Monetary integration and currency substitution in the EMS: The case for a European monetary...

EUROPEAN

K%?Mc ELSEVIER European Economic Review 41 (1997) 1403-1419

Monetary integration and currency substitution in the EMS: The case for a European monetary

aggregate

Peter Spencer Uniuersiry of London, Birkbeck College, Department of Economics, 7- 15 Gresse Street, London WlP

2LL, UK

Received 15 November 1995; revised 15 October 1996

Abstract

This paper examines the phenomenon of cross-currency substitution and the validity of monetary aggregation within the European Monetary System using non-parametric methods which exploit the relationship between intergroup substitution and separability. The separa-

bility property is tested using the methodology developed by Varian (‘The Nonparametric Approach to Demand Analysis’, Econometrica 50,945-974, 1982; ‘Nonparametric Tests of Consumer Behaviour’, Review of Economic Studies 51, 99-l 10, 1983) and is violated in the case of most European countries, suggesting that the hypothesis of zero currency substitution which forms the basis of national monetary aggregation should be rejected. The results do however validate the use of monetary aggregates for the EMS as a whole, and I use a standard parametric demand for money approach to compare the performance of an EMS divisia index with that of simple sum aggregates during the 1980s and 1990s. 0 1997 Elsevier Science B.V.

JEL classification: C14; C43; Dll; EAI; E52; F41; F33; F36

Keywords: Demand for money; Aggregation: Divisia index; Currency substitution; Separability;

European Monetary System

1. Introduction

Monetary economists have long agonised over the appropriate empirical defini-

tion of ‘money’ at the national level. Now, following Girton and Roper (1981) and

0014.2921/97/$17.00 0 1997 Elsevier Science B.V. All rights reserved.

PU SOO14-2921(96)00065-7

1404 P. Spencer/ European Economic Review 41 (1997) 1403-1419

McKinnon (1982) many argue that an international approach to monetary aggregation is warranted. There is mounting evidence in favour of this approach in Europe. Several reports have documented the growth of cross-border trade and currency holdings. The Deutsche Bundesbank (1995b) estimates that between 30 and 40 percent of Deutschmark notes are held outside Germany. Holdings of other European currencies are particularly significant in the case of Italy (Angeloni et

al., 1991). Reflecting these observations, Artis et al. (1993) provide convincing evidence

of currency substitution between Lira and Deutschmark in the case of Italy. Evidence of substitution by German nationals between Lira, the European Cur- rency Unit (ECU) and Deutschmark is also reported by a recent Deutsche Bundesbank (1995a) econometric study. Artis et al find, following similar studies by Kremers and Lane (1991) and Monticelli and Strauss-Kahn (1991) that the demand for money is more stable at the European than at the national level. These results raise important questions about the level of monetary integration in Europe, suggesting that the level of currency substitution is high enough to require a European rather than a national approach to monetary aggregation.

The aggregation question is approached in this paper using a non-parametric approach based upon microeconomic theory, which says that an economic aggregate will behave like an elemental good, providing that its components are weakly

separable from other goods and services. ’ This separability property can be tested

following the Varian (1982) observation that a necessary condition for this to occur is that the data should be consistent with the Generalised Axiom of Revealed Preference (GARP). Heuristically, this axiom rules out preference reversal. So for example, if budget-constrained choice reveals that quantity basket i is preferred to basket j given prices i, then if a new set of prices allows the agent to afford bundle j as well as a new basket which it has been revealed is not inferior to i, then basket j will not be chosen (Varian, 1982).

Monetary economists have come to regard the GARP test as an important preliminary check when analysing the behaviour of a monetary aggregate. 2 If the components of the aggregate are weakly separable, it is possible to consider stability, controllability and other issues knowing that the aggregate behaves as an elemental good (at least in the historical data). Otherwise, changes in composition

’ These components are weakly separable from the rest if the conditional preference ordering over

these components is independent of the particular values taken by the other items (Deaton and

Muellbauer, 1980, sctn. 5.2).

2 There are nevertheless, a number of practical problems with this test, particularly when it is applied

to monetary assets. The maintained hypothesis is based on the assumption that observed quantities are

the result of a process of continuous optimisation which is not subject to structural change, and testing

for measurement error and other forms of misspecification require Monte-Carlo methods which are

time consuming (Patterson, 1991). Non-parametric methods may be more robust in these respects, but

are subject to their own problems, in particular the specification of functional form.

