On currency substitution and money demand instability

ON CURRENCY SUBSTITUTION AND

MONEY DEMAND INSTABILITY

ALI F. DARRAT, AHMED AL-MUTAWA,

and OMAR M. BENKATO

ABSTRACT

Based on the asset portfolio-balance theory and on modeling in the context of cointegrated systems, this

paper develops a broad money demand function for Japan, giving a special attention to the role of interna-

tional currency substitution. The empirical results from quarterly time series suggest that a meaningful and

structurally stable broad money demand equation can be estimated for Japan but only if currency substi-

tution is taken into account. Therefore, Japanese monetary authorities should make allowance for the

important effect of currency substitution on domestic money holdings when formulating policies. More-

over, the results further imply that Japan and the US. in particular may need to coordinate their monetary

actions as a “bloc” rather than pursue independent policies.

I. INTRODUCTION

The possibility that currencies of different countries are considered possible substitutes by

domestic asset holders has received considerable attention in recent years. If it exists, this currency substitution (CS) phenomenon has profound economic and policy implications. Significant CS could render domestic money demand very sensitive to developments in for-

eign exchange markets, in turn exposing the domestic economy to shocks from abroad. Sev-

Direct a11 correspondence to: Ali F. Darrat, Department of Economics and Finance, Louisiana Tech University, Ruston, LA 71272; Ahmed Al-Mutawa, Department of Economics, The United Arab Emirates University;

Omar M. Benkato, Department of Finance, Ball State University.

International Review of Economics and Finance, S(3): 321-334 Copyright 0 1996 by JAI Press Inc. ISSN: 1059-0560 All rights of reproduction in anv form reserved.

321

322 ALI F. DARRAT, AHMED AL-MUTAWA, and OMAR M. BENKATO

era1 related consequences would then emerge that require careful analysis. For example, monetary independence across nations whose currencies are sufficiently substitutable may no longer be possible, thus jeopardizing the chief advantage of the current flexible exchange rate regime; namely, that it insulates domestic economies from external shocks (see Miles, 1978, 1981; McKinnon, 1982; Melvin, 1985; Ghosh, 1989). Proponents of the pre- 1973 fixed exchange rate system may then gain some grounds.’ Moreover, conventional money demand functions may also become structurally unstable for they often fail to recognize the

underlying currency substitution process (see Brittan, 1981; Bordo & Choudhri, 1982; Cud- dington, 1983; Ortiz, 1983; Batten & Hafer, 1984; Joines, 1985; Browne, 1986). This latter implication is especially important for the proper conduct of monetary policy which criti- cally hinges on the specification of a stable money demand function.*

Two curious aspects of the CS literature in regards to the money demand stability issue are noteworthy. First, most studies discuss the stability issue rather indirectly on the basis of examining whether some proxy for CS proves statistically significant in money demand regressions. Secondly, these studies generally overlook the CS implications and changes in the yen/dollar exchange value for Japan’s money demand relationships.

The main purpose of this paper is to fill both gaps in the literature. In view of the importance of the Japanese economy and the prominent role of the bilateral yen/dollar exchange value in foreign trade and the international capital markets, we investigate the currency substitution process in the context of the Japanese money demand function. In so doing, we identify a proper money demand model for Japan using recent advances in error-correction and cointegrated systems. We then directly subject the money demand model to a battery of stability tests to investigate the implications of currency substitution for the stability of the Japanese money demand function. The remainder of the paper is organized as follows. Section II briefly discusses the CS phenomenon and describes the proposed money demand model. Section III examines some relevant estimation issues. Section IV reports the empirical results and the stability findings. Finally, section V concludes.

II. THE MODEL

We examine the implications of currency substitution for Japan using the standard asset portfolio-balance theory (see Laidler, 1993). In such a framework, economic agents allocate their wealth among alternative domestic and foreign financial assets.3 For the purpose of this paper, we assume that the holder of wealth distributes it among four components:

W=M+B+xMf+xBf (1)

where W is total wealth, M is domestic money, B is domestic bonds, MS represents foreign monies, Bf represents foreign bonds, and x is the nominal exchange rate (the price of foreign currency in terms of domestic currency).4 Following the portfolio balance theory, the demand for each asset depends, among other things, on the assets’ respective returns.

Since money is a riskless (mostly non-interest bearing) asset, it becomes interesting to analyze how residents allocate their wealth between, say, M and Mf. Often employed is the criterion based on relative expected returns expressed in a common measurement unit (e.g., domestic currency). If the expected real return on domestic money is the negative of the

Currency Substitution and Money Demand Instability 323

expected rate of domestic inflation, then the expected real return on foreign money is the

difference between the expected change in the exchange rate minus the expected rate of

domestic inflation. In light of the above, and based on the returns on various types of assets given in equa-

tion (1) along with the usual substitutability assumption, the demand function for domestic

real money (aggregating over all economic agents) denoted by (M&becomes:

(MIP)d =fi, i, 7f, if, Ax-“)

where y is a scale variable approximated by aggregate income, i is the nominal interest rate

on domestic bonds, rce is expected inflation, if is the nominal interest rate on foreign bonds,

and Axe is the expected change in the exchange rate. The relevant partial derivatives of the

money demand function in equation (2) are:

a(M/P)d

---Jr

a(M/P)d a(M/P)d tl(M/P)d a(M/P)d < o

The second and fourth partial derivatives with regard to the returns on bonds show that

the demand for domestic real balances depends negatively on the nominal returns on both

domestic and foreign bonds. That is, higher returns on bonds (domestic and foreign) tend to

reduce the demand for real money. The third partial derivative indicates that expectations

of higher inflation make holding cash balances more costly, thus leading to a reduction in

the money demand. The fifth partial derivative represents the currency substitution phe-

nomenon in the demand function. If domestic currency is expected to depreciate (i.e., the

exchange rate rises), domestic residents would tend to increase their holdings of foreign

monies relative to domestic currency.5 When currency depreciates, wealth holders can

minimize the cost of holding currencies (M + xMf) by substituting foreign for domestic

money in their portfolio. Thus, while the nominal returns on both monies are exogenously

set to zero, expectations of changes in the exchange rate can make holding one currency

more attractive than the other.

Capital mobility in money demand functions of the type given in equation (2) is taken

into account by the inclusion of the real returns on foreign bonds (of). However, the return

on foreign bonds is also affected by changes in the exchange rate. Therefore, the expected

return on foreign bonds will be given by the return on such bonds plus the expected change

in the exchange rate (if+Ax’).6 The preceding discussion yields the following (linearlized) money demand function:

(M/P)d = a0 + a,y + a,i + cx3n’ + a,(%+ Ax’) + cx5Axe

where

aI > 0 and a2, a3, a,, a5 < 0

In highly integrated global financial markets, intense international arbitrage activities have often worked to equalize interest rates in different currencies. This equilibrium con-

ALI F. DARRAT, AHMED AL-MUTAWA, and OMAR M. BENKATO

dition, known as the Interest Rate Parity (IRP), is perhaps one of the most well-established hypotheses in the international finance literature (see Giddy, 1993; Madura, 1995).’ Although it may seem overly restrictive, and some departures have indeed been docu- mented, numerous empirical studies have provided credence to the theoretical notion of IRP (e.g., Branson, 1979; Frenkel & Levich, 1979; Boothe, 1991).

The IRP implies that:

Substituting from equation (4) into equation (3) yields:

(M/Pjd = PO + p,y + p2i + P3nr + P4Axe

where B, captures the effects of the returns on both domestic and foreign bonds. It should be noted that dropping foreign interest rates from the money demand function can also find support from the analysis in Hamburger (1977). He argues that foreign and domestic interest rates exhibit a high degree of multicollinearity and that domestic money holdings in open economies are primarily sensitive to domestic (not foreign) interest rates. Such observation is further strengthened by the high degree of integration in contemporary international financial markets. It has thus become the practice of some recent empirical studies (e.g., Bahmani-Oskooee; Pourheydarian, 1990) to avoid the multicollinearity problem by including only the domestic interest rate in money demand equations of open

economies. Equation (5) then represents the money demand model in the presence of currency sub-

stitution. The possibility of currency substitution is typically tested by examining whether expected exchange rate change (AX’) proves a significant argument in the demand for

domestic money.

III. SOME ESTIMATION ISSUES

To transform the model into a statistical (estimatable) form, some comments are in order. The first issue pertains to which money measure(s) should be used for the dependent variable.8 In the case of Japan, and for several reasons, the most appropriate definition is perhaps the broad measure (M2 + CDs). This aggregate seems to play a central role in formulating the Japanese monetary policy as could be gleaned from the discussion in Ito (1982), Okina (1985), and Suzuki, Kuroda, and Shirakawa (1988). Indeed, the Bank of Japan began announcing in 1978 official forecasts of the growth in (M2 + CDs), and started conducting open market operations in CDs in 1986. Furthermore, as Corker (1990) argues, this broad money stock fulfills both the transaction needs as well as the asset motive. Con- sequently, this aggregate should be particularly useful for wealth-oriented portfolio models. To obtain real money balances, the money stock (M2 + CDs) is deflated by the GNP deflator (P) to yield the dependent variable.

The second issue concerns the empirical proxies used for the scale variable and the expected returns on the alternative financial assets. Extensively employed in the literature is real GNP (y) for the scale variable.’ The return on domestic bonds is proxied by the inter-


est-rate differential (R), defined as the short-term (call money) rate minus the own interest

rate paid on the interest-bearing components of the broad money stock. Expected inflation is proxied by the change in the rate of inflation (AX). Finally, following the purchasing

power parity, expected changes in the exchange rate (representing currency substitution,

CS) is measured by the differential between inflation rates in Japan and in the United

States, using the wholesale price index to calculate the inflation rates in both countries.” Note, thirdly, that desired domestic money holdings (M@ are not observable. To

replace them with observable values, the underlying adjustment process should be dis-

cussed. A common method used is the Koyck partial adjustment scheme. This scheme

postulates that, due to the forces of adjustment costs, economic agents are only partially

successful in adjusting their actual holdings of real money balances to the desired levels in

each period. It can be shown that this entire partial adjustment process could be represented in the money demand function by the inclusion of a once-lagged dependent variable as an

additional regressor.’ ’ Taking account of the preceding issues, the operational form of the proposed money

demand model is:

where all variables are defined as before, and E is a normally distributed error term assumed

to be serially uncorrelated with constant variance. One of the critical issue in any empirical estimation is determining the appropriate func-

tional form of the equation to be estimated, and this issue represents the fourth and final

technical comment. Popularized by Goldfeld (1973, 1976), many previous money demand studies have employed the log-level form for its convenience since the estimated parame-

ters become direct measures of elasticities, and also for the advantage of the logarithms in

correcting possible heteroscedasticity. Nevertheless, imposing the log-level for all vari-

ables might jeopardize the credibility of the results if some or all of the variables are

nonstationary in that form, leading to the “spurious regression phenomenon” of Granger

and Newbold (1974), and Phillips (1986). The phrase “spurious regression” usually refers

to regressions that are characterized by exaggerated R2 values and meaningless t- and F- statistics with non-standard distributions. To avoid such a problem, recent econometric

studies often employ first-differences because they generally generate stationary economic

time series. However, as Engle and Granger (1987) point out, first-differencing can also fil-

ter out important low frequency (long-run) information when the variables in log-levels are

cointegrated. Put differently, although first-differenced regressions may not suffer from

spurious correlation, they might be plagued with potential misspecification problems of

their own. Consequently, more recent research has focused on consolidating the two alter-

native specifications (log-levels and first-differences) using cointegration analysis and

error-correction models. The first step is to examine the stationarity of each of the economic time series contained

in equation (6). We do this by utilizing the Augmented Dickey-Fuller test of unit roots. If

the test statistics are significant, the null hypothesis that the variable in question contains a unit root (nonstationary) is rejected in favor of the alternative hypothesis that the series is

stationary. Two outcomes are possible. First, some of the variables may be stationary in

log-levels, but others are stationary when differenced once or higher. In this case, the vari-


ables in the model cannot be cointegrated, i.e. do not enjoy a long-run relationship.

Secondly, the variables are all stationary at the same degree of differencing (same order of

integration), and as such they may be cointegrated. In the latter case, another testing step is required to see if the variables share a common

unit root (cointegrated) or possess different unit roots (not cointegrated). We do that by

estimating the cointegrating equation using the nonstationary form of the variables. The

resultant residuals are then checked for stationarity using the ADF test. ‘* If these residuals

prove stationary, the variables are said to be cointegrated, and the final stage in the model

building process involves the specification of a dynamic error-correction equation. This is

done by adding the (once-lagged) residuals obtained from the cointegrating equation as

another regressor into the equation, while the remaining variables are introduced in their

stationary form. The lagged residuals are called the error correction (EC) term which rep-

resents the process by which short-run real money demand adjusts to its long-run

equilibrium position.

IV. EMPIRICAL FINDINGS AND STABILITY RESULTS

We use quarterly data available for Japan over the period 1963:l through 1992:3 (before

adjustments for lags) to estimate the money demand equation. The sources of all time

series are the data tapes of the Citibank and the International Financial Statistics.

We begin by testing the stationarity of our variables by the ADF procedure.13 The test

results suggest rejection of the null hypothesis of nonstationarity for all variables only if

expressed in logarithmic first-differences (denoted by Aln). The relevant test-statistics are

-3.12 for Aln(M/P); -5.36 for Alny ; -4.29 for AlnR; -9.03 for Arc-; and 10.08 for AlnCS.

Compared to the 5% critical value of -3.45, each of these statistics indicates stationarity for

each variable in the logarithmic first-differences. These stationarity findings are then used to formulate cointegrating equations from the

(nonstationary) log-levels of the variables. Following Hall (1986) and Miller (1991), we

select the left-hand-side conditioning variable based on maximizing adjusted-R2, which

found to be the log-level of real broad money. We retrieve the estimated residuals from the

selected cointegration equation and inspect them for stationarity to see whether the nonsta-

tionary variables share a common unit root (residuals are stationary and thus

cointegratedness is indicated) or perhaps they exhibit different unit roots (nonstationary

residuals indicating noncointegrated variables). Results from the ADF test reject the null

hypothesis of nonstationary residuals and thus suggest cointegration for the variables. This

finding implies the presence of long-run equilibrium relationship among the proposed vari-

ables in the money demand model. The estimated residuals from the cointegrating money demand equation represent the

underlying long-run relationship. These residuals are added to the equation to construct a

dynamic (error-correction) model for the Japanese money demand of the form:

Aln(M/P), = b, + blAln(M/P),_ 1 + b,Alny, + b3Alni,

+ b4ArcI + b,Aln CS, + b,EC, _ 1 + pr (7)


where all variables are defined as before, A In denotes the logarithmic first-difference oper-

ator, EC is the error-correction term defined as the (once-lagged) residuals obtained from

the underlying cointegrating equation, and u is a white-noise error term. The prior theoret-

ical predictions for the estimated parameters are:

bl, b2 > 0; b3, b4, b,, be, < 0

Table 1 reports the empirical results from model (7). To enhance statistical efficiency,

we employ the White procedure that yields consistent estimates of the underlying covari-

ante matrix.14 The estimated model seems promising: adjusted R2 is sufficiently high

(F = 17.59),15 and the standard error of the estimated model (=0.012) is less than the stan-

dard deviation of the dependent variable (=0.017). Plots of actual and predicted real broad

money demand, not shown here to conserve space, also suggest a good fit as the proposed

model traces quite well real broad money holdings and their turning points over the estima-

tion period. Further evidence supporting the statistical adequacy of the estimated model is

revealed through the results from several diagnostic tests. The Durbin-Watson statistic sug-

gests no significant serial correlation, a finding confirmed by the score of the Geary-tau

nonparametric statistic and by the insignificant profile of the residuals’ autoregressions.16

The Glejser test of heteroscedasticity also indicates homoscedastic errors with a very insig-

nificant F-value. Finally, Ramsey’s RESET test of misspecification yields an insignificant

F-value, suggesting a proper specification. Equally important, the empirical results from model (7) are also encouraging in that all

proposed variables in the money demand equation appear with correctly signed coeffi-

cients that are statistically significant. With the exception of real income whose coefficient

is significant at the 10% level, all other variables have coefficients that are highly signifi-

cant at the 5% level or better.t7 These empirical results show that real income, representing the scale variable, exerts

some effect on real money holdings in Japan with a relatively small short-term income

elasticity of about 0.118. However, in the presence of an apparently slow adjustment

(=0.091), the long-run income elasticity (=1.30) is not significantly different from unity

(t = 2.73). Of course, a unitary income elasticity of real money holdings is consistent both

with the prediction of the portfolio theoretical model of Tobin (1969), and also with Laid-

Table 1. ECM of Real Broad Money Demand for Japan, 1963:1-1992:3

Variable CoeJZ+vzt Estimates f-Statistics

Constant -0.0002 4.09

Aln[(M2 + CDs)/plr , 0.909 7.40**

Alny, 0.118 1.22* AlnR, -0.0122 -1.93*

A% -0.713 -5.66** AlnCS, -0.124 -2.49**

EC,-, a.970 -5.92**

Nures: The dependent variable is Aln [(MZ + CDs)/P],. The test statistics are discussed in the text. An * indicates statistical significance at the 10% level, while ** indicates significance at the 5% level.

Summary and Test Sratistics: RZ = 0.48, SE. = 0.01226, F = 17.59, D. W. = I .86, Geary-Taw55, Glejser-F = 0.08,

Ramsey RESET-F = -2.13


ler’s (1993) overall assessment of available empirical evidence on the demand functions for broad money stock across many developed countries.

The results further reveal that both measures of the opportunity costs; namely, interest-

rate differential and expected inflation, exert important negative impacts upon the Japanese broad money holdings. The error-correction term also plays a highly significant role in

determining money demand in Japan, and appears with the correct negative sign. There- fore, when money supply exceeds the long-run money demand, and with a stabilizing

monetary policy, money stock growth should fall. The highly significant error-correction term indicates the presence of important feedbacks toward long-run equilibrium in the Jap-

anese monetary dynamics. Without the error-correction term, the Japanese broad money

demand would thus be seriously misspecified. The focus of this paper is of course on the impact of currency substitution in determining

domestic money holdings. The empirical results support the currency substitution phenomenon, and show that the CS proxy plays a highly significant role in the determination of

domestic money demand in Japan. The importance of CS is revealed in at least three different respects. First, the coefficient on the CS variable exhibits the theoretically correct

(negative) sign which is statistically highly significant at better than the 5% level. Sec- ondly, the presence of the CS proxy also seems required for a proper specification of the

Japanese money demand function. Dropping the CS proxy seriously biases money demand function according to the Ramsey RESET test (F-value = 20.08, and the 5% critical

F-value = 2.71). Finally, and perhaps more importantly, the Japanese money demand relationship appears structurally stable only if it incorporates the CS proxy, as we discuss

below in more details. To sum up, the empirical results from the proposed Japanese money demand model seem

quite satisfactory and, in fact, appealing from the theoretical and statistical viewpoints. Nevertheless, usefulness of these results for policy analysis hinges on the structural stabil-

ity of the estimated money demand model. Indeed, as the equation of exchange suggests, the demand for money (velocity) provides the link between monetary policy and the rest of

the economy. Without a stable money demand, policy-makers cannot adequately predict

the impact of a given change in money supply on other macroeconomic variables such as

inflation and unemployment. As Boughton (198 1) points out, it is advisable to subject the estimated money demand

equation to several stability tests, since each testing procedure is designed to address a particular aspect of structural instability. In this paper, we apply three alternative stability techniques; namely, the Chow (1960) test, the Farley-Hinich (1970) test, and the Dufour

(1980) test. The well-known Chow procedure is a test against a sudden (one time) shift in the parameters at a given point in time. We investigate three possible breaking points: the

midpoint to maximize the empirical power of the test;‘* 1973:3 to represent the first oil shock as well as the start of the flexible exchange rate regime; and 1974:4 to account for the start of financial deregulation in Japan (Moreno & Kim, 1993). We also apply the Farley- Hinich procedure which does not require splitting the sample at a given date, but rather applies to the full sample period. As such, the Farley-Hinich test investigates whether the equation has undergone a gradual shift over time. The procedure is based on adding a time- slope term for every regressor. A significant t-ratio for any time-slope indicates structural instability. The Dufour test is a variant of the Chow test which is based on adding dummy variables introduced for each observation over the time interval for which structural insta-


bility is suspected. To allow instability a maximum opportunity to appear, we introduce

shift dummies for the entire quarterly period of 1980s and early 1990s (50 dummies). Sig-

nificant dummy slopes suggest structural instability over the associated time period(s). Table 2 reports the stability results for the estimated money demand model from the three

stability tests. All three tests systematically indicate that the Japanese real broad money

demand equation is structurally stable. For the Chow test, none of the alternative breaking

dates yields significant F-statistics, and none of the time-slope coefficients of the Farley-Hin- ich test proves statistically significant. As to the Dufour test, the vast majority of the Dufour

dummies (47 out of 50) have insignificant slopes. It can thus be blithely stated that the Jap-

anese real broad money demand model estimated in this paper is structurally stable. What is the contribution of including the CS proxy for the stability issue? To address

this, we reestimate the broad money demand equation (7), but without the CS proxy. We

Table 2. Test Results of Structural Instability (Incorporating Currency Substitution)

(A) The Chow Test IB) The Farlev-Hinich Test

Breaking Dates

Mid-point

1973:3

1979:4

F-Stafistic

1.14

1.19

1.29

I i

Time-Slope of

Lagged Dependent Variable

Real Income

Interest Rate Differential

Expected Inflation

Currency Substitution

Error-Correction Term

t-Stutistic

-1.04

0.83

0.79

0.84

a.17

1.38

(C) The Dufour Test

Dummy-Slope for t-Statistic Dummy-Slope fbr

1980:2 0.30 1986:3

1980:3 0.64 1986:4

1980:4 1.36 1987: 1 1981:l 0.85 1987:2

1981:2 1.79 1987:3

1981:3 0.98 1987:4

1981:4 0.18 1988: I 1982: 1 0.14 1988:2

1982:2 0.15 1988:3

1982:3 0.87 1988:4

1982:4 0.10 1989: 1

1983:l 1.48 1989:2

1983:2 0.60 1989:3

1983:3 0.58 1989:4

1983:4 0.79 1990: 1 1984: 1 1.12 1990:2

1984:2 0.53 1990:3

1984:3 0.64 1990:4

1984:4 0.46 1991:l

1985:l 0.25 1991:2

1985:2 0.37 1991:3

1985:3 0.90 1991:4

1985:4 0.84 1992: 1

1986: I 0.49 199212

1986:2 0.72 1992:3

t-Statistic

0.0

0.65

0.87

0.55

0.22

1.59

0.15

0.05

0.57

0.70

0.09

0.97

0.85

1.76

2.00*

2.29*

0.61

0.25

0.29

0.29

0.17

0.42

I .40

1.58

2.59*

ALI F. DARRAT, AHMED AL-MUTAWA, and OMAR M. BENKATO

then explore the stability property of the resultant “standard” money demand equation. Table 3 displays the results from the three stability tests. It is interesting to note that the

Japanese broad money demand model is rendered quite unstable when the currency substi-

tution phenomenon is ignored. Specifically, the Chow test suggests that the equation without the CS proxy is structurally unstable across all three breaking dates. This instability

evidence is corroborated by the Farley-Hinich test. The latter test further suggests that the

adjustment process, in particular, undergoes significant drifts over time [see the highly significant t-statistics on the time-slope coefficients of Aln[(M/P), _ 1 and EC, _ 1]. And in the

Dufour test, most of the Dufour dummies (35 out of 50) are statistically significant in the absence of a variable representing currency substitution. Taken together, all three tests con- firm the important contribution of C’S to the stability property of the Japanese real broad

money demand relationship.”

Table 3. Test Results of Structural Instability (Excluding Currency Substitution)

(A) The Chow Test (B) The Furley-Hinich Test

Breaking Dates F-Statistic Time-Slope of f-Statistic

Mid-point 9.65**

1973:3 12.38**

1979:4 9.94**

Lagged Dependent Variable

Real Income

Interest Rate Differential

Expected Inflation

Currency Substitution

Error-Correction Term

-1.91**

I .60*

2.22**

1.82**

2.70**

(C) The Dufour Test

Dummy-Slope.for t-Statistic Dummy-Slope for t-Statistic

1980:2 1.65

1980:3 2.36*

1980:4 4.43**

1981:l 3.41**

1981:2 9.39**

1981:3 2.71**

1981:4 0.11

1982:l 1.06

1982:2 0.94

1982:3 4.98**

1982:4 0.36

1983: 1 5.82**

1983:2 2.53*

1983:3 4.24**

1983:4 3.44**

1984: 1 7.56**

1984:2 3.27**

1984:3 4.24**

1984:4 2.02*

1985: 1 2.70**

1985:2 1.15

1985:3 5.59**

1985:4 4.02**

1986: 1 2.16*

1986:2 2.61*

1986:3 0.07

1986:4 3.77**

1987: 1 3.95**

1987:2 1.93*

1987:3 1.85

1987:4 s.1s*

1988: 1 1.18

1988:2 0.60

1988:3 2.51*

1988:4 6.64*

1989:l 0.54

1989:2 4.64*

1989:3 2.54*

1989:4 10.08*

199O:l 5.37*

1990:2 6.59*

1990:3 1.93*

1990:4 0.94

1991:l 2.17*

1991:2 0.36

1991:3 1.19

1991:4 1.70

1992: 1 6.46*

1992:2 4.90**

1992:3 7.22%


V. CONCLUSIONS

In this paper we have examined empirically the potential influence of currency substitution on the Japanese broad money demand function. Based on the underlying asset portfolio theory, we develop the money demand function and examine several pertinent estimation issues. In so doing, we focus on the role of currency substitution to the structural stability

of the estimated money demand function. The empirical results systematically suggest that currency substitution represents an

important determinant of broad money demand in Japan. The CS proxy exhibits a correctly signed and highly significant coefficient. Results from specification tests suggest serious misspecification if the Japanese money demand function fails to recognize the influence of currency substitution. A battery of stability tests consistently reveal that the Japanese broad

money demand function is structurally stable only if it allows for the effect of currency substitution. Therefore, currency substitution represents a key channel of external financial interdependence for Japan. Consequently, when formulating policy actions, the Bank of Japan must take into account the significant impact of currency substitution on domestic money holdings. It also seems advisable for Japan and the U.S. to coordinate their monetary actions as a “bloc” rather than pursue independent policies.

ACKNOWLEDGMENTS

We would like to thank, without implicating, an anonymous referee for helpful sugges-

tions. Darrat also wishes to acknowledge with appreciation financial support from the Research Council of the Louisiana Tech University.

NOTES

1. Note that the presence of CS only requires that some (but not necessarily all) residents in a country consider foreign money as a good substitute to domestic money. See Miles (1984).

2. The presence of CS may also reduce the relative efficacy of monetary (versus fiscal) policy (Mundell, 1963), and can further diminish government revenues from seignior- age (Tanzi & Blejer, 1982).

3. For simplicity, wealth is assumed to be allocated only among purely financial assets. Thus, wealth in the forms of real goods (e.g., fixed capital) or shares that represent them are not considered.

4. Typically, the portfolio balance model assumes that individuals do not hold foreign currency (i.e., Mf = 0), which represents the point of departure of the CS model from the standard portfolio balance model.

5. Expectations about exchange rate changes may also indirectly affect the return on foreign bonds. We return to this issue below when we discuss capital mobility.

6. For a terse account of this issue, see Cuddington (1983). 7. Note that ZRP is primarily applicable to short-term securities of one-year or less

since forward contracts are not typically available for longer periods.


8. See Laidler (1993) and the references cited therein for a useful discussion. 9. Data limitation precludes the use of (non-human) wealth to measure the scale vat%

able for Japan. A permanent income measure can also be constructed through an error

learning approach. However, as Laidler (1993) points out, such a measure is econometri-

cally similar to including the current income plus a lagged adjustment as is done here. 10. For an elaborate account of the issues involved and the underlying rationale, see

Ramirez-Rojas (1985). Note that associating differential movements in wholesale price

indices with expected changes in nominal exchange rate does not necessarily assume any

particular causal ordering between the two factors. Moreover, other proxies for CS were

used but produced qualitatively very similar results to those obtained on the basis of the

wholesale inflation differential. A full account of these results are available from the

authors upon request. 11. For a derivation, see Maddala (1992). 12. For example, if all variables are integrated of degree one [denoted - Z(l)], the

residuals should be - Z(0) for the variables to be cointegrated. 13. The testing equations include a time trend (if needed) and a once-lagged dependent

variable. 14. Of course, using single-equation (ULS) procedures may yield simultaneous-equa-

tion bias. However, the alternative procedure of two-stage least squares (instrumental vari-

ables) could also produce seriously biased results due to an improper choice of instruments,

as Cooley and LeRoy (198 1) have argued. In any case, simultaneity bias has not proven a

major problem in estimating money demand equations across different countries. On this,

see Poloz (1980), Bordo and Choudhri (1982) and Laidler (1993). 15. Note that the model is estimated in growth rates and, as such, the value of R2 is

genuine in that it does not suffer from the “spurious correlation” bias. 16. The reliability of the D. W. test is questionable in the presence of a lagged dependent

variable among the regressors. The alternative Durbin-h could not be computed due to a high

value of the relevant standard error. The Geary procedure may be particularly attractive since

it is nonparametric and could thus test for a general (unspecified) form of autocorrelation. 17. Since the underlying theory imposes specific sign restrictions, one-tailed tests are

used. 18. See Farley, Hinich, and McGuire (1975). 19. Like most other testing procedures, these tests can only provide statistical evidence

on the merit of the null hypothesis of parameter stability.

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