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The Response of Interest Rates to Unexpected Weekly Money: Are Policy Changes Important?* R. W. HAFER Southern Illinois University at Edwardsville Edwardsville, Illinois

RICHARD G. SHEEHAN Notre Dame University Notre Dame, Indiana

I. Introduction

The evidence presented in prior studies of the interest rate-weekly money relationship generally indicates that financial markets react only to unexpected money changes. The data also support the hypothesis that the response of interest rates to the money announcement significantly changed in October 1979 and in October 1982.

On 6 October 1979 the Federal Reserve announced that it would henceforth place more

policy emphasis on the behavior of the money stock and reduce the weight given to fluctuations in the federal funds rate. In October 1982 the Fed again changed its operating procedure, shifting away from targeting the monetary aggregates to emphasizing the importance of discount window

borrowing. The evidence to date from almost every study rejects the hypothesis of a stable money announcement interest rate relationship in favor of the alternative that both October policy regime changes represent structural breaks in the estimated regressions.

Reading through this growing literature, we were struck by the noticeable absence of tests for breaks in the regressions other than the above-mentioned regime shifts.' Previous researchers

generally have ignored the potentially destabilizing impacts stemming from the numerous financial innovations that occurred after 1979 such as the redefinition of the monetary aggregates in 1980 and the nationwide legalization of NOW accounts in 1981. The effect of policy events like the

special credit control program of 1980 and shifts in the weight attached to M1 as a policy target also have not been adequately treated in tests of structural stability. Given the numerous financial

changes occurring between 1980 and 1982, it should not be surprising that selecting a break point

*We would like to thank Jerry Dwyer, Barry Falk, Gikas Hardouvelis, Scott Hein, Doug Pearce, participants in seminars at the 1986 Southern Economics Association and at Washington University-St. Louis and an anonymous referee for comments on an earlier version. Nancy Juen provided excellent research assistance. An earlier version of this paper was completed while Hafer was at the Federal Reserve Bank of St. Louis. The usual disclaimer applies.

1. This statement should be tempered somewhat. As we will show in section III, Roley [21; 22] tested for a break in February 1980 when the money announcement was moved from Thursday to Friday and when the expected money survey was changed from twice a week (Tuesday and Thursday) to once a week (Thursday). In addition, several others [4; 14; 17] use approaches similar to that employed here.

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578 R. W. Hafer and Richard G. Sheehan

immediately before this period (say, October 1979) or after (say, October 1982) produces results that suggest these to be the true breaks, even if they are not. This result could easily occur if the

technique used to detect structural change, such as the standard F-test, is not sensitive to break

point selection.2 Our purpose in this paper, therefore, is to re-examine the temporal stability of the interest

rate-weekly money relationship. To this end, we use a time varying parameter estimation technique in conjunction with Quandt's [18; 19] procedure to detect structural change. The time-varying parameter approach is useful because, in addition to testing for overall parameter stability, it

provides a plot of the estimated parameter's evolution across the sample period. Since our sample runs from September 1977 through December 1985, this visual aid allows us to ascertain whether the behavior of the parameter changes appreciably at the time of policy regime changes. Indeed, applying this procedure to a collection of interest rates across the maturity spectrum, we find that the most likely break points occur at times other than October 1979 and October 1982. Based on evidence from several sample periods that permit the data to isolate either October regime change as the most likely break point, the results from the Quandt procedure also reject the contention that the relationship between interest rates and unanticipated changes in weekly M1 was significantly affected by apparent changes in monetary operating procedures.

The remainder of the paper is structured as follows. The estimated equation and data are

presented in section II. Section III contains ordinary least squares estimates of the model for the full period and for relevant subperiods. It provides a basis to compare our estimates with previous results. We also present time varying parameter and Quandt test results. The paper closes with our conclusions.

II. The Model and Data

The relation between changes in interest rates and the weekly money announcement generally has been analyzed by estimating the equation:

Ait = p0 + 3UM, + t, (1)

where A i is the change in the interest rate and UM is the unexpected change in the money stock

(M1). Researchers often use only the three-month T-bill rate to measure the effects of unanticipated

money announcements on short-term rates. In our empirical work we investigate the temporal behavior of /, using the entire maturity spectrum of interest rates as dependent variables. We examine the effect on changes in the federal funds rate and the three-month T-bill rate. To determine the impact of unexpected M1 announcements on longer-term rates we use Shiller, Campbell and Schoenholtz's [27] procedure to generate seven implicit forward rates.3 This approach ensures that results for long-term securities do not merely reflect the response of embedded shorter-term

2. Farle. Hinich and McGuire [11] point out that the power of the Chow test declines rapidly when the shift in the relation does not occur at the midpoint of the sample. In terms of testing the 6 October 1979 break, such an effect

may be important. For example, based on the sample used in this paper, breaking the sample at 6 October 1979 produces samples of 105 and 322 weeks.

3. The implicit forward rates used are 3-6 month, 6-12 month, 1-2 year, 2-3 year, 3-5 year, 5-7 year, 7-10 year, 10-20 year and 20-30 year. Although the correct description of, say, the 20-30 year forward rate is the implicit return on a 10 year bond 20 years hence, we use the grammatically easier 20-30 year forward rate in our textual discussion.

INTEREST RATES AND UNEXPECTED MONEY

rates. In all cases, the dependent variable is measured from the market close on the day of the

money announcement to the close on the following business day. While we would prefer to measure the interest rate response from just before to immediately following the announcement, data

availability for rates on maturities longer than three months dictates that we use the close-to-close

change in rates. In this regard, we are fully cognizant of the criticisms raised by, among others, Falk and Orazem [10] and Belongia and Sheehan [3].

The unexpected change in money is generated using the expected money series taken from the

Money Market Services, Inc. weekly survey of market participants. We use the median forecast of this survey to measure expected money. The expected change is subtracted from the actual

change to construct the unexpected change in M1. Actual changes in M1 taken from the Federal Reserve's H.6 statistical release are calculated as first announced less first revised values. Because the definition of Ml changes across the sample, the following procedure is used. The "old" definition of Ml is employed until February 1980. From that point through November 1981 the definition is M1B not adjusted for NOW accounts. The current definition of M1 is used from November 1981 to the end of the sample. This approach yields a measure of M1 that corresponds to the measure survey respondents were asked to forecast. Note, however, that this Ml was not the only narrowly-defined monetary aggregate announced by the Fed at various times during the

sample. There were, for example, "shift-adjusted" M1B and M A measures.4

Equation (1) differs noticeably from some other specifications by omitting expected money (EM) from the right hand side. We omit EM because we focus exclusively on the temporal impact of unexpected money to determine the importance of policy regime shifts in explaining previously noted structural breaks. Some evidence also exists suggesting that the estimated effect of unex-

pected money is not sensitive to the specification used. Belongia and Sheehan [3] test alternative versions of equation (1). While they investigate the significance of expected money and the inter-

pretation of this result, they also find that the estimated coefficient on unexpected money appears to be unaffected when expected money is included. Because unexpected money's coefficient does not appear to be sensitive to the inclusion or exclusion of expected money and since we are not

specifically concerned with re-testing the efficient markets hypothesis, we estimate equation (1).5

III. Empirical Results

Ordinary Least Squares Estimation

Before turning to the empirical results, it is useful to consider the popularity of the October break

points in the existing literature. To do this, we collected a sample of previous studies of the interest

rate-weekly money relationship.6 The authors of these articles are listed in Table I, along with the

sample periods examined and the break points tested. October 1979 and October 1982 commonly are cited as policy regime changes. The "other" category illustrates the non-policy related breaks that have been tested. The number of candidates in the "other" category are almost non-existent.

4. The "shift-adjustment" to M1B was done to correct for inflows to NOW accounts that originated in non-Mi assets. The basis for this adjustment was survey data accumulated by the Board of Governors. Bennett [5] provides a discussion. Additional evidence and discussion is found in Broaddus and Goodfriend [6] and Rasche and Johannes [20].

5. We did experiment with a version of equation (1) that includes expected money and found that our results were little affected. To better focus the paper on time-varying coefficients, however, we report the evidence from equation (1).

6. Numerous other articles exist dealing with the impact of weekly MI on stock prices, exchange rates and other financial variables. Sheehan [26] surveys this literature.

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Table I. Summary of Empirical Results: Break Points Tested

Potential Break Points Tested

Author Sample Period October 1979 October 1982 February 1984 Other

Cornell [7] Jan 78-Aug 81 Yes N.A. N.A. None Deaves, Melino and Pesando [9] Oct 79-Sept 85 N.A.a Yes Yes None Deaves [8] Oct 82-Sept 85 N.A. N.A.a Yes None Hardouvelis [13] Oct 77-Nov 82 Yes No N.A. None Hein [14] Oct 79-Oct 83 N.A.a Yes N.A. Ashley Testb Huizinga-Leiderman [15] Oct 79-Feb 84 N.A.a Yes N.A.c None Judd [16] Sept 77-Feb 84 Yes Yes N.A.c None Loeys [17] Nov 77-Dec 83 Yes No N.A. Rolling-sampled Roley [21] Sept 77-Nov 81 Yes N.A. N.A. Feb 80c Roley [22] Sept 77-Oct 82 Yes N.A.c N.A. Feb 80e Roley-Walsh [24, 25] Sept 77-Oct 82 Yes N.A.c N.A. None

a. Sample start based on previous evidence of regime shift. b. Sample is divided into 10 subsamples. Evidence suggests weakening of relationship beginning mid-1982. c. Sample ends based on previous evidence of regime shift. d. In addition to October 1979, procedure suggests possible breaks for different interest rates. The (forward) rates and additional breaks are:

3-month (August 1981); 3-12 month (December 1981); 1-3 year (December 1981); 3-7 year (January 1981 and June 1982); and 7-30 year (August 1981 and June 1982).

e. February 8, 1980 marks the time that (a) Money Market Services switched from a semi-weekly (Tuesday and Thursday) survey to a weekly survey (Tuesday only), and (b) the date when the Federal Reserve moved the announcement of Ml from Thursday to Friday.


Except for Loeys [17] and Hein [14], both of whom use procedures that do not require a priori specification of the breaks, only Roley [21; 22] specifically tests for an alternative break in the

relationship. It is more troubling that several studies simply begin [8; 9; 14; 15], or end [15; 16; 24; 25] their sample based on the maintained hypothesis of a regime change. We ask whether the data actually supports such an a priori selection of estimation periods.

For purposes of comparison it is useful to first estimate equation (1) by ordinary least squares (OLS) using the popular subperiods: pre-October 1979, October 1979 through September 1982 and post-September 1982. The estimation results for the different interest rates are presented in Table II. The results are in accord with previous evidence that the estimated slope coefficients

(jl1 's) generally are larger during the 1979-82 sample than before or after. Moreover, during the 1979-82 subperiod the coefficient on unexpected money is significant (5 percent level) for all but the longest forward rates, compared with only 4 of 11 significant based on the pre-October 1979 data. Evidence from the post-1982 period, however, indicates that the coefficient on unexpected money is insignificant in the federal funds rate and long forward rate equations. The estimated

magnitude of the effect for the 1979-82 period generally is larger the shorter the maturity of the interest rate. In contrast, the results for the pre-1979 period show a mix of effects across the

maturity of rates (both in significance and magnitude), while the post-1982 evidence suggests similar size impacts across the maturity spectrum.

One notable aspect of the OLS evidence is the consistent finding that unexpected changes in M1 do not influence the 7-10, 10-20 and 20-30 year forward rates. This result is consistent with other studies that have used implicit forward rates [13; 16; 27], but contradicts those using actual long-term rates [7].7 Because actual long-term rates overlap shorter-term rates, the evidence

using the forward rates arguably provides clearer evidence on the impacts of unexpected changes in weekly money on long-term rates. Based on the evidence in Table II and in previous studies, unexpected money appears to have no effect on the longer-term implicit forward rates. Conse-

quently, in the following analysis we only examine the time varying response of rates up to the 5-7 year forward rate.

Time Varying Parameter Estimation

The temporal stability of unexpected money's effect on our set of interest rates is analyzed using the time-varying parameter model suggested by Garbade [12]. The procedure provides a useful

approach to explore the changing response of market rates to announced or perceived changes in

policy regime. Since it does not impose a priori breaks in the estimated sample, this procedure has the flexibility to assess the market's response not only to announced policy changes, but also to the other institutional and regulatory changes that occurred during the period. Plotting the time

varying coefficient for unexpected money across our sample will be of special interest.8 To estimate the time varying parameter model, equation (1) is rewritten as:

Ait = 0,t +1 I,tUMt + vt, (2)

7. We also estimated the equations using actual spot rates. Doing so yielded a statistically significant coefficient on unexpected money for the 1979-82 and 1982-85 samples for all interest rates.

8. The time varying parameter approach also has been used to test the hypothesis that the change in policy procedures in October 1979 accounts for the apparent rise and increased volatility in the short-term real rate of interest [1].

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Table II. Estimates (OLS) of Equation (1): Ait = fo + P, UMt + Et, Alternative Sample Periodsa

29 Sept. 1977 - 4 Oct. 1979 11 Oct. 1979 - 29 Sept. 1982 1 Oct. 1982 - 27 Dec. 1985 Rate b /3 ' 2 -2 -2 Rateb tPo 31 R PQ3i0 R30 /31 R

Fed funds .008 (.84) .003 (.46) .00 .034 (.64) .086 (3.58) .07 -.032 (1.47) .010 (.83) .00 3-month .028 (2.93) .017 (2.68) .06 .045 (1.59) .079 (6.18) .20 .007 (.83) .020 (4.40) .10 3-6month .015 (2.27) .011 (2.69) .06 .044 (1.76) .070 (6.13) .19 .006 (.55) .026 (4.57) .11 6-12 month .020 (3.05) .015 (3.58) .10 -.009 (.53) .052 (6.41) .21 -.012 (1.37) .020 (4.09) .09 1-2 years .010 (1.59) .002 (.43) .00 .011 (.45) .071 (6.45) .21 .002 (.20) .022 (3.77) .07 2-3 years .003 (.38) .010 (2.05) .03 .020 (.92) .049 (5.03) .19 -.003 (.28) .016 (2.36) .03 3-5 years -.001 (.10) .007 (1.70) .02 .057 (3.35) .022 (2.90) .05 .004 (.38) .022 (3.93) .08 5-7 years .004 (.53) -.003 (.04) .00 .027 (1.47) .026 (3.18) .06 .013 (1.08) .018 (2.71) .04 7-10 years .005 (.59) -.009 (1.71) .02 .049 (2.60) -.011 (1.23) .00 -.030 (.99) .023 (1.36) .01 10-20 years .003 (.53) .001 (.25) .00 .043 (2.04) .009 (.94) .00 .020 (.97) -.003 (.24) .00 20-30years -.100 (1.27) -.041 (.83) .00 -.060 (.72) .054 (1.44) .01 -.033 (1.64) -.003 (.23) .00

Absolute values of t-statistics appear in parentheses. a. The number of observations are 105, 153 and 169, respectively. b. The interest rates used are forward rates, except for the federal funds rate and the three-month Treasury bill rate. Forward rates are calculated

using the Shiller, Campbell, and Schoenholtz [27] linearization procedure.


Table III. Time Varying Parameter Estimation Results, Sample Period: 29 September 1977 - 27 December 1985

Rate ML (OLS) ML x2 (ld.f.) "ML OLS p

Fed funds 348.906 352.846 7.88* .178 .189 .0006 3-month 599.319 608.003 17.37* .054 .057 .0006 3-6 month 661.692 667.924 12.46* .042 .044 .0003 6-12 month 785.395 792.277 13.76* .023 .025 .0005 1-2 years 593.996 607.079 26.17* .053 .060 .0014 2-3 years 649.700 654.719 10.04* .044 .046 .0006 3-5 years 740.262 - - 0 5-7 years 653.133 663.726 21.19* .037 .046 .0080

*Indicates significance at the 1 percent level.

where the time path of,/ 1, is assumed to be a random walk process without drift; that is,

P1 t = PI I,t-1 Pi,t where p ,t is distributed N(0, o2p). Constraining the conditioning P-matrix to zero reduces the estimation to OLS. To test for parameter stability, a maximum likelihood (ML) estimate of equation (2) is calculated for the case in which P is constrained to zero and compared to a ML estimate without the constraint. Rejecting the restriction P = 0 implies rejecting the

hypothesis of constant coefficients. The relevant stability test results are reported in Table III. Comparing the constrained esti-

mates (ML(OLS)) to the unconstrained (ML) estimates yields a x2 statistic.9 The evidence re-

ported in Table III indicates that the hypothesis of a constant coefficient on unexpected money is

rejected at very high levels for all short and medium term forward rates except the 3-5 year rate. For the 3-5 year rate the ML procedure yields the OLS result. Thus, like previous studies, the

evidence in Table III indicates that interest rates generally have responded differently over time to

unexpected money changes. The time varying parameter estimation procedure also provides a visual portrayal of the

parameter's evolution over time. Inspection of the plots of/31, allows us to gauge whether changes in policy regime and changes in the response of interest rates coincide. More specifically, if

marked movements in the estimated coefficient on unexpected money do not occur at the popular October 1979 and October 1982 break points, then we should re-evaluate previous evidence

supporting the traditional regime-shift view.

The time-varying behavior of the coefficients on unexpected money are depicted in Figures 1-7, with the ordering of the figures corresponding to Table III. The bands in the figures cor-

respond to a one standard deviation confidence interval about the estimated coefficients. Given

the evidence in Table III, the 3-5 year forward rate is excluded. The plots for the federal funds

rate and the 3-month T-bill rate corroborate one aspect of both previous research and the OLS results in Table II: Unexpected money had a greater average impact during the period 1979-82 than either before or after. Indeed, the finding of greater response during 1979-82 is consistent across rates.

The time series of the estimated UM coefficients illustrated in Figures 1-7 do not support the popular notion that October 1979 marked a significant shift in the impact of unexpected

9. The test procedure is based on a concentrated log-likelihood function that produces a maximum likelihood (ML) estimator for P and a time series of the coefficients. The ML value P is compared to the OLS value using the statistic -2(L(Po) - L(P)) where L(Po) is the likelihood function estimated at P = 0 and L(P) is the likelihood function evaluated at the ML value of P. The resultant statistic is distributed as a X 2 with one degree of freedom.

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A um

0.18 ? i

0.16 ,,i

, , g i

0.14 - ,s *

0.08 : :

0.02 /// V'

0.00, '

" ...

-0.02

,,-

-0.04 -

-0.06

Jan 77 Jan 79 Jan 81

Vertical lines represent Oct. 6, 1979 & Oct. 15, 1982.

Figure 1. Federal Funds Rate

J a 7 7 Ja 9Jn8 Vericl

lne

r e rsn Oc.6 97 c. 5 92 F~i g ure1 FdralFnsRt

A um

0.12

Vertical lines represent Oct. 6, 1979 & (

Figure 2. 3-Month T-Bill

Jan 87

Jan 77 Jan 79 Jan 81 Vertical lines represent Oct. 6, 1979 & Oct. 15, 1982.

Figure 3. 3-Month to 6-Month Forward Rate

A um

0.08

-0.02 -


Figure 4. 6-Month to 1-Year Forward Rate

EST RATES AND UNEXPECTED MONEY 585

X,*.

\

''""

..,

Jan 83 Jan 85 Jan 87


A um

0.16

0.12

0.08 a /- 0 0

0.08- / f/ ". ,

i

0.00 I

-0.04 - /

a a

-0.12-

-0.16-


Figure 5. 1-Year to 2-Year Forward Rate

A um

0.10



Jan 83 Jan 85

87

Jan 87


Jan 77 Jan 79 Jan 81 Jan 83 Jan 85 Jan 87


money on interest rates. In every instance the coefficient on unexpected money begins to increase well before October 1979. For example, when the change in federal funds rate is the dependent variable, the estimated coefficient on unexpected money increases from 0.031 in January 1979 to 0.068 just before the announced October 1979 policy shift. Similarly, the impact of unexpected money on the 3-month T-bill rate (Figure 2) more than doubles between mid-1978 and September 1979. Indeed, this general pattern prevails for every interest rate examined.'0 The hypothesis that the change in the empirical relationship occurred at the October 1979 policy shift is contradicted

by the time-varying parameter results. Not only does the change occur earlier, but also there is no clear evidence of an aberration in the estimated coefficients in October 1979.

Figures 1-7 also clearly indicate no significant change in the behavior of the estimated UM coefficients in October 1982. While the coefficients do decline somewhat at this point, the reductions generally begin much earlier than the October 1982 change in operating procedures. For example, Figures 2, 3 and 4 reveal decreases in the coefficients beginning in 1981 for the 3-month T-bill rate, the 3-6 month and 6-12 month forward rates, respectively. Moreover, the coefficient magnitudes slowly decay from as early as 1980 using the 1-2 and 2-3 year forward rates (Figures 5 and 6). The 5-7 year forward rate coefficient in Figure 7 changes little after an initial run up in early 1980.

The response of the federal funds rate (Figure 1) appears to be much more variable across the sample than that of other rates. The size of the coefficient drops sharply following a peak in

10. Although Loey's [17] graphical evidence suggests some upward movement in the coefficient on UM before October 1979, his rolling-regression procedure makes it difficult to accurately time the shift. For example, the largest impact of unexpected money on the 3-month T-bill occurs for the sample that runs from May 1979 through April 1980. Although this peak effect is attributed to the October policy shift, the sample also includes (1) the February 1980 change in announcement days and consequent change in the MMS survey and (2) the beginning of the special credit control program.

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early 1980, rises again from mid-1981 through early 1982 and decays from that point through the end of the sample period. Although the timing differs somewhat from the other rates, the response of the funds rate apparently also does not change with the October 1982 change in operating procedures."

The visual evidence in Figures 1-7 demonstrates that the shifts in monetary policy operating procedures do not coincide with changes in the effect of unexpected money on interest rates. What then could explain the behavior depicted in the figures? The initial increase in the estimated coefficients begins in mid-1978 for all rates except the 1-2 year and 2-3 year forward rates.

Beginning in 1979, the Fed abandoned its quarterly setting of monetary growth targets and instituted the annual growth targets required under the Humphrey-Hawkins Act of 1978. This

change forced the Fed to set its targets from the fourth quarter to the fourth quarter to lessen the problem of base drift inherent in its previous approach.'2 Even though the Fed had stated

monetary growth targets prior to January 1979, the 1978 Congressional passage of a new operating constraint clearly represented a change in the Fed's policy environment. The increasing response of interest rates to unexpected money changes following this procedural change suggests that financial market participants placed increasing importance on monetary developments before the October 1979 change in operating procedures.

The figures also indicate that money announcements have their peak effects in early 1980 for all rates except the 3-6 month forward rate; its peak occurs in 1981. The early 1980 peak coincides with three events of possible significance for monetary policy. First, the special credit control program was initiated by the Carter administration with its implementation by the Fed in March 1980 and its phase-out in July 1980.13 Second, on 7 February 1980, the Fed also announced new definitions of the monetary aggregates, designating two narrow measures of money-MlA and M1B-even though policy continued to be discussed primarily in terms of M1B.'4 And third, Congress enacted the Depository Institution Deregulation and Monetary Control Act in March 1980, legislation that would have wide ranging effects on the overall financial environment. The market's increased response to unexpected money changes in early 1980 could be due to any combination of these three changes.

Another important event that coincides with substantial changes in the estimated UM coefficient is the period in early 1981 that encompasses the nationwide introduction of NOW accounts.'5

11. Roley [23] provides a theoretical model that suggests that the switch in February 1984 from lagged to contem-

poraneous reserve accounting provides another plausible break in the empirical relationship. The evidence in the time series plots show no aberration of the coefficients path that supports this hypothesis. Note, however, that using February 1984 as a break point between October 1982 and December 1985 may yield a significant result based on a standard F-test

procedure given the time path of the coefficients. 12. For a useful discussion of base drift and monetary targeting by the Fed, see [6]. 13. The special credit control program effectively increased institutions' borrowing costs through the imposition of

a discount rate surcharge on large banks that were frequent discount window borrowers and influenced the flow of credit between institutions. The impact of the controls on the behavior of money was dramatic. During April 1980, M1A and M1B declined at annual rates of 18.5 percent and 14.5 percent and showed essentially no growth in May.

14. The new monetary definitions were announced because recent financial innovations were blurring the distinction between transactions and savings deposits. Policy deliberations by the FOMC consequently were set forth in terms of M A, M 1B and M2. Because respondents to the Money Market Services survey only were asked to forecast MIB changes, the increased sensitivity of interest rates to unexpected money changes may reflect increased uncertainty by market

participants over which monetary measure was important in Fed policy discussions. Indeed, little has been written on this

important institutional aspect, even though it clearly may affect the stability of the weekly interest rate-money relationship. 15. To our knowledge, no other study has examined the effects of this financial innovation on the money announce-

ment-interest rate relationship. As with the announcement of the new monetary aggregates in February 1980, the nationwide legalization of NOW accounts produced yet another monetary aggregate discussed in Fed policy deliberations-


This effect is especially noticeable for the shorter-term rates. For example, using the often-studied 3-month T-bill rate, a sharp increase in the coefficient occurs from the beginning of 1981 through mid-1981 when the coefficient begins a descent that continues through the remainder of the

sample.16 For both the 3-6 month and the 6-12 month forward rates, the episode of the NOW account introduction is associated with the beginning of the decline in the estimated coefficient on unexpected money. The decline in the estimated coefficients also corresponds to the period when Ml was being downplayed as a policy target.l7 Thus, the decline may be due to the Federal Reserve changing weights on alternate monetary aggregates or to financial market participants changing their interpretation of unexpected deviations of the monetary aggregates.

The results in Figures 1-7 may be due to the increased variability of interest rates immedi-

ately after October 1979 rather than being due to differential impacts of monetary policy. The time

varying parameter procedures described in equation (2) can be modified to deal with heteroscedas-

ticity. The varying parameter procedure employed above maximizes the likelihood function with

respect to the k diagonal elements of the P matrix. This matrix is assumed constant over time, thus imposing homoscedasticity on the time varying process. If a break occurred, that is, if we allow heteroscedasticity due to a single change in the variance at a known point, then we would need to search over 2k parameters, the k diagonal elements before and after the break. If the break point is unknown, we must search for 2k + 1 parameters. With n unknown breakpoints, the number of parameters in the nonlinear search increases to (n + 1)k + n.

Given the time consuming nature of the numerical procedure required to achieve conver-

gence, it is unfortunately impractical to allow even a single breakpoint. As an alternative, we

simply constrained the values for pi in the post-October 1979 period to be double their earlier values and then re-optimized. This assumption is equivalent to doubling the variance after Octo- ber 1979. This modification leaves the basic results unaffected, and the plots are not noticeably altered.'8

In summary, Figures 1-7 provide little support for the hypothesis that the October 1979 and October 1982 policy regime changes correspond to structural breaks in the estimated interest rate-

weekly money relationship. Based on the time varying parameter results, the points of change

namely, "shift-adjusted" MlB. Because market participants again did not know with certainty which measure domi- nated policy decisions, the apparent changing response of market rates to unexpected money (M1B) changes may not be

surprising. 16. It is interesting to note that the apparent portfolio shift that gave rise to the construction of "shift-adjusted"

M1B had probably run its course by April 1981. This is indicated by comparing the monthly growth rates of M1B and

shift-adjusted M1B during 1981: From December 1980 through April 1981, the growth rates diverge, on average, by 7 percentage points. During the next several months (April-October), in contrast, the average difference is less than 1

percentage point. Since the portfolio shifting appeared to have been completed by mid-1981, the Fed discontinued its

shift-adjusted M1B series in January 1982. In making this announcement, the Fed also renamed M1B to MI and ceased

publication of M A. 17. In particular, short-run Ml targets were eliminated at the FOMC's 5 October 1982 meeting. Nevertheless, only

M1 numbers were released weekly during the period covered in this study, and to the extent that they convey a signal to market participants about M2 as well as M1, they may influence spot and forward interest rates even if the FOMC is

targeting on M2 rather than M1. 18. As an additional test for the importance of heteroscedasticity on the results, we employ Ashley's [2] stabilogram

test breaking each year into its first and second halves and including eight sets (for eight years) of slope and intercept dummy variables. This procedure assumes that the response differs discretely rather than potentially varying continuously over time. The Ashley procedure uses ordinary least squares. Our results for the stabilogram are entirely consistent with the TVP tests. We mention the stabilogram tests only because they readily can be modified using Bartlett's procedure to account for heteroscedasticity since they are estimated using OLS. This modification results in no appreciable difference in the estimated coefficients or their implications.

589


in the empirical relationship are better aligned with other institutional changes such as the credit control period and the nationwide legalization of NOW accounts.

Quandt Test Results

The time-varying parameter results indicate that changes in the unanticipated money-interest rate

relationship do not occur at times of announced policy regime shifts. Although the coefficient

plots are suggestive, they suffer from one drawback: there is no statistical test associated with the evolution of the parameter estimates that allows one to pin-point the likely break. To further

investigate whether the October 1979 and October 1982 policy shifts are largely responsible for observed changes in the interest rate-weekly money estimates, Quandt's [18; 19] test is used to locate the most likely shift point.19

We test two hypotheses using the Quandt test. First, we consider parameter stability within the commonly used subperiods September 1977 through September 1979, October 1979 through September 1982 and the post-September 1982 sample. Because these periods form a basis of

comparison to previous work, it is useful to determine if the relationship is stable within these

periods. Rejection of the null hypothesis of stability would suggest that the October regime shifts are not unique.

Second, we test for the most likely break point using two periods, one running from Septem- ber 1977 to September 1982 and the other from October 1979 to March 1985, to see if the Quandt procedure isolates either October regime shift.20 If the commonly adopted break is correct, the

Quandt test should single out the 6 October 1979 date as the "most likely" break point in the first

period and the October 1982 policy shift in the second. The Quandt test results are reported in Table IV. For each interest rate the first set of results

tests the hypothesis of stability within subperiods. The second set of results tests the hypothesis that the October regime shifts are the most likely break points.

The results testing the hypothesis of within subperiod stability suggest that we can reject stability in the September 1977-September 1979 and October 1982-March 1985 subperiods for

every interest rate. The most likely break point chosen for the October 1979-September 1982

sample generally is insignificant, however. The exceptions are the results using the 5-7 year forward rate, where the break point of 6 April 1981 is significant at the 10 percent level, and the 3-6 month forward rate where a break point of 23 January 1981 is significant at the 5 percent level. In all other cases, we cannot reject stability during the October 1979 to September 1982

period. We now turn to the issue of the likelihood of breaks in October 1979 and October 1982.

The Quandt test results in Table IV for the September 1977 to September 1982 period report no evidence of a likely break near the 6 October 1979 policy change regardless of the interest rate used. While no unique break point is chosen for every rate, 12 April 1979 is selected as the most

likely break in five out of eight instances.

19. The Quandt test is used to identify the most likely break point under the maintained hypothesis that there is

only one break in the sample. In the table, we also report the outcome of a standard F-test using the point chosen by the Quandt test as the break. The F-test results are presented for heuristic purposes. This is because: (a) The break

point selected by the Quandt procedure is the point that also should maximize the F-value. If the calculated F-statistic is insignificant, then one has relatively strong evidence against a break. If the F-value is large, then one only has met a

necessary condition to argue for instability. (b) Because of the points raised in footnote 2, and because the Quandt test

points generally are not mid-sample points, the power of the F-test may be questioned. We would like to thank Jerry Dwyer for bringing the first point to our attention.

20. The sample was truncated at 21 March 1985 rather than extending until December 1985 based upon maximum number of observations allowed by the Quandt routine employed.


Table IV. Quandt Test Results

Rate Test period Date of max (L) F (sig)

Federal funds 9/77 - 9/79 10/79- 9/82 10/82- 3/85

9/77- 9/82 10/79- 2/85

9/77 - 9/79 10/79 - 9/82

10/82- 12/85

9/77- 9/82 10/79- 3/85

3-month T-bill

3-6 month

6-12 month

9/77 - 9/79 10/79- 9/82

10/82- 12/85

9/77- 9/82 10/79 - 3/85

9/77 - 9/79 10/79- 9/82

10/82- 12/85

9/77 - 9/82 10/79- 3/85

9/77 - 9/79 10/79- 9/82

10/82- 12/85

9/77 - 9/82 10/79- 3/85

9/77- 9/79 10/79- 9/82

10/82- 12/85

1-2 year

2-3 year

5-7 year

9/77 - 9/82 10/79- 3/85

9/77 - 9/79 10/79- 9/82

10/82- 12/85

9/77 - 9/82 10/79- 2/85

4/12/79 5/9/80

5/27/83

5/9/80 8/5/83

4/12/79 10/9/81 3/18/83

4/12/79 2/25/83

9/14/78 1/23/81 3/18/83

1/16/81 1/16/81

10/20/77 7/31/81 3/18/83

4/12/79 10/21/83

4/12/79 10/9/81 3/18/83

4/12/79 10/21/83

4/12/79 5/9/80

3/18/83

4/12/79 10/21/83

4/19/79 4/6/81

6/17/83

11/6/81 8/5/83

.0330

.9193

.0022

.0111

.0544

.0302

.1557

.0142

.0521

.0473

.0116

.0391

.0108

.0057

.0792

.0716

.1444

.0018

.0024

.0268

.0464

.2326

.0062

.0205

.0592

.0505

.8559

.0060

.0091

.0179

.0252

.0747

.0014

.0084

.1404

Studies using periods like October 1979 to March 1985 generally have tested for a sample break in October 1982.21 As shown in Table IV, for no interest rate does Quandt's procedure select

a date in 1982, again providing no evidence to corroborate the use of an October 1982 break

21. See Table I for evidence. More recently, the February 1984 shift to contemporaneous reserve accounting has received attention [15; 16; 23; 25]. In each instance, however, this event is used as justification to terminate the estimation

period.

591


point. The most common break detected is 21 October 1983, which is selected using the 6-12 month, 1-2 year, 2-3 year and 5-7 year forward rates.

The evidence from the Quandt tests corroborates the time varying parameter evidence in two ways. First, the relation between unexpected money and interest rates is unstable over time, a finding that is consistent with virtually all previous research. Second, in no instance does the Quandt test validate the use of either October policy change as the most likely break in the estimated relationship. Like the time varying parameter evidence, the Quandt test results suggest examining other, more likely break points.

IV. Conclusion

The purpose of this paper is to re-examine the importance of the commonly accepted October 1979 and 1982 policy announcements as break points in the empirical relationship between unexpected weekly money and interest rates. Because these dates coincide with monetary policy regime changes, they are plausible choices to assess the varying effects of unexpected money on interest rates. In fact, the results reported in previous studies suggest that the subperiods delineated by the October breaks are statistically different.

The evidence we present suggests a different interpretation of the data. The time varying parameter results indicate that major changes in the evolution of the estimated parameter on unexpected money do not occur in either October 1979 or October 1982. The evidence points to changes in the coefficient's time path that better coincide with various institutional innovations that

transpired during the sample. These include the passage and implementation of the Humphrey- Hawkins Act, the imposition and removal of credit controls, the introduction of NOW accounts, the redefinition of monetary aggregates and changing FOMC emphasis on Ml as a target. To determine if our time varying parameter results are technique specific, we also employed the Quandt procedure. These results again suggest that October 1979 and October 1982 are not the most likely break points.

The rejection of the October policy announcements as break points raises an important and broader question. How credible are announced policy shifts in explaining the behavior of financial variables? The results presented here suggest that they have relatively little influence.

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