P. Spencer/European Economic Review 41 (1997) 1403-1419 1405

can cause the level of transactions in the economy to move independently of the aggregate.

The GARP test was first used in the monetary area by Swafford and Whitney (1987) and Belongia and Chalfant (1989) using US data. Many other national data sets have been tested for weak separability since, with varying degrees of success. This national approach is appropriate if agents in each country only use the national currency, but Section 2 of this paper shows how the theoretical relationship between intergroup substitution and separability can be used to test this assumption. The results validate the use of monetary aggregates for the European Monetary System (EMS) as a whole, but with the exception of Germany and France do not justify the use of national monetary aggregates. Section 3 shows how nonparametric methods can be used to develop a divisia index of monetary services for the EMS bloc. Section 4 uses a standard parametric demand for money approach to compare the behaviour of this divisia measure with that of simple sum aggregates during the 1980s and 1990s. Section 5 offers a brief

summary of the main conclusions.

2. Nonparametric analysis of the demand for money

We begin by considering the optimisation problem faced by a consumer (or

producer) at the national level. In the case of country j (j = 1,. . . ,n>, we assume a regular utility function:

U’ = lJj( M’, Zj) ,

where Mj = (M’j, A4*j,. . . , M”j), (1)

zj = (Z”, z*‘, . . . , Z’j) )

defined over real consumption (or production) variables Zj and real monetary holdings Mj in that country. If these data are defined at the level of the individual agent and are generated by optimal behaviour then they will satisfy the GARP (Varian, 1983, sctn. 1). At the aggregate level, the optimality and aggregation conditions can be tested by checking whether the aggregate components of Mj and are consistent with the GARP, as in Patterson (1991). However, this requires the Zj variables to be fully specified, and monetary analysts invariably proceed directly to test whether the components of Mj are consistent with the GARP, which is the basic condition for monetary aggregation. 3 In this case, the vector

3 Formally, admissible aggregation requires weak separability and linear homogeneity or homoth-

eticity. The latter assumption ensures that if the components grow at the same rate then so does the

aggregate. However, this requirement can be relaxed using the distance function approach (Deaton and

Muellbauer, 1980; Barnett, 1987). This method yields an index which is linear homogeneous, but base-period dependent. Weak separability is thus necessary and sufficient for this type of aggregation.

1406 P. Spencer/European Economic Review 41 (1997) 1403-1419

Mj can be replaced in Eq. (1) by the scalar aggregate Mj = gj(Mlj M2j Mm’). The researcher can then check the validity of sub-aggregates, such as’n&ow monetary aggregates, using similar test procedures (Varian, 1983).

In the present context, we need to consider a more general problem, faced by a hypothetical European consumer (or trader), moving between all j countries or currencies. In this case the utility function can be defined over real and monetary variables in all countries:

U=lJ(M,Z),

where M= (M’, M2 ,..., M”),

Z=(Z’,Z2, . . . . zn>.

(4

As at the national level, the validity of monetary aggregation checked by testing the restriction: M = g(M”, Mzl,. . , M”‘, . . , Ml”, . . . , Mm”>. However, an additional question arises in this context, one which concerns the degree of cross border interaction. If as is often supposed, agents remain within national bound- aries, then a series of separate national utility functions of the form (1) would rationalise the data as well as a pan-European function of the form (2). Formally, Eq. (2) would be additively (or strongly) separable in the national currency blocs: M = C,g’(M’j,. . . , M”j). 4 This in turn implies weak separability: M =

g(g’(M”, .., Mm’) ,..., g”(M’” ,..., Mm”)). Both of these specifications imply that the national currency groupings each individually satisfy the GARP (Varian, 1983, sctns. 4 and 5). So if we find that the European but not the national data pass the GARP test, this provides evidence of currency substitution.

In order to implement this procedure we need to consider the cost of holding assets denominated in different currencies. First consider the net local currency cost of holding a nominal unit of monetary asset i = 1, . . . , m for one period in country j, financed locally at the benchmark rate Rj. This requires a gross repayment of (1 + Rj) at the end of the holding period, offset by a receipt of (1 + r’j). Discounting the net cost back to the beginning of the period at the benchmark rate gives the nominal unit user cost:

4 The relationship between intergroup substitution and separability is explained in Deaton and

Muellbauer (1980, sctns. 5.2 and 5.3). They show that demands for items in any weakly separable

group can be represented in terms of the total group expenditure and the prices of those items alone. Thus substitution between all items in group g and those in group h depend upon a common

intergroup substitution factor which reflects changes in group expenditures. Strong separability occurs

when this intergroup substitution term is common to all pairs of groups g, h. If this intergroup

substitution term is zero then no intergroup substitution effects occur. In the present context, this is the

case of zero currency substitution.

P. Spencer/ European Economic Review 41 (1997) 1403-1419 1407

Multiplying this by the nominal holding N’j and summing over all assets i gives the total nominal cost of monetary holdings in country j:

cj = C{( Rj - ?‘)/( 1 + Rj)}N’j. (3)

Dividing by Pj, the local consumer price index (CPU, then gives the real cost,

which may be written as the sum of the products of the real holdings (N’j/Pj) and the real rentals (Rj - r’j)/(l + Rj):

C’/Pj = C{( R’- rij)/( 1 + Rj)){N”/P’}. (4)

The nonparametric analysis is conducted using real variables (Swafford and Whitney, 1987). At the national level, Eq. (4) is used as the budget identity and the programme NONPAR developed by Varian (based upon his 1983 paper) is used to see whether the real holdings and real rentals (shown in curly brackets) are consistent with a national utility function (described in Eq. (1)) which is weakly separable in real money holdings. It does this by asking whether these price-quan-

tity data are consistent with the GARP. The basis of this test is to note that under the optimisation assumption, the choice of quantity basket i reveals that it is at preferred to any other basket j affordable with money to spare at prices i. So if a new set of prices allows the agent to afford bundle j as well as a new basket which has been revealed through choice, either directly or indirectly, to be superior to i, then basket j will rtot be chosen (Varian, 1982). NONPAR checks these inequalities and counts the number of violations of the GARP which occur.

In order to examine this property at the European level I aggregate the local nominal costs given by Eq. (3) and then convert them into a common currency using the nominal ECU exchange rate Xj:

c=CxjC[(Rj-rij)/(i +@)]Wj. (5)

Dividing by the appropriate European CPI index P gives the real cost, which may be conveniently written ’ as

C/P= c[ PjXJ/P]c[( R’- #j)/(l +R’)] N”/P’. .i i

This forms the budget identity. Differentiating this with respect to the real holding M’j = N’j/Pj we see that from the perspective of a pan-European trader the real ECU rental K’j = [ PjX’/P][(Rj - r’j)/(l + Rj)] is the product of the real

5 This may be rearranged to show the real cost as the sum Zj[C’/ Pl][ P’X’/ P] of the real national

costs (Eq. (2)) weighted up by real exchange rates [ PjXj/ PI.


exchange rate and the real local rental. This is how I define real holdings and rentals at the European level. 6,7

In view of the difficulty of identifying the components of Z, I follow the standard approach and proceed directly to see whether the European monetary data satisfy the GARP, and are thus consistent with utility (or at least monetary sub-utility) maximisation by a representative European agent. Technically, NON- PAR is used to check whether these monetary data are consistent with a utility function of the form (21, where M = g(g’(M*t, M”, . . . , Mm’), . . . , gYM’“,

I . . , FM”‘)), and so is weakly separable in real money holdings 8 (I will use fi, to denote the derivative of the aggregator function M with respect to M”.) If this is the case, monetary aggregation is valid at the supra-national level. The validity of national monetary aggregation can then be investigated by asking whether the aggregator function is weakly separable in the different currency blocs. As

explained earlier, this provides the basic test for currency substitution. The empirical analysis used data for the principal members of the Exchange

Rate Mechanism (ERM) of the EMS: Germany, the Netherlands, Denmark, France, Italy, Ireland, Belgium and (latterly) Spain and Austria. I thus exclude Britain, but include Austria which has been a member since the beginning of 1995, and operated a shadow ERM policy over most of the period under examination in this paper. The data set is described in Appendix A. I use the Bundesbank’s monetary data for Germany, which gives me three aggregates, but have only one monetary series (Ml) for Belgium. The IMF’s IFS data is employed for other countries, and gives two aggregates for each, money (seasonally adjusted) and

quasi-money (which, like the interest rate data, is non-seasonal in nature). Pooling these data gives a total of 18 monetary balances.

In the non-parametric tests, I allowed for German Monetary Union (GMU) by

splitting the test period in July 1990. 9 The analysis initially focused upon two

6 A referee has pointed out that there is an ambiguity here, since I could alternatively have adjusted

the real holdings rather than user costs by the real exchange rate, thus defining real holdings as

[N”/PJ][ P’Xj/P] and user costs as [C”]. These data can be used to test the validity of a utility

functionoftheseparableform:U=U(~(N”E1/P,...,N”’E’/P,...,N’”E”/P,...,Nm”E”/P),Z).

However, this specification, while formally distinct, gives results which are identical to those reported

below.

’ The assumption of local currency finance eliminates the effect of exchange rate expectations and

implies that changes in monetary balances can be financed by changes in bond holdings or borrowing

in the same currency (Thomas, 1985). This is why, even in the presence of currency substitution, it is

not necessary to use a common benchmark.

a Euler’s law (M = C,,fk,Mk’) together with the optimality conditions (discussed below), allows

this to be interpreted alternatively as the real cost of monetary services. The next section shows that it

can be written as the written as the ratio of divisia weighted nominal money service (N) and price (P) indices.

9 Program size limitations made it difficult to test alternatives like the splicing method used in the next section.


separate sub-periods: first, a pre-GMU period (January 1985-June 1990) and second, a post-GMU period (July 1990-January 1995). I began the analysis at the supra-national level, by checking whether each of these data sets was itself consistent with the GARP. Although this was not the case for the first period, all of the violations occurred during the last fifteen months, perhaps as a result of the turmoil in eastern Europe, so consistency with the Generalised Axiom was achieved by redefining the pre-GMU period as January 1985-March 1989. Surprisingly, no such violations were found in the immediate post-GMU period, although one inconsistent data pair was found a year later (May-June 19911,

which was also dropped from the data. Next, I test for weak separability of subsets of monetary assets over these

periods. Table 1 summarises the results of the basic GARP test. First, I find that the monetary data for a group of six core countries, Germany, Austria, the Netherlands, Denmark, France and Italy are also consistent with the GARP. Together with the ERM-wide result, this immediately validates the use of ERM or core-ERM aggregates defined at the broad money level. Second, non-interest bearing (NIB) assets satisfy the GARP, at least in the post-GMU period, support- ing the use of an Ml money aggregate. The interest bearing (IB) money subgroup failed marginally. Finally, although the national data subsets for France and (with the exception of a minor violation in the second period) Germany satisfy the

GARP, those for other countries do not. As the table shows, NONPAR identifies many violations of the GARP: instances of apparent preference reversal in which a

Table 1

Non-parametric tests

Number of violations of GARP at:

(1) National level 6) Germany

(ii) Austria (iii) Netherlands

(iv) France

(VI Italy

(vi) Denmark (vii) Spain

(viii) Ireland

(ix) Belgium

(a) Pre-EMU

0

35

31

0

4

34 10

49 _

(b) Post-EMU

2

0

0

0

34

I

803

2

(2) Country groups (all assets)

@(vi)

(i)-(ix)

0 0

0 0

(3) Asset groups (all countries)

NIB assets

IB assets

11 0 17 2

1410 P. Spencer/European Economic Review 4if1997) 1403-1419

national currency basket which is chosen in one period in preference to another affordable at the prevailing prices, is then revealed in other periods to be inferior to that alternative.

There are many logical reasons why such violations could occur at the national

level. Individuals may fail to optimise, or their demand functions may not

aggregate properly (German, 1953). Different types of money may be used to finance different types of expenditure, consistent with the results found by Patterson for the UK (1992). Whatever the explanation for these results, they

suggest that national monetary aggregation is no longer valid. Moreover, when seen in context of the positive results for the EMS as a whole, these results indicate that failure is due to substitution between items denominated in different currencies, consistent with the evidence reported in the introduction.

3. A non-parametric index of monetary services

Index number theory can now be used to derive EMS-wide indices which approximate the aggregator function and its nominal money and price components. I will assume for simplicity that the aggregator function is linearly homogeneous

in monetary balances. lo Total differentiation of the aggregator function M then

gives

dlogM=~ijdMij/M=C;f,jMijdlogMiJ DghMgh , ij ij il 1 8

where the denominator in the final expression follows from Euler’s law given linear homogeneity. In equilibrium, the marginal rate of substitution of any two assets must equal the ratio of their prices, or f.,/fgh = Kij/Kgh. This allows the (percentage) change in the aggregator to be written as the change in a divisia monetary service indicator (N) less the change in a similar goods price index (P):

dlog M = CKijMii log Mij

ij /[sKghMgh] =dlogN-dlogP,

where dlog N = Cijs’jdlog NiJ; dlog P = Cijs’jdlog P’j and where the weights can be reduced to the nominal rental expenditure shares:

,ij =_y~[(Rj - rij)/(l +Rj)]Nij/c. (6)

(These weights are invariant to the currency used as numeraire because a change of currency in any period simply rescales all Xj by the same constant, having the same effect on both numerator and denominator.) Approximating these differen-

lo As noted earlier, this assumption can be relaxed using minimum distance methods (Barnett, 1987).


tials by discrete (monthly) changes, we can now decompose changes in the nominal cost (C, described in Eq. (5)) into those of the service index N cumulated from

Alog N=xs’jAlog N’j (7) ij

and those of a dual price index D which is similarly obtained from

A log D =&‘jAlog[Xj(Rj-r’j)/(l +~j)]

=&jAlogXj+Cs”Alog[(H’-r’j)/(l +R’)], (8)

j ij

where Sj = Cisij. This representation shows that the change in the dual depends upon changes in the component exchange rates against the ECU as well as changes in the local currency user costs. However, it is appropriate to define the base currency in terms of a divisia-weighted 9-currency ECU. ” Following this approach, the component changes sum to zero and changes in D only depend upon those in local currency user costs (A log D = Cijsij A log[( Rj - rij)/( 1 + Rj)]). The indices N and D follow from Eqs. (7) and (8) by cumulation.

In line with the results of the nonparametric tests, I include in the divisia

aggregates the 18 monetary components for Germany, Austria, the Netherlands, Denmark, France, Italy, Ireland, Belgium and Spain. At this stage, GMU is allowed for by splicing the West German and pan-German series together in July 1990. These variable weight indices should in principle be used in conjunction with variable weight measures of the factors influencing the demand for money. But in this study, I have used the divisia index in conjunction with the fixed

weight CPI and GDP measures described in the appendix. l2 Dividing the divisia index by this CPI measure gives the real divisia index used in the empirical analysis of the next section.

Under the maintained assumptions, the divisia index gives a second order approximation to the aggregator function and its components. This is a compelling argument for using divisia. Yet many economists (Fisher et al., 1993; Drake and Mullineux, 1995) argue that the assumptions conventionally used to implement the divisia approach are unrealistic. The assumption that portfolios are always and everywhere in equilibrium is clearly at odds with the lags observed in practice, which make it appropriate to smooth the user costs appropriately (Spencer, 1994). The assumption that observed user costs accurately reflect the marginal cost of

” This has a dollar value updated using the rule Alog E = EjSJAlog E’ where E’ is the jth dollar

exchange rate. Then, Alog X’ = Alog E’ - Alog E, which sums to zero over j. ” This was partly in order to avoid the computational burden of constructing additional indices and

partly to facilitate comparison with fixed weight simple sum money measures.


liquidity services should also be modified if banks offer implicit interest (to offset account charges) or offer services which vary with the level of outstanding monetary balances. Bamett and Liu (1995) also show that the user costs should be modified if there is uncertainty and risk aversion. Finally, the choice of the benchmark asset (which should be capital safe, with a maturity matching that of the monetary assets, and yet yield no liquidity service) may also be problematic in many European countries. l3

4. Parametric demand for money analysis

Much work will be required to develop divisia measures which properly allow for these criticisms. In view of these problems, economists who are close to policy (for example Fisher et al., 1993; Herrmann et al., 19941, argue that the choice of aggregate must remain an open empirical question. They assess the behaviour of divisia aggregates using simple sum measures as a benchmark, despite the strong theoretical objections to simple sum measures. l4 The analogue of the simple sum index in the European context is a base-weighted common currency sum index. Consequently I compare the behaviour of divisia with Ml and M3 currency sum measures which are described in the appendix.

Figs. 1 and 2 compare the annual growth in divisia with that in Ml and M3. Historically, these aggregates have behaved in a similar way. I5 This finding

reflects the relatively low level of interest payments on bank deposits which means that simple sum measures give changes in monetary balances a weight which is similar to that implied by the divisia approach. I6

All three series show a general monetary deceleration during the 1980s followed by a marked upsurge in the early 1990s. The main differences lie in the relative sensitivity to output and interest rates. Periods of high interest rates and falling output (1980-1982 and 1992-1993) tend to depress MI by more than divisia, but depress divisia more than M3. Fig. 3 shows the dual user cost index alongside the measures of short and long term interest rates.

The econometric analysis of these data used the programme PC-GIVE. German reintegration was allowed for by splicing price and income series together in

13 This is however possible in the UK system, where the Local Authority deposit market with its

wide maturity range in capital-safe but non-marketable securities offers a suitable benchmark.

I4 Sum aggregation is only valid if the components are perfect substitutes, which they cannot be if

their user costs differ.

r5 Herrmann et al. (1994) obtain a similar result for Germany, finding plausible models for both

divisia and simple sum measures.

I6 Inspection of Eq. (1) shows immediately that if no interest is paid on deposits, then the simple sum and divisia growth rates are the same at the national level. If the benchmark rates in different countries

are equalised, this also true at the European level CEq. (4)).


2-

81 82 83 84 85 86 87 88 89 90 91 92 93 94 95

Fig. 1. Growth in divisia and M 1.

199Oq3. I began with a series of Augmented Dickey-Fuller (ADF) tests which determine the stationarity properties of the data over the estimation period 1980m9-1995ml. In each case I selected the lag length using the Hannan-Quinn information criterion available on this program. The results are shown in Table 2.

All data are in logarithms and seasonally adjusted where appropriate. The ADF

6-

4-

Dmla ’

2- t,

‘=_

0 87 82 a3 84 a5 86 a7 00 89 90 91 92 93 94 95

Fig. 2. Growth in divisia and M3.

1414

80

60

P. Spencer/European Economic Review 41 (19971 1403-1419

Short Term Rate CL H.?cale ----/

I’

. , , , I , / / , , ! / , , , ,(R~‘$Sc~~), / / , ,

a2 84 86 88 90 92 94

14

12

10

8

6

4

Fig. 3. Interest rates and the divisia price dual

statistics show that all three monetary stock series are non-stationary I(1) pro- cesses, as are the aggregate interest rate, dual price and output series.

Given these results, it is important to ask whether there are any cointegrating relationships among these variables. In particular, it is important to know whether the relationship between the divisia index and the macroeconomic variables of

interest to the monetary authorities is in any sense better than that for conventional simple sum aggregates. To approach this question I used the Johansen-Juselius maximum likelihood approach (discussed for example in Cheung and Lai (1993)).

The first stage of this procedure is to estimate an unrestricted Vector Autoregres- sion @AR) explaining the real money stock, output and interest rate or user cost

Table 2

Augmented Dickey-Fuller tests for orders of integration 1980m8-1995ml ’

Levels Differences

Real divisia ADF(S,T)= - 1.31 (-3.34) ADF(3) = - 6.00 (- 2.87) Real M3 ADF(6, T) = - 2.28 (- 3.43) ADF(2) = - 6.76 (- 2.87) Real M 1 ADF(7,T) = - 2.35 (- 3.43) ADF(4) = - 8.41 (- 2.87) Price dual ADF(2) = - 1.64 (- 2.87) ADF(2) = - 5.39 (- 2.87) Long rate ADF(2) = - 1.67 (- 2.87) ADF(l)= -8.52(-2.87) Short rate ADF(2) = - 1.41 (- 2.87) ADF(l)= -9.32(-2.87) Real GDP ADF(3) = - 1.59 (- 2.87) ADF(2) = - 6.36 (- 2.87)

a ADF(X): ADF with X differences and no time trend: ADF(X, T): above with time trend: other

parentheses indicate 95% critical values.


Table 3

Johansen-Juselius estimates 1980m9-1995ml

Divisia M3 Ml

Trace test

Null:

r=O

rll

r12

(95% critical value in parenthesis)

33.41 38.05

(29.7) (34.6)

14.72 16.00

(15.4) (18.2)

0.59 2.45

(3.8) (3.7)

37.81

(34.6)

16.83

(18.2)

3.29

(3.7)

Coefficients of principal eigenvector

Real money - 1.0

Real income 1.208

User cost - 0.244 a

Time trend (-jd

-1.0 - 1.0

0.586 3.474

-0.142 b -0.451 c

-0.017 - 0.027

a Divisia price dual.

b Long rate.

’ Short rate.

d The time trend was not significant in this relationship and was excluded.

variables in terms of their own lagged values. I7 The appropriate lag length is again decided using the Hannan-Quinn criterion. In the case of divisia money, this

suggested a lag length of five, with six lags indicated for broad money and seven for Ml. The Johansen-Juselius algorithm then factorises this VAR model and establishes the number of distinct cointegrating vectors.

The results are reported in Table 3. The trace test results shown in the top panel indicate that there is just one cointegrating vector for each aggregate which is significant at the conventional 95 percent level. The parameters of the cointegrat-

ing vectors are reported in the lower half of the table, normalised to give a unit coefficient on the real service index. Since these variables are logarithms, the coefficients in the divisia relationship indicate an income elasticity of 1.2 and a dual price elasticity of ( -)0.24, both of which seem reasonable on a priori grounds, The precision of these coefficients was checked by restricting them sequentially to zero. In the case of the price dual, this restriction gave a x2(1> test value of 4.27, indicating that the coefficient is well determined and that the demand for monetary services could be controlled through the effect of interest rates on the divisia price dual. It is worth noting that the time trend in the divisia

” Zero-one intercept dummy variables were introduced to check the validity of the 1990 German

data splice and to test for seasonal effects. These were not significant and were dropped from the

reported results.

1416 P. Spencer/ European Economic Review 41 (1997) 1403-1419

equation was not significant, reflecting the relative stability of divisia velocity over

time. The broad money relationship is shown in the second column of the table. This

cointegrating vector relates real balances to real income and the long-term interest rate, a conventional measure of the opportunity cost. ‘*,r9 This model also

incorporates a deterministic time trend, which indicates a gradual but significant upward shift in M3 velocity over time. The final column shows the results of a cointegration analysis of the Ml aggregate. This yielded a single cointegrating vector at the 95% level, relating real Ml balances to real income and the short-term interest rate, representing the opportunity cost of NIB balances. As in the case of broad money, the model also includes a deterministic time trend.

The long-run income elasticities of these Ml and M3 equations suggest that narrow money behaves like a ‘luxury’ good and broad money as a ‘necessity’, which is the reverse of our a priori expectation. This finding appears to be robust, albeit surprising. The precision of the Johansen cointegrating vector was in each case tested using exclusion and other restrictions. For example, restricting the

income elasticities of Ml and M3 to unity gave x*(1> values respectively of 5.33 and 3.09, suggesting that these results are significant. Moreover, a second series of checks using the Granger-Engel OLS regression approach yielded similar cointe-

grating vectors. These results clearly reflect the remarks made in the first part of this section about the relative sensitivity of the aggregates to a recession in output. Although they are by no means conclusive, they do seen to favour the divisia approach to monetary aggregation.

5. Conclusion

Conventional monetary analysis is conducted at the national level on the assumption that there is little scope for substituting different currencies when making transactions.

This implies additive separability of preferences over assets in different national currencies. If aggregate data is consistent with optimisation by a representative agent, then these national currency data sets will each satisfy the GARP, allowing a simple non-parametric test of the zero substitution hypothesis to be satisfied.

18 The initial exploratory research on this relationship also employed the short-run interest rate, and

found two significant cointegrating vectors. However, it was found that these could be restricted by

excluding the short rate from the demand for money (which yields a vector similar to the one reported) and then representing the second relationship as a simple rate spread.

r9 Although the opportunity cost term is smaller than in the case of Divisia money, this coefficient

suggests that long-term interest rates have a powerful influence over the broad monetary aggregate. For

example, the long tern-rate increased from 6.1 to 8.4 percent during 1994, enough to reduce broad

money by 4 percent over the long run.

P. Spencer/European Economic Review 41(1997) 1403-1419 1417

The results of this test, described in the first part of this paper, neatly complement those of recent parametric studies of the demand for money in Europe, indicating that in many smaller countries monetary aggregation is no longer appropriate at the national level. I find that the EMS data as a whole can be rationalised in terms of utility maximisation by a representative agent, validating the use of EMS-wide monetary aggregates. However, with the exception of France and Germany, the underlying utility or aggregator function is not separable in the different currency blocs. This restriction is violated for other countries, usually in both pre- and post-GMU data samples, suggesting that currency substitution is an important factor in Europe.

The second part of the paper uses nonparametric methods to develop a divisia

approximation of the EMS-wide aggregator function and its nominal money and price components using standard assumptions. In view of the questionable nature of some of these assumptions, I assess the behaviour of the EMS divisia aggregate

using simple sum Ml and M3 measures as benchmarks. This analysis suggests that the demand for money is well defined in the ERM, at least if a divisia aggregate is employed. I was able to identify a plausible model of the demand for divisia monetary services. However, the responsiveness of the simple sum aggregates to changes in real incomes seems to be at odds with a priori expectations. These results do lend some support to the theoretical argument for divisia aggregation, but apart from this there appears to be little to distinguish the behaviour of these measures empirically.

The broad similarity between divisia and sum aggregates clearly reflects the relative immunity of the European money markets to financial innovation and in particular the relatively low level of interest paid on bank deposits. However,

Europe’s monetary authorities may not be able to rely on this immunity in the future, especially if a single currency and a unified settlement system make the money markets more competitive. They may soon have to decide between the

alternative approaches to aggregation, largely it seems on a priori grounds. In this case, the strong theoretical foundations of the divisia approach would surely put it in a good position for consideration by the European Central Bank. This position would be strengthened if, in the meantime, research effort were to help resolve the problems of implementation.

Acknowledgements

This research was originally prepared for the CEPR conference, What policy

for the ECB?, held in Frankfurt, June 1995. The present version of the paper has benefited greatly from suggestions made by Franqois Bourguignon and two anonymous referees. I am also grateful for comments by Alec Chrystal, Andy Mullineux and Karl-Heinz Toedter, although this does not of course imply that they agree with all of the views expressed in this paper.


Appendix A. Data sources and definitions

In order to standardise definitions the monetary data were largely taken from the IMF’s International Financial Statistics (IFS) which gives seasonally adjusted figures for Ml (line 34b), and, quasi money, (line 35). *’ The main exceptions were Germany where I used the standard s.a. Bundesbank series and Belgium, where I was only able to obtain a single Ml aggregate (sa., from using national

sources). We also used national banking statistics for M2 (s.a.) where this was available (the Netherlands, France, Denmark, and Italy). In these countries quasi- money was replaced by M2 - Ml, to give a measure of retail IB balances.

In the interests of consistency, the benchmark interest rate was the local IFS government bond rate series in all cases. The local user cost for Ml was assumed to be the benchmark bond rate (line 60d) and for quasi-money (or M2 - Ml), the benchmark rate less the retail time deposit rate (line 60h). None of these series need to be seasonally adjusted.

Data for all countries CPIs, GDP (or GNP) and exchange rates were also taken from IFS. Data for GDP is only available on a quarterly basis and I interpolated this using the monthly industrial production (s.a.) figures, employing the method of Chow and Lin (1987). Apart from the divisia indices I follow the simple sum aggregation procedure described in Artis et al, annex A. This employs nominal exchange rate based weights for the sum Ml and broad M2/M3 aggregates. National GDP/GNP statistics were aggregated using the same weights. Following

Artis et al. (1993) interest rates were ECU-weighted and Consumer Price Indices were aggregated using GDP weights. These weights were based on 1980 data, as reported in Artis et al. (1993, annex A), supplemented with the IMF figures for Austria. These data are available in ASCII text format by sending a 3’/, inch diskette to the author.

References

Angeloni, I., C. Cottarelli and A. Levy, 1991, Cross border deposits and monetary aggregates in the

transition to EMU, Working paper 114 (IMF, Washington, DC).

Artis M.J., R.C. Bladen Hovel1 and W. Zhang, 1993, A European money demand function, In: Masson

and Taylor, eds., Policy issues in the operation of currency union (Cambridge University Press,

Cambridge). Barnett, W.A., 1987, The microeconomic theory of monetary agggregation, In: W.A. Bamett, ