopen for business, all but one granted arequest for a personal interview. As a re-sult, we have second-wave interview datafor 99.8 percent of the restaurants that re-sponded in the first wave of the survey, andinformation on closure status for 100 per-cent of the sample.Table 2 presents the means for severalkey variables in our data set, averaged overthe subset of nonmissing responses for eachvariable. In constructing the means, employ-ment in wave 2 is set to 0 for the perma-

6~esponserates per call-back were almost identicalin the two states. Among New Jersey stores, 44.5percent responded on the first call, and 72.0 percentresponded after at most two call-backs. Among Penn-sylvania stores 42.2 percent responded on the first call,and 71.6 percent responded after at most two call-backs.7 ~ sof April 1993 the store closed because of roadconstruction and one of the stores closed for renova-tion had reopened. The store closed by fire was openwhen our telephone interviewer called in November1992 but refused the interview. By the time of thefollow-up personal interview a mall fire had closed thestore.

nently closed stores but is treated as missingfor the temporarily closed stores. (Full-time-equivalent [FTE] employment was cal-culated as the number of full-time workers[including managers] plus 0.5 times thenumber of part-time workers.)' Means arepresented separately for stores in New Jer-sey and Pennsylvania, along with t statisticsfor the null hypothesis that the means areequal in the two states.Rows la-e show the distribution of storesby chain and ownership status (company-owned versus franchisee-owned). TheBurger King, Roy Rogers, and Wendy'sstores in our sample have similar averagefood prices, store hours, and employmentlevels. The KFC stores are smaller and areopen for fewer hours. They also offer amore expensive main course than stores inthe other chains (chicken vs, hamburgers).In wave 1, average employment was 23.3full-time equivalent workers per store inPennsylvania, compared with an average of20.4 in New Jersey. Starting wages werevery similar among stores in the two states,although the average price of a "full meal"(medium soda, small fries, and an entree)was significantly higher in New Jersey. Therewere no significant cross-state differences inaverage hours of operation, the fraction offull-time workers, or the prevalence of bonusprograms to recruit new worker^.^The average starting wage at fast-foodrestaurants in New Jersey increased by 10percent following the rise in the minimumwage. Further insight into this change isprovided in Figure 1, which shows the dis-tributions of starting wages in the two statesbefore and after the rise. In wave 1, thedistributions in New Jersey and Pennsylva-nia were very similar. By wave 2 virtually all

'we discuss the sensitivity of our results to alterna-tive assumptions on the measurement of employmentin Section 111-C.' ~ h e s e programs offer current employees a cash"bounty" for recruiting any new employee who stayson the job for a minimum period of time. Typicalbounties are $50-$75. Recruiting programs that awardthe recruiter with an "employee of the month" desig-nation or other noncash bonuses are excluded from ourtabulations.


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THE AMERICAN ECONOMIC REVIEW SEPTEMBER 1994

Variable1. Distribution of Store Types (percentages):

a. Burger Kingb. KFCc. Roy Rogersd. Wendy'se. Company-owned2. Means in Wav e I:

a. F TE employmentb. Percen tage full-time employeesc. Starting waged. Wage = $4.25 (percen tage)e. Price of full mealf. Hour s open (weekday)g. Recruiting bonus

3. Means in Ware 2:a. FTE employmentb. Percentage full-time employeesc. Starting waged. Wage = $4.25 (percenta ge)e . Wage = $5.05 (percenta ge)f. Price of full mea lg. H ours open (weekday)h. Recruiting bonus

Stores in: NJ PA t a

20.4(0.51)32.8(1.3)4.61(0.02)30.5(2.5)

21.0 21.2 - 0.2(0.52) (0.94)35.9 30.4 1.8(1.4) (2.8)5.08 4.62 10.8(0.01) (0.04)0.0 25.3 -(4.9)85.2 1.3 36.1(2.0) (1.3)3.41 3.03 5.0(0.04) (0.07)14.4 14.7 -0. 8(0.2) (0.3)20.3 23.4 - 0.6(2.3) (4.9)

Notes: See text for definitions. S tandard errors a re given in parenthe ses.aT es t of equality of means in New Jersey a nd Pennsylvania.

restaurants in New Jersey that had beenpaying less than $5.05 per hour reported astarting wage equal to th e new rate. Inter-estingly, the minimum-wage increase had n oapp are nt "spillover" on higher-wage restau-rants in the state: the mean percentage wagechange for these stores was -3.1 percent.

Despite the increase in wages, full-time-equivalent employment increased in NewJersey relative to Pennsylvania. WhereasNew Jersey stores were initially smaller,employment gains in New Jersey coupledwith losses in Pennsylvania led to a smalland statistically insignificant interstate


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VO L. 84 NO. 4 CARD AND KRUEGER: MINIMUM WAGE AND EMPLOYMENT

February 1 9 9 2

Wage Range

November 1 9 9 2

Wage RangeNew Jerse y Pennsylvania

FIGURE1. DISTRIBUTION WAGERATESOF STARTING


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778 T HE AM ERIC AN EC ON OM IC REVIEW SEPTEM BER I994difference in wave 2. Only two other vari-ables show a relative change between waves1 and 2: the fraction of full-time employeesand the price of a meal. Both variablesincreased in New Jersey relative to Pennsyl-vania.W e c an assess the reliability of our surveyquestionnaire by comparing the responsesof 11 stores that were inadvertently inter-viewed twice in th e first wave of th e survey.10Assuming that measurement errors in thetwo interviews are independent of eachother and independent of the true variable,the correlation between responses gives anestimate of the "reliability ratio" (the ratioof the variance of the signal to the com-bined variance of the signal and noise). Theestimated reliability ratios are fairly high,ranging from 0.70 for full-time equivalentemploym ent to 0.98 for the price of a meal."W e have also checked whether stores withmissing data for any key variables are dif-ferent from restaurants with complete re-sponses. We find that stores with missingdata on employment, wages, or prices aresimilar in othe r respects t o stores with com-plete dat a. T he re is a significant size differ-ential associated with the likelihood of thestore closing after wave 1. The six storesthat closed were smaller than other stores(with an average employment of only 12.4full-time-equivalent employees in wave 1).12

111. Employment Effects of theMinimum-Wage IncreaseA. Differences in Differences

Table 3 summarizes the levels andchanges in average employment per store in10Thes e restaurants were interviewed twice b ecausethe i r phone numbers appeared in more than one phonebook, and neither the interviewer nor the respondentnoticed that they were previously interviewed.11Similar reliability ratios for very similar questionswere obtained by Katz and Krueger (1992).''A probit analysis of the probability of closureshows that the initial size of the store is a significantpredictor of closure. The level of starting wages has anumerically small and statistically insignificant coeffi-

cient in the probit model.

our survey. We present data by state incolumns (i) and (ii), and for stores in NewJersey classified by whether the startingwage in wave 1 was exactly $4.25 per hour[column (iv)] between $4.26 and $4.99 perhour [column (v)] or $5.00 or m ore per hour[column (vi)]. W e also show t he differencesin average employment between New Jerseyand Pennsylvania stores [column (iii)] andbetween stores in the various wage rangesin New Jersey [colum ns (viil-(viii)].Row 3 of the table presents the changesin average employment between waves 1and 2. These entries are simply the differ-ences between the averages for the twowaves (i.e., row 2 minus row 1). A n a lterna-tive estimate of the change is presented inrow 4: here we have computed the changein employment over the subsample of storesthat reported valid employment data in bothwaves. We refer to this group of stores asthe balanced subsample. Finally, row 5 pre-sents the average change in em ployment inthe balanced subsample, treating wave-2employment at the four temporarily closedstores as zero, rath er than as missing.As noted in Table 2, New Jersey storeswere initially smaller than their Pennsylva-nia counterparts but grew relative to Penn-sylvania stores after the rise in the mini-mum wage. The relative gain (the "dif-ference in differences" of the changes inemployment) is 2.76 FTE employees (or 13percent), with a t statistic of 2.03. Inspec-tion of the averages in rows 4 and 5 showsthat the relative change between New Jer-sey and Pennsylvania stores is virtually ide n-tical when the analysis is restricted to thebalanced subsample, and it is only slightlysmaller when wave-2 employment at thetemporarily closed stores is treated as zero.Within New Jersey, employment ex-panded at th e low-wage stores (those paying$4.25 per hour in wave 1 ) and contracted a tthe high-wage stores (those paying $5.00 ormo re per hour). Indeed, th e average changein employment at the high-wage stores( - 2.16 FTE employees) is almost identicalto the change among Pennsylvania stores(- 2.28 FTE employees). Since high-wagestores in New Jersey should have been


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V O L . 84 NO. 4 CARD AND KRUEGER: MINIMUM WAG E AND EMPLOYMENT

largely unaffected by the new minimumwage, this comparison provides a specifica-tion test of the validity of the Pennsylvaniacontrol group. The test is clearly passed.Regardless of whether the affected storesare compared to stores in Pennsylvania orhigh-wage stores in New Jersey, the esti-mated employment effect of the minimumwage is similar.The results in Table 3 suggest that em-ployment contracted between February andNovember of 1992 at fast-food store s tha twere unaffected by the rise in the minimumwage (stores in Pennsylvania and stores inNew Jersey paying $5.00 per hour or morein wave 1). We suspect that the reason forthis contraction was the continued worsen-ing of the economies of the middle-Atlanticstates d uring 1992.13 Unemploym ent ratesin New Jersey, Pennsylvania, and New Yorkall trended upward between 1991 and 1993,with a larger increase in New Jersey thanPennsylvania during 1992. Since sales offranchised fast-food restaurants are pro-cyclical, the rise in unemployment would beexpected to lower fast-food employment inthe absence of other factors.14

B. Regression-Adjusted ModelsThe comparisons in Table 3 make noallowance for other sources of variation inemployment growth, such as differencesacross chains. Th ese a re incorporated in theestimates in Table 4. The entries in thistable are regression coefficients from mod-

13An alternative possibility is that seasonal factorsproduce higher employment at fast-food restaurants inFebruary and March than in November and December.An analysis of national employment data for foodpreparation and service workers, however, shows higheraverage employment in the fourth quarter than in thefirst quarter.14To investigate the cyclicality of fast-food restau-rant sales we regressed the year-to-year change in U.S.sales of the McDonald's restaurant chain from1976-1991 on the corresponding change in the unem-ployment rate. The regression results show that a1-percentage-point increase in the unemployment ratereduces sales by $257 million, with a t statistic of 3.0.

els of the form:( la ) A E , = a + b X i + c N J i + ~ ,

( l b ) AE , = a' +b lXi+ clGAPi+ E{where AE, is the change in employmentfrom wave 1 to wave 2 at store i, Xi is a setof characteristics of store i, and NJ, is adummy variable that equals 1 for stores inNew Jersey. GAP, is an alternative measureof th e impact of th e minimum wage a t storei based on the initial wage at that store(W,,):GAP, = 0 for stor es in Pennsylvania

= 0 for sto res in New Jersey with

for othe r stores in New Jersey.GAP, is the proportional increase in wagesat store i necessary to meet the new mini-mum rate. Variation in GAP, reflects bothth e New Jersey-Pennsylvania con trast an ddifferences within New Jersey based on re-ported starting wages in wave 1. Indeed, th evalue of GAP, is a strong predictor of theactual proportional wage change betweenwaves 1 and 2 ( R *= 0.75), and conditionalon GAP, there is no difference in wagebehavior between stores in New Jersey andPennsylvania. l5The estimate in column (i) of Table 4is directly comparable to the simpledifference-in-differences of employmentchanges in column (iv), row 4 of Table 3.T h e d i s c r e p a n c y b e t w e e n t h e t w oestimates is due to the restricted sample inTable 4. In Table 4 and the remaining ta-bles in this section we restrict our analysisto the set of stores with available employ-ment and wage data in both waves of the

1 5 ~regression of the proportional wage change be-tween waves 1 and 2 on GAP, has a coefficient of 1.03.


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THE AMERICAN ECONOMIC REVlEW SEPTEMBER 1994

TABLE3-AVERAGE EMPLOYMENT THE R IS EPER STOREBEFOREAND I ~ E R IN NEW JERSEYM I N I M U MW A G E

Stores by state Stores in New Jersey a Differences within N J ~

Variable PA(i ) NJ(ii) Difference,N J - P A(iii) Wage=$4.25(iv) Wage

=$4.26-$4.99(v) Wager$5.00(vi) Low-high(vii) Midrange-high(viii)

1. FTE employment before,all available observations2. FT E employment after,all available observations3. C hange in mean FT Eemployment4. Change in mean FTEemploym ent, balancedsample of storesC5. Change in mean FT E

employment, settingFT E at temporarilyclosed stores to O dNotes: Sta ndar d errors ar e shown in parentheses. Th e sample consists of all stores with available data on employment. FTE(full-time-equivalent) employment counts each part-time worker as half a full-time worker. Employment at six closed storesis set to zero. Employment at four temporarily closed stores is treated as missing.as ta re s in New Jersey were classified by whether starting wage in wave 1 equals $4.25 per ho ur ( N = 101), is between$4.26 and $4.99 per hour ( N = 140), or is $5.00 per hour o r higher ( N = 73).b ~ if f e r e n c ein employment between low-wage ($4.25 per ho ur) and high-wage ( 2$5.00 per hour) stores; and differencein employment between midrange ($4.26-$4.99 per hour) and high-wage stores.'Subset of stores with available employm ent da ta in wave 1 and wave 2.this row only, wave-2 employment at four temporarily closed stores is set to 0. Employment changes a re based o n thesubset of stores with available employment d ata in wave 1 and wave 2.

TABLE4-REDUCED-FORM MODELSFOR CHANGEIN EMPLOYMENTM o d e l

Independ en t var iab le (i) (ii) (iii) (iv) (v)1. New Jersey dumm y 2.33 2.30 - - -(1.19) (1.20)2. Initial wage gapa - - 15.65 14.92 11.91(6.08) (6.21) (7.39)3 . Controls for chain and no yes no yes yesownersh ipb4. Controls for regionC5. Standard e rror of regress ion6. Probability value for controlsdNotes: Standard errors are g iven in parentheses . The sample consis ts of 357 s toreswith avai lable data on employment and s tart ing wages in waves 1 and 2 . Thedependen t var iab le in a l l model s i s change in FTE employmen t . The mean andstan dard deviat ion of the de pend ent variable ar e -0 .237 and 8 .825, respect ively. Allmodels include an unres tr icted constant (not reported).aProport ional increase in s tart ing wage necessary to raise s tart ing wage to newminimu m rate. For s tores in Pennsylvania the wage g ap is 0 .b ~ h r e edummy variables for chain type and whether or not the s tore is company-owned a re inc luded.'Dummy variables for two regions of New Jersey and two regions of eas ternPennsylvania are included.d~ ro ba bi l i tyvalue of jo in t F test for exclusion of all control variables.


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781V O L . 84 NO . 4 CARD AND KRUEGER: MINIMUM WAG E AND EMPLOYMENT

survey. This restriction results in a slightlysmaller estimate of the relative increase inemployment in New Jersey.The model in column (ii) introduces aset of four control variables: dummies forthree of the chains and ano ther dummy forcompany-owned stores. As shown by theprobability values in row 6, these covariatesadd little to the model and have no effecton the size of the estimated New Jerseydummy.T h e specifications in columns (iiil-(v) usethe GAP variable to measure the effect ofthe minimum wage. This variable gives aslightly better fit than the simple New Jer-sey dummy, although its implications for theNew Jersey-Pennsylvania comparison aresimilar. The mean value of GAPi amongNew Jersey stores is 0.11. Thu s the estimatein column (iii) implies a 1.72 increase inFTE employment in New Jersey relative toPennsylvania.Since GA P, varies within N ew Jersey, it ispossible to add both GAP, and NJ, to theemployment model. The estimated coeffi-cient of the New Jersey dummy then pro-vides a test of the Pennsylvania controlgroup. When we estimate these models, thecoefficient of the New Jersey dummy is in-significant (with t ratios of 0.3-0.7), imply-ing that inferences about the effect of theminimum wage are similar whether thecomparison is made across states or acrossstores in New Jersey with higher and lowerinitial wages.An even stronger test is provided in col-umn (v), where we have added dummiesrepresenting three regions of New Jersey(North, Central, and Sou th) and two regionsof eastern Pennsylvania (Allentown-Eastonand the northern suburbs of Philadelphia).These dummies control for any region-s~ e c i f i cdemand shocks and identifv the ef-feet of the minimum wage byemployment changes at higher- and lower-wage within the same region of NewJersey. Th e probability value in row 6 showsno evidence of regional components in em-ployment growth. The addition of the re-gion dummies attenuates the GAP coeffi-cient and raises its standard error, however,making it no longer possible to reject the

null hypothesis of a zero employment effectof the minimum wage. One explanation forthis attenuatio n is the presence of measure-ment error in the starting wage. Even ifemployment growth has no regional compo-nent, the addition of region dummies willlead to some attenuation of the estimatedGAP coefficient if some of the true varia-tion in GAP is explained by region. Indeed,calculations based on the estimated reliabil-ity of the G A P variable (from the set of 1 1double interviews) suggest that the fall inthe estim ated G A P coefficient from column(iv) to column (v ) is just equal to the ex-pected change attributable to measurementerror.16We have also estimated the models inTable 4 using as a dependent variable theproportional change in employment at eachstore.17 The estimated coefficients of theNew Jersey dummy and the GAP variableare uniformly positive in these models butinsignificantly different from 0 at conven-tional levels. The implied employment ef-fects of the minimum wage are also smallerwhen t he de pend ent variable is expressed inproportional terms. For example, the GAPcoefficient in column (iii) of Table 4 impliesthat the increase in minimum wages raisedemployment at New Jersey stores that wereinitially paying $4.25 per hour by 14 per-cent. The estimated GAP coefficient from acorresponding proportional model impliesan effect of only 7 per cen t. Th e difference isattributable to heterogeneity in the effect ofthe minimum wage at larger and smallerstores. Weighted versions of the propor-tional-change models (using initial employ-ment as a weight) give rise to wage elastici-

16In a regression model without other controls theexpected attenuation of the GAP coefficient due tomeasurement error is the reliability ratio of GAP (yo),which we estimate at 0.70. The expected attenuationfactor when region dummies are added to the model isy l = (Yo- ~ 2) / ( 1 - ~ 2 ) ,where ~2 is the R-squarestatistic of a regression of GAP on region effects (equalto 0.30). Thus, we expect the estimated GAP coeffi-cient to fa ll by a factor of Y I / Y O = 0.8 when regiondummies are added to a regression model." ~ h e s e specifications are reported in table 4 ofCard and Krueger (1993).


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782 THE AMERICAN ECONOMIC REVIEW SEPTEMBER I994ties similar to t he elasticities implied by theestimates in T able 4 (see below).

C . Specification TestsThe results in Tables 3 and 4 seem tocontradict the standard prediction that arise in the minimum wage will reduce em-ployment. Table 5 presents som e alternativespecifications that probe the robustness ofthis conclusion. For completeness, we re-port estimates of models for the change inemployment [columns (i) and (ii)] and esti-mates of models for the prop ortional changein employm ent [columns (iii) an d (iv)].18 T hefirst row of the table reproduces the "basespecification" from colum ns (ii) and (iv) ofTable 4. (Note that these models includechain dummies and a dummy for company-owned stores). Row 2 presents an alterna-tive set of estimates when we set wave-2employment at th e temporarily closed storesto 0 (expanding our sample size by 4). Thischange has a sm all attenuatin g effect o n thecoefficient of the New Jersey dummy (sinceall four stores are in New Jersey) but lesseffect on the GAP coefficient (since the size

of GAP is uncorrelated with the probabilityof a temporary closure within New Jersey).Rows 3-5 presen t estimation results us-ing alternative measures of full-time-equiv-alent employment. In row 3, employment isredefined to exclude management employ-ees. This change has no effect relative tothe base specification. In rows 4 and 5, weinclude managers in FTE employment butreweight part-tim e workers as either 40 per-cent or 60 percent of full-time workers (in-stead of 50 percent).19 These changes have

18The proportional change in employment is de-fined as the change in employment divided by theaverage level of employment in waves 1 and 2. Thisresults in very similar coefficients but smaller standarderrors than the alternative of dividing by wave-1 em-ployment. For closed store s we set the prop ortionalchange in employment to - 1.19Analysis of th e 1991 Curre nt Population Surveyreveals that part-time workers in the restaurant indus-try work about 46 percent as many hours as full-timeworkers. Katz and Krueger (1992) report that the ratioof par t-tim e worke rs' hours to full-time workers' h our sin the fast-food industry is 0.57.

little effect on the models for the level ofemployment but yield slightly smaller pointestimates in the proportional-employment-change models.In row 6 we present estimates obtainedfrom a su bsamp le that excludes 35 stores intowns along the New Jersey shore. The ex-clusion of these stores, which may have adifferent seasonal pattern than other storesin our sample, leads to slightly larger mini-mum -wage effects. A similar finding emerg esin row 7 when we add a set of dummyvariables that indicate the week of thewave-2 inter vie^.^'As noted earlier, we made an extra effortto obtain responses from New Jersey storesin the first wave of our survey. The fractionof stores called three or more times to ob-tain an interview was higher in New Jerseythan in Pennsylvania. To check the sensitiv-ity of our results to this sampling feature,we reestimated our models on a subsamplethat excludes any stores that were calledback mo re th an twice. T he results, in row 8,are very similar to the base specification.Row 9 presents weighted estimation re-sults for the proportional-employment-

change models, using as weights the initiallevels of employment in each store. Sincethe proportional change in average employ-ment is an employment-weighted average ofthe proportional changes at each store, aweighted version of the prop ortional-changemodel should give rise to elasticities thatare similar to th e implied elasticities arisingfrom the levels models. Co nsistent with thisexpectation, the weighted estimates arelarger than the unweighted estimates, andsignificantly different from 0 at conventionallevels. The weighted estimate of the NewJersey dummy (0.13) implies a 13-percentrelative increase in New Jersey employment-the sam e prop ortion al employ ment effectimplied by the simple difference-in-dif-ferences in Table 3. Similarly, the weightedestimate of the GAP coefficient in theproportional-change model (0.81) is close to

20We also added dummies for the interview datesfor the wave-1 survey, but these were insignificant anddid not change the estimated minimum-wage effects.


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783VOL. 84 NO. 4 CARD AND KRUEGER: MINIMUM WAGE AND EMPLOYMENT

Proportional changeChange in employment in employmentNJ dummy Gap measure NJ dummy Gap measureSpecification (i) (ii) (iii) (iv)

1. Base specification 2.30 14.92(1.19) (6.21)2. Treat four temporarily closed stores as permanently closeda 3. Exclude managers in employment countb 4. Weight part-time as 0.4x full-timec5. Weight part-time as 0.6X full-timed6. Exclude stores in NJ shore areae7. Add controls for wave-2 interview dateE 8. Exclude stores called more than twice in wave l g 9. Weight by initial employmenth

10. Stores in towns around Newark' - 33.75(16.75)11. Stores in towns around CamdenJ - 10.91(14.09)12. Pennsylvania stores only - - 0.30(22.00)Notes: Standard errors are given in parentheses. Entries represent estimated coefficient of New Jersey dummy[columns (i) and (iii)] or initial wage gap [columns (ii) and (iv)] in regression models for the change in employmentor the percentage change in employment. All models also include chain dummies and an indicator for company-owned stores.aWave-2 employment at four temporarily closed stores is set to 0 (rather than missing). b~ull-timeequivalent employment excludes managers and assistant managers. CFull-time equivalent employment equals number of managers, assistant managers, and full-time nonmanagement workers, plus 0.4 times the number of part-time nonmanagement workers.d~ull-timeequivalent employment equals number of managers, assistant managers, and full-time nonmanage-ment workers, plus 0.6 times the number of part-time nonmanagement workers.eSample excludes 35 stores located in towns along the New Jersey shore.' ~ o d e l sinclude three dummy variables identifying week of wave-2 interview in November-December 1992.gSample excludes 70 stores (69 in New Jersey) that were contacted three or more times before obtaining thewave-1 interview.h~egressio nmodel is estimated by weighted least squares, using employment in wave 1 as a weight.. Subsample of 51 stores in towns around Newark.Subsample of 54 stores in town around Camden.Subsample of Pennsylvania stores only. Wage gap is defined as percentage increase in starting wage necessaryto raise starting wage to $5.05.

i


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784 THE AME RICAN ECONOM IC REVIE W SEPTEMBER 1994

the implied elasticity of employment withrespect to wages from the basic levels speci-fication in row 1, column (iiI2l These find-ings suggest that the proportional effect ofthe rise in the minimum wage was concen-trated among larger stores.One explanation for our finding that arise in the minimum wage has a positiveemployment effect is that unobserved de-man d shocks within New Jersey outweighedth e negative employm ent effect of t he m ini-mum wage. T o address this possibility, rows10 and 11 present estimation results basedon subsamples of stores in two narrowlydefined areas: towns around Newark (row10) and towns around Camden (row 11). Ineach case the sample area is identified bythe first three digits of th e store's zip code.22Within both areas the change in employ-ment is positively correlated with the GAPvariable, although in neither case is theeffect statistically significant. To the extentthat fast-food product market conditions areconstant within local areas, these resultssuggest that our findings are not driven byunobserved dem and shocks. Ou r analysis ofprice changes (reported below) also sup-ports this conclusion.A final specification check is presented inrow 12 of Tab le 5. In this row we excludestores in New Jersey and (incorrectly) de-fine the GAP variable for Pennsylvaniastores as the proportional increase in wagesnecessary to raise the wage to $5.05 perhour. In principle the size of the wage gapfor stores in Pennsylvania should have nosystematic relation with employm ent growth.In practice, this is the case. There is noindication that the wage gap is spuriouslyrelated to employment growth.

21~ssumingaverage employment of 20.4 in NewJersey, the 14.92 GAP coefficient in row 1, column (ii)im lies an employment elasticity of 0.73."The "070" three-digit zip-code area (aroundNewark) and the "080" three-digit zip-code area(around Camden) have by far the largest numbers ofstores among three-digit zip-code areas in New Jersey,and together they account for 36 percent of New Jerseystores in our sample.

We have also investigated whether thefirst-differenced specification used in ouremployment models is appropriate. Afirst-differenced model implies that the levelof employment in period t is related to thelagged level of employment with a coeffi-cient of 1. If sho rt-run em ploym ent fluctua-tions are smoothed, however, the true co-efficient of lagged employment may be lessthan 1. Imposing the assumption of a unitcoefficient may then lead to biases. To testthe first-differenced specification we reesti-mated models for the change in employ-ment including wave-1 employment as anadditional explanatory variable. To over-come any mechanical correlation betweenbase-period employment and the change inemployment (attr ibutable to measurementerror) we instrumented wave-1 employmentwith the number of cash registers in thestore in wave 1 and the number of registersin the store that were open at 11:OO A.M. Inall of the specifications the coefficient ofwave-1 employment is close to zero. Forexample, in a specification including theGAP variable and ownership and chaindummies, the coefficient of wave-1 employ-ment is 0.04, with a standard error of 0.24.We conclude tha t th e first-differenced spec-ification is appropriate.

D . Full-Time and Part-Time SubstitutionOur analysis so far has concentrated onfull-time-equivalent employment and ig-nored possible changes in the distributionof full- and part-time workers. An increasein the minimum wage could lead to an in-crease in full-time employment relative to

part-time employment for at least two rea-sons. First, in a conventional model onewould expect a minimum-wage increase toinduce employers to su bstitute skilled work-ers and capital for minimum-wage workers.Full-time workers in fast-food restaurantsare typically older and may well possesshigher skills than part-time w orkers. Thus, aconventional model predicts that stores mayrespond to an increase in the minimumwage by increasing the proportion of full-time workers. Nevertheless, 81 percent ofrestaurants paid full-time and part-time


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Outcome measureStore Characteristics:

Mean cNJ(i)

hange inPA(ii)

outcomeNJ -PA(iii)

Regression of change inoutcome variable on:NJ dummy Wage gapa Wage gapb(iv) (v) (vi)

1. Fraction full-time workersc (percentage)2. Number of hours open per weekday3. Number of cash registers4. Number of cash registers openat 11:OOA.M.

Employee Meal Programs:5. Low-price meal program (percentage)6. Free meal program (percentage)7. Combination of low-price and freemeals (percentage)

Wage Profile:8. Time to first raise (weeks)9. Usual amount of first raise (cents)

10. Slope of wage profile (percentper week)Notes: Entries in columns (i) and (ii) represent mean changes in the outcome variable indicated by the row headingfor stores with available data on the outcome in waves 1 and 2. Entries in columns (iv)-(vi) represent estimatedregression coefficients of indicated variable (NJ dummy or initial wage gap) in models for the change in theoutcome variable. Regression models include chain dummies and an indicator for company-owned stores.aThe wage gap is the proportional increase in starting wage necessary to raise the wage to the new minimumrate. For stores in Pennsylvania, the wage gap is zero.b ~ o d e l sin column (vi) include dummies for two regions of New Jersey and two regions of eastern Pennsylvania.'Fraction of part-time employees in total full-time-equivalent employment.

workers exactly the same starting wage in workers are more productive (but equallywave 1 of o ur survey.23 This suggests eit her paid), there may be a second reason forthat full-time workers have the same skills stores to substitute full-time workers foras part-time wo rkers or that equity concerns part-time workers; namely, a minimum-wagelead restaurants to pay equal wages for un- increase enables the industry to attract moreequally productive workers. If full-time full-time workers, and stores would natu-rally want to hire a greater proportion offull-time workers if they are more produc-tive.231n the other 19 percent of stores, full-time workers Row 1 of Table 6 presents the meanare paid more, typically 10 percent more. changes in the proportion of full-time work-


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786 THE AMERICAN ECONOMIC REVIEW SEPTEMBER 1994

ers in New Jersey and Pennsylvania be-tween waves 1 and 2 of our survey, andcoefficient estimates from regressions of thechange in the proportion of full-time work-ers on the wage-gap variable, chain dum-mies, a company-ownership dummy, and re-gion dumm ies [in column ( 4 1 . Th e resultsare ambiguous. The fraction of full-timeworkers increased in New Jersey relative toPennsylvania by 7.3 percent (t ratio = 1.841,but regressions on the wage-gap variableshow no significant shift in the fraction offull-time workers.24E. Other Employment-Related M easuresRows 2-4 of Table 6 present results forother outcome variables that we expect tobe related to the level of restaurant employ-ment. In particular, we examine whetherthe rise in the minimum wage is associatedwith a change in the number of hours arestaurant is open on a weekday, the num-ber of cash registers in the restaurant, andthe number of cash registers typically inoperation in the restaurant at 11:OO A.M.Consistent with our employment results,

none of these variables shows a statisticallysignificant decline in New Jersey relative toPennsylvania. Similarly, regressions includ-ing the gap variable provide no evidencethat the minimum-wage increase led to asystematic change in any of these variables[see columns (v) and (vi)].IV. Nonwage Offsets

One explanation of our finding that a risein the minimum wage does not lower em-ployment is that restaurants can offset theeffect of the minimum wage by reducingnonwage compensation. For example, ifworkers value fringe benefits and wagesequally, employers can simply reduce thelevel of fringe benefits by the amount of theminimum-wage increase, leaving their em-

2 4 ~ i t h i nNew Jersey, the fraction of full- t ime em-ployees increased ab out as quickly at stores with higherand lower wages in wave 1.

ployment costs unchanged. T he m ain fringebenefits for fast-food employees are freeand reduced-price meals. In the first waveof our survey about 19 perce nt of fast-foodrestaurants offered workers free meals. 72percent offered reduced-price meals, ai d 9percent offered a combination of both freeand reduced-price meals. Low-price mealsare an obvious fringe benefit to cut if theminimum-wage increase forces restaurantsto pay h igher wages.Rows 5 and 6 of Table 6 present esti-mates of the effect of the minimum-wageincrease on the incidence of free meals andreduced-price meals. Th e proportion of res-taurants offering reduced-price meals fellin both New Jersey and Pennsylvania afterthe minimum wage increased, with a some-what greater decline in New Jersey. Con-trary to an offset story, however, the reduc-tion in reduced-price meal programs wasaccompanied by an increase in the fractionof stores offering free meals. Relative tostores in Pennsylvania, New Jersey employ-ers actually shifted toward more generousfringe benefits (i.e., free meals rather thanreduced-price meals). However, the relativeshift is not statistically significant.We continue to find a statistically in-significant effect of the minimum-wage in-crease o n the likelihood of receiving free o rreduc ed-pric e meals in colum ns (v) and (vi),where we report coefficient estimates of theGAP variable from regression models forthe change in the incidence of these pro-grams. The results provide no evidence thatemployers offset the minimum-wage in-crease by reducing free or reduced-pricemeals.Another possibility is that employers re-sponded to the increase in the minimumwage by reducing on-the-job training andflattening the tenure-wage profile (seeJacob Mincer and Linda Leighton, 1981).Indeed, one manager told our interviewer inwave 1 that her workers were forgoing ordi-nary scheduled raises because the minimumwage was about to rise, and this wouldprovide a raise for all her workers. To de-termine whether this phenomenon occurredmore generally, we analyzed store man-agers' responses to questions on the amount


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VOL. 84 NO. 4 CARD AND KRUEGER: MINIMUM WAGE AND EMPLOYMENT 787

of time before a normal wage increase andthe usual amount of such raises. In rows 8and 9 we report the average changes be-tween waves 1 and 2 for these two variables,as well as regression coefficients from mod-els that include th e wage-gap variable.25Al-though the average time to the first payraise increased by 2.5 weeks in New Jerseyrelative to Pennsylvania, th e increase is notstatistically significant. Furthermore, thereis only a trivial difference in the relativechange in the amo unt of th e first pay incre-ment between New Jersey and Pennsylvaniastores.Finally, we examined a related variable:the "slope" of the wage profile, which weme asur e by the ra tio of th e typical first raiseto th e am ount of time until th e first raise isgiven. As shown in row 10, the slope of thewage profile flattened in both New Jerseyand Pennsylvania, with no significant rela-tive difference between states. The changein the slope is also uncorrelated with theGAP variable. In summary, we can find noindication that New Jersey employerschanged either their fringe benefits or theirwage profiles to offset the rise in the mini-mum wage.26

V. Price Effects of the Minimum-Wage Increase A final issue we examine is the effect ofthe minimum w age on the prices of meals atfast-food restaurants. A competitive modelof the fast-food industry implies that anincrease in the minimum wage will lead toan increase in product prices. If we assumeconstant returns to scale in th e industry, theincrease in price should be proportional tothe share of minimum-wage labor in total

2 5 ~ nwave 1, the average time to a first wage in-crease was 18.9 weeks, and the average amount of thefirst increase was $0.21 per hour.2 6 ~ a t zand Krueger (1992) report that a significantfraction of fast-food stores in Texas responded to anincrease in the minimum wage by raising wages forworkers who were initially earning more than the newminimum rate. Our results on the slope of the tenureprofile are consistent with their findings.

factor cost. The average restaurant in NewJersey initially paid about half its workersless than the new minimum wage. If wagesrose by roughly 15 percent for these work-ers, and if labor's share of total costs is 30percent, we would expect prices to rise byabout 2.2 percent ( = 0.15 X 0.5 X 0.3) du e tothe minimum-wage rise.27In each wave of our survey we askedmanagers for the prices of three standarditems: a medium soda, a small order offrench fries, and a main course. The maincourse was a basic hamburger at BurgerKing, Roy Rogers, and W endy's restaurants,and two pieces of chicken at KFC stores.We define "full meal" price as the after-taxprice of a medium soda, a small order offrench fries, and a main course.Table 7 presents reduced-form estimatesof the effect of the minimum-wage increaseon prices. The dependent variable in thesemodels is the change in th e logarithm of th eprice of a full meal at each store. The keyindependent variable is either a dummy in-dicating whether t he store is located in NewJersey or the proportional wage increaserequired to meet the minimum wage (theG AP variable defined above).The estimated New Jersey dummy in col-umn (i) shows that after-tax meal pricesrose 3.2-percent faster in New Jersey thanin Pennsylvania between February andN o v e m b e r 1 9 9 2 . ~ ~The effect is slightlylarger controlling for chain and company-ow nersh ip [se e column ($1. Since th eNew Jersey sales tax rate fell by 1 percent-age point between the waves of our survey,these estimates suggest that pretax pricesrose 4-percent faster as a result of the

" ~c co r d i n ~to the McDonald's Corporation 1991Annual Report. payroll and benefits are 31.3 percent ofoperating costs at company-owned stores. This calcula-tion is only approximate because minimum-wage work-ers make up less than half of payroll even though theyare about half of workers, and because a rise in theminimum wage causes some employers to increase thepay of other higher-wage workers in order to maintainrelative pay differentials.he effect is attributable to a 2.0-percent increasein prices in New Jersey and a 1.0-percent decrease inprices in Pennsylvania.


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- - - - - - - - - - - - - - - -

THE AMERICAN ECONOMIC REVIEW SEPTEMBER 1994TABLE7-REDUCED-FORMMODELS I N THE PRICEOF A FULLMEALFOR CHANGE

Dependent variable: change in the log priceof a full mealIndependent variable (i) (ii) (iii) (iv) (v)1. New Jersey dummy 0.033 0.037 - - -(0.014) (0.014)2. Initial wage gapa - - 0.077 0.146 0.063(0.075) (0.074) (0.089)3. Controls for chain andb no yes no yes Yesownership4. Controls for regionC no no no no yes5. Standard error of regression 0.101 0.097 0.102 0.098 0.097Notes: Standard errors are given in parentheses. Entries are estimated regressioncoefficients for models fit to the change in the log price of a full meal (entrCe, mediumsoda, small fries). The sample contains 315 stores with valid data on prices, wages, andemployment for waves 1 and 2. The mean and standard deviation of the dependentvariable are 0.0173 and 0.1017, respectively.aProportional increase in starting wage necessary to raise the wage to the newminimum-wage rate. For stores in Pennsylvania the wage gap is 0.bThree dummy variables for chain type and whether or not the store is company-owned are included.'Dummy variables for two regions of New Jersey and two regions of easternPennsylvania are included.

minimum-wage increase in New Jersey- One potential explanation for the latterslightly more than the increase needed to finding is that stores in New Jersey com petepass through t he cost increase caused by the in the same product market. As a result,minimum-wage hike. restaurants that are most affected by theThe pattern of price changes within New minimum wage are unable to increase theirJersey is less consistent with a simple produ ct prices faster than th eir competitors." pass-through" view of minimum-wage cost In contrast, stores in New Jersey and Penn-increases. In fact, meal prices rose at sylvania are in separate product markets,approximately the same rate at stores in enabling prices to rise in New Jersey rela-New Jersey with differing levels of initial tive to Pennsylvania when overall costs risewages. Inspection of the estimated GAP in New Jersey. Note that this explanationcoefficients in column (v) of Table 7 con- seems to rule out the possibility that store-firms th at within regions of New Jersey, th e specific dem and shocks can accoun t for th eG A P variable is statistically insignificant. anomalous rise in employment at stores inIn sum, these results provide mixed evi- New Jersey with lower initial wages.dence that higher minimum wages result inhigher fast-food prices. The strongest evi- VI. Store Openingsdence emerges from a comparison of NewJersey and Pennsylvania stores. The magni- An important potential effect of highertude of the price increase is consistent with minimum wages is to discourage the open-predictions from a conven tional model of a ing of new businesses. Alth oug h our samp lecompetitive industry. On the other h and, we design allows us to es tima te the effect of thefind no evidence that prices rose faster minimum wage on existing restaurants inamong stores in New Jersey that were most New Jersey, we canno t address th e effect ofaffected by th e rise in the minimum wage. the higher minimum wage on potential


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789VO L. 84 NO. 4 CARD AND KRUEGER: MINIMUM WAGE AND EMPLOYMENTentra nts.2 9 T o assess th e likely size of suchan effect, we used national restaurant direc-tories for the McDonald's restaurant chainto compare the numbers of operatingrestaurants and the numbers of newlyopened restaurants in different states overthe 1986-1991 period. Many states raisedtheir minimum wages during this period. Inaddition, the federal minimum wage in-creased in the early 1990's from $3.35 to$4.25, with d iffering effects in differe nt sta tesdepending on the level of wages in thestate. These policies create an opportunityto measure the impact of minimum-wagelaws on sto re openin g rates across states.

The results of our analysis are presentedin Table 8. We regressed the growth rate inthe number of McDonald's stores in eachstate on two alternative measures of theminimum wage in the state and a set ofother control variables (population growthand the change in the state unemploymentrate). The first minimum-wage measure isthe fraction of workers in the state's retailtra de industry in 1986 whose wages fell be-tween the existing federal minimum wage in1986 ($3.35 per hour) and the effective min-imum wage in the state in April 1990 (themaximum of the federal minimum wage andth e state minimum wages as of April 1990)."The second is the ratio of the state's effec-tive minimum wage in 1990 to the averagehourly wage of retail trade workers in thestate in 1986. Both of these m easures ar edesigned to gauge the degree of upwardwage pressure exerted by state or federalminimum-wage changes between 1986 and1990.The results provide no evidence thathigher minimum-wage rates (relative to theretail-trade wages in a state) exert a nega-

29 Direct inquiries to the chains in our sample re-vealed that Wendy's opened two stores in New Jerseyin 1992 and one store in Pennsylvania. The otherchains were unwilling to provide information on newopenings.30We used the 1986 Current Population Survey(merged monthly file) to construct the minimum-wagevariables. State minimum-wage rates in 1990 were ob-tained from the Bureau of National Affairs LaborRelations Reporter Wages and Hours Manual (undated).

tive effect on either the net number ofrestaurants or the rate of new openings. Tothe contrary, all the estimates show positiceeffects of higher minimum wages on thenum ber of operating or newly open ed stores,although many of the point estimates areinsignificantly different from zero . While thisevidence is limited, we conclude that theeffects of minimum wages on fast-food storeopenin g rates are probably small.

VII. Broader Evidence on Employment Changes in New Jersey O ur establishment-level analysis suggests

that the rise in the minimum wage in NewJersey may have increased employment inthe fast-food industry. Is this just an anomalyassociated with our particular sample, or aphenomenon unique to the fast-food indus-try? Data from the monthly Current Popu-lation Survey (CPS) allow us to comparestate-wide employment trends in New Jer-sey and the surrounding states, providing acheck on the interpretation of our findings.Using monthly CPS files for 1991 and 1992,we computed employment-population ratesfor teenagers and adults (age 25 and older)for New Jersey, Pennsylvania, New York,and th e entire United States. Since the NewJersey minimum wage rose on April 1, 1992,we computed the employment rates forApril-December of bot h 1991 and 1992.Th e relative changes in employment in NewJersey and the surrounding states then givean indication of the effect of the new law.A comparison of changes in adult em-ployment rates show that the New Jerseylabor market fared slightly worse over the1991-1992 period th an eith er the U.S. labormarket as a whole or labor markets inPennsylvania or New York (see Card andKrueger, 1993 table 9 l3 ' Among teenagers,however, the situation was reversed. In NewJersey, teenage employment rates fell by 0.7percent from 1991 to 1992. In New York,

31The employment rate of individuals age 25 andolder fell by 2.6 percent in New Jersey between 1991and 1992, while it rose by 0.3 percent in Pennsylvania,and fell by 0.2 percent in the United States as a whole.


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790 THE AMERICA N ECONOMIC REVIEW SEPTEMBER 1994

Dep enden t variable:Depe nden t variable: propo rtional (num ber of newly opened stores )+increase in number of stores (number in 1986)Ind ep en de nt variable (i) (ii) (iii) (iv) (v) (vi) (vii) (viii)

Minim um- Wage Variable:1. Fraction of retail workers 0.33 - 0.13 - 0.37 - 0.16 -in affected wage range 1986" (0.20) (0.19) (0.22) (0.21)2. (State minimum wage in 1991)+ - 0.38 - 0.47 - 0.47 - 0.56(average retail wage in 1986Ib (0.22) (0.22) (0.23) (0.24)Other Control Variables:3. Proportional growth in - - 0.88 1.03 - - 0.86 1.04popu lation, 1986-1991 (0.23) (0.23) (0.25) (0.25)4. Change in unemployment - - -1.78 -1.40 - - - 1.85 - 1.40rates, 1986-1991 (0.62) (0.61) (0.68) (0.65)5. Stan dard err or of regression 0.083 0.083 0.071 0.068 0.088 0.088 0.077 0.073Notes: Standard errors are shown in parentheses. The sample contains 51 state-level observations (including theDistrict of Columbia) on th e num ber of McDo nald's restaurants o pen in 1986 and 1 991. Th e depen den t variable incolumns (i)-(iv) is the prop ortional increase in the num ber of restaurants o pen . Th e mean and sta nd ard deviationare 0.246 and 0.085, respectively. The dependent variable in columns (v)-(viii) is the ratio of the number of newstores opened between 1986 and 1991 to the number op en in 1986. Th e mean and standard deviation are 0.293 and0.091, respectively. All regressions are weighted by the state population in 1986.aFraction of all workers in retail trade in the state in 1986 earning an hourly wage between $3.35 per hour andthe "effective" state m inimum wage in 1990 (i.e., the maximu m of the fede ral minimum wage in 1990 ($3.80) andthe state minimum wage as of April 1, 1990).b ~ a x i m u mof state and federal minimum wage as of April 1, 1990, divided by the average hourly wage ofworkers in retail trade in the state in 1986.

Pennsylvania, and the United States as a summarize the predictions of the standardwhole, teenage employment rates dropped model a nd som e simple alternatives, and wefaster. Relative to teenagers in Pennsylva- highlight the difficulties posed by our find-nia, for example, the teenage employment ings.rate in New Jersey rose by 2.0 percentagepoints. While this point estimate is consis- A. Standard Competitive Modeltent with our findings for the fast-food in-dustry, the standard error is too large (3.2 A standard competitive model predictspercent) to allow any confident assessment. tha t establishm ent-level employm ent will fallif the wage is exogenously raised. For anVIII. Interpretation entire industry, total employment is pre-dicted to fall, and product price is predictedAs in the earlier study by Katz and to rise in response to an increase in a bind-Krueger (1992), our empirical findings on ing minimum wage. Estimates from theth e effects of th e New Jersey minimum w age time-series literature on minimum-wage ef-are inconsistent with the predictions of a fects can be used to get a rough idea of theconventional competitive model of the fast- elasticity of low-wage employment to thefood industry. Our employment results are minimum wage. The surveys by Brown et al.consistent with several alternative models, (1982. 1983) conclude that a 1 0 -~ er ce n tin -although none of these models can also crease in th e cove rage-adjusted minimumexplain th e ap pare nt rise in fast-food prices wage will reduce teenage employment ratesin New Jersey. In this section we briefly by 1-3 percent. Since this effect is for all


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teenagers, and not just those employed inlow-wage industries, it is surely a lowerbound on the magnitude of the effect forfast-food workers. The 18-percent increasein the New Jersey minimum wage is there-fore predicted to reduce employment atfast-food stores by 0.4-1.0 emplo yees perstore. Our empirical results clearly rejectthe upper range of these estimates, al-though we cannot reject a small negativeeffect in some of our specifications.A possible defense of the competitivemodel is that unobserved demand shocksaffected certain stores in New Jersey-specifically, those stores that were initiallypaying wages less than $5.00 per hour. How-ever, such localized demand shocks shouldalso affect product prices. (In fact, in acompetitive model, product demand shockswork through a rise in prices.) Althoughlower-wage stores in New Jersey had rela-tive employment gains, they did not haverelative price increases. Furthermore, ouranalysis of employment changes in two ma-jor suburban areas (around Newark andCamden) reveals that, even within localareas, employment rose faster at the storesthat had to increase wages th e most becauseof the new minimum wage.

B. Alternative ModelsAn alternative to the conventional com-petitive model is one in which firms areprice-takers in the product market but havesome degree of market power in the labormarket. If fast-food stores face an upward-sloping labor-supply schedule, a rise in theminimum wage can potentially increase em -ployment at affected firms and in the indus-try as a whole.32This same basic insight emerges from anequilibrium search model in which firmspost wages and employees search amongposted offers (see Dale T . Mortensen, 1988).Ken neth Burdett and M ortensen (1989) de-

" ~ a n i e l G. Sullivan (1989) and Michael R Ransom(1993) present empirical results for nurses and univer-sity teachers that suggest monopsony-like behavior ofemployers.

rive the equilibrium wage distribution for anoncooperative wage-search/wage-postingmodel and show that the imposition of abinding minimum wage can increase bothwages and em ployment relative to the initialequilibrium. Furthermore, their model pre-dicts that the minimum wage will increaseemployment the most at firms that initiallypaid th e lowest wages.Although monopsonistic and job-searchmodels provide a potential explanation forthe observed employment effects of the NewJersey minimum wage, they cannot explainthe observed price effects. In these models,industry prices should have fallen in NewJersey relative to Pennsylvania, and at low-wage stores in New Jersey relative to high-wage stores in New Jersey. Neither predic-tion is confirmed: indeed, prices rose fasterin New Jersey than in Pennsylvania, al-though at about t he same rate at high- andlow-wage stores in New Jersey. Anotherpuzzle for equilibrium search models is theabsence of wage increases at firms that wereinitially paying $5.05 or more per hour.The strict link between the employmentand price effects of a rise in the minimumwage may be broken if fast-food stores canvary th e quality of service (e.g., th e len gth ofthe queue at peak hours, or the cleanlinessof stores). Another possibility is that storesaltered the relative prices of their variousmenu items. Comparisons of price changesfor th e th ree items in o ur survey show slightdeclines (-1.5 percent) in the price offrench fries and so da in New Jersey relativeto Pennsylvania, coupled with a relative in-crease (8 percent) in entrCe prices. Theselimited data suggest a possible role for rela-tive price cha nge s within th e fast-food in-dustry following the rise in the minimumwage.O ne way to test a monopsony mo del is toidentify stores that were initially "supply-constrained" in the labor market and testfor employment gains at these stores rela-tive to o ther stores. A potential indicator ofmarket power is the use of recruitmentbonuses. As we noted in Table 2, about 25percent of stores in wave 1 were offeringcash bonuses to employees who helped finda new worker. We compared employment


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792 THE AMERICAN ECONOM IC REVIEW SEPTEMBER 1994

changes at New Jersey stores that were of-fering recruitment bonuses in wave 1, andalso interacted the GAP variable with adummy for recruitment bonuses in severalemployment-change models. We do not findfaster (or slower) employment growth at theNew Jersey stores that were initially usingrecruitment bonuses, or any evidence thatthe GAP variable had a larger effect forstores that were using bonuses.

IX. ConclusionsContrary to the central prediction of thetextbook model of the minimum wage, but

consistent with a number of recent studiesbased on cross-sectional time-series com-parisons of affected and unaffected marketsor employers, we find no evidence that therise in New Jersey's minimum wage reducedemployment at fast-food restaurants in thestate. Regardless of whether we comparestores in New Jersey that were affected bythe $5.05 minimum to stores in easternPennsylvania (where the minimum wage wasconstant at $4.25 per hour) or to stores inNew Jersey that were initially paying $5.00per hour or more (and were largely unaf-fected by the new law), we find that theincrease in the minimum wage increasedemployment. We present a wide variety ofalternative specifications to probe the ro-bustness of this conclusion. None of thealternatives shows a negative employmenteffect. We also check our findings for thefast-food industry by comparing changes inteenage employment rates in New Jersey,Pennsylvania, and New York in the yearfollowing the increase in th e minimum w age.Again, these results point toward a relativeincrease in e mploy ment of low-wage work-ers in New Jersey. W e also find no evidencethat minimum-wage increases negativelyaffect the number of McDonald's outletsopened in a state.Finally, we find that prices of fast-foodmeals increased in New Jersey relative toPennsylvania, suggesting that much of theburden of the minimum-wage rise waspassed on to consumers. Within New Jer-sey, however, we find no evidence th at pricesincreased more in stores that were most

affected by the minimum-wage rise. Takenas a whole, these findings are difficult toexplain with th e stand ard competitive modelor with models in which employers facesupply constraints (e.g., monopsony or equi-librium search models).

R E F E R E N C E SBrown, Charles; Gilroy, Curtis and Kohen, An-

drew. "The Effect of the Minimum Wageon Employment and Unemployment."Jou rnal of Economic Literature, Jun e 1982,20(2), pp. 487-528.. "Time Series Evidence on the Ef-fect of the Minimum Wage on YouthEmployment and Unemployment." Jour-nal of Human Resources, Winter 1983,18(1), pp. 3-31.

Burdett, Kenneth and Mortensen, Dale T."Equilibrium Wage Differentials andEmployer Size." Cen ter for MathematicalStudies in Economics and ManagementScience Discussion Paper No. 860, North-western University, October 1989.Bureau of National Affairs. Daily Labor Re-

port. Washington, DC: Bureau of Na-tional Affairs, 5 May 1990.. Lab or Relations Reporter Wages an dHours Manual. Washington, DC: Bureauof National Affairs, irregular.Card, David. "Using Regional Variation inWages To Measure the Effects of theFederal Minimum Wage." Industrial andLabor Relations Reuiew, October 1992a,46(1), pp. 22-37.. "Do Minimum Wages Reduce Em-ployment? A Case Study of California,1987-89." Industrial and Labo r RelationsReuiew, October 1992b, 46(1), pp. 38-54.Card, David and Krueger, Alan B. "MinimumWages an d Em ployment: A C ase Study ofth e Fast Food Industry in New Jersey andPennsylvania." National Bureau of Eco-nomic Research (Cambridge, MA) Work-ing Paper No. 4509, October 1993.Katz, Lawrence F. and Krueger, Alan B. "TheEffect of the Minimum Wage on the FastFood Industry." Industrial and Labor Re-lations Reuiew, October 1992, 46(1), pp.6-21.


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793V O L . 84 NO . 4 CARD AND KRUEGER: MINIMUM WAG E AND EMPLOYMENT

Lester, Richard A. "Employment Effects ofMinimum Wages." Industrial and LaborRelations Reuiew, January 1960, 13, pp.254-64.. The economics of labo r, 2nd E d.New York: Macmillan, 1964.

Machin, Stephen and Manning, Alan. "TheEffects of Minimum Wages on Wage Dis-persion and Employment: Evidence fromthe U.K. Wage Councils." Industrial andLabor Relations Reuiew, January 1994,47(2), pp. 319-29.McDonald's Corporation. 1991 An nual report.Chicago, 1991.Mincer, Jacob and Leighton, Linda. "The Ef-fects of M inimum W ages on H um an C ap-ital Formation," in Simon Rottenberg,ed ., The economics of legal minimumwages. Washington, DC : American En ter-prise In stitute, 1981, pp. 155-73.Mortensen, Dale T. "Equilibrium Wage Dis-tributions: A Synthesis." Center forMathematical Studies in Economics and

Management Science Discussion PaperNo. 811, Northwestern University, March1988.Ransom, Michael R. "Seniority and Monop-sony in the Academic Labor Market."American Economic Reuiew, March 1993,83(1), pp. 221-33.Stigler, George J. "The Economics of Mini-mum Wage Legislation." American Eco-nomic Reciew, June 1946, 36(3), pp.358-65.Sullivan, Daniel G . "Monopsony Power in theMarket for Nurses." Journal of Law andEconomics, October 1989, Part 2, 32(2),pp. S135-78.U.S. Department of Commerce. 198 7 Census ofretail trade: Miscellaneous subjects. W ash-ington, DC: U.S. Government PrintingOffice, October 1990.Wellington, Alison J. "Effects of the Mini-mum Wage on the Employment Status ofYouths: An Update." Journal of HumanResources, Winter 1991, 26(1), pp. 27-46.


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Minimum Wages and Employment: A Case Study of the Fast-Food Industry in New Jerseyand Pennsylvania

David Card; Alan B. Krueger

The American Economic Review, Vol. 84, No. 4. (Sep., 1994), pp. 772-793.

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[Footnotes]

1 The Effect of The Minimum Wage on Employment and Unemployment

Charles Brown; Curtis Gilroy; Andrew Kohen

Journal of Economic Literature, Vol. 20, No. 2. (Jun., 1982), pp. 487-528.

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1 Time-Series Evidence of the Effect of the Minimum Wage on Youth Employment andUnemployment


The Journal of Human Resources, Vol. 18, No. 1. (Winter, 1983), pp. 3-31.

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1 Effects of the Minimum Wage on the Employment Status of Youths: An UpdateAlison J. Wellington


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3 The Effect of the Minimum Wage on the Fast-Food Industry

Lawrence F. Katz; Alan B. Krueger

Industrial and Labor Relations Review, Vol. 46, No. 1. (Oct., 1992), pp. 6-21.

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11The Effect of the Minimum Wage on the Fast-Food Industry

Lawrence F. Katz; Alan B. KruegerIndustrial and Labor Relations Review, Vol. 46, No. 1. (Oct., 1992), pp. 6-21.

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32 Monopsony Power in the Market for Nurses

Daniel Sullivan

Journal of Law and Economics, Vol. 32, No. 2, Part 2, Empirical Approaches to Market Power: AConference Sponsored by the University of Illinois and the Federal Trade Commission. (Oct., 1989),pp. S135-S178.

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32 Seniority and Monopsony in the Academic Labor Market

Michael R. Ransom

The American Economic Review, Vol. 83, No. 1. (Mar., 1993), pp. 221-233.

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References

The Effect of The Minimum Wage on Employment and Unemployment


Journal of Economic Literature, Vol. 20, No. 2. (Jun., 1982), pp. 487-528.

Stable URL:

http://links.jstor.org/sici?sici=0022-0515%28198206%2920%3A2%3C487%3ATEOTMW%3E2.0.CO%3B2-C

Time-Series Evidence of the Effect of the Minimum Wage on Youth Employment andUnemployment



Stable URL:

http://links.jstor.org/sici?sici=0022-166X%28198324%2918%3A1%3C3%3ATEOTEO%3E2.0.CO%3B2-Q

Using Regional Variation in Wages to Measure the Effects of the Federal Minimum Wage

David CardIndustrial and Labor Relations Review, Vol. 46, No. 1. (Oct., 1992), pp. 22-37.

Stable URL:

http://links.jstor.org/sici?sici=0019-7939%28199210%2946%3A1%3C22%3AURVIWT%3E2.0.CO%3B2-3

Do Minimum Wages Reduce Employment? A Case Study of California, 1987-89

David Card


Stable URL:

http://links.jstor.org/sici?sici=0019-7939%28199210%2946%3A1%3C38%3ADMWREA%3E2.0.CO%3B2-%23

The Effect of the Minimum Wage on the Fast-Food Industry



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Employment Effects of Minimum Wages

Richard A. Lester

Industrial and Labor Relations Review, Vol. 13, No. 2. (Jan., 1960), pp. 254-264.

Stable URL:

http://links.jstor.org/sici?sici=0019-7939%28196001%2913%3A2%3C254%3AEEOMW%3E2.0.CO%3B2-3

The Effects of Minimum Wages on Wage Dispersion and Employment: Evidence from the

U.K. Wages CouncilsStephen Machin; Alan Manning

Industrial and Labor Relations Review, Vol. 47, No. 2. (Jan., 1994), pp. 319-329.

Stable URL:

http://links.jstor.org/sici?sici=0019-7939%28199401%2947%3A2%3C319%3ATEOMWO%3E2.0.CO%3B2-7

Seniority and Monopsony in the Academic Labor Market

Michael R. Ransom

The American Economic Review, Vol. 83, No. 1. (Mar., 1993), pp. 221-233.

Stable URL:

http://links.jstor.org/sici?sici=0002-8282%28199303%2983%3A1%3C221%3ASAMITA%3E2.0.CO%3B2-J

The Economics of Minimum Wage Legislation

George J. Stigler

The American Economic Review, Vol. 36, No. 3. (Jun., 1946), pp. 358-365.

Stable URL:

http://links.jstor.org/sici?sici=0002-8282%28194606%2936%3A3%3C358%3ATEOMWL%3E2.0.CO%3B2-A

Monopsony Power in the Market for Nurses

Daniel Sullivan

Journal of Law and Economics, Vol. 32, No. 2, Part 2, Empirical Approaches to Market Power: AConference Sponsored by the University of Illinois and the Federal Trade Commission. (Oct., 1989),pp. S135-S178.

Stable URL:

http://links.jstor.org/sici?sici=0022-2186%28198910%2932%3A2%3CS135%3AMPITMF%3E2.0.CO%3B2-1

Effects of the Minimum Wage on the Employment Status of Youths: An Update

Alison J. Wellington


Stable URL:

http://links.jstor.org/sici?sici=0022-166X%28199124%2926%3A1%3C27%3AEOTMWO%3E2.0.CO%3B2-5


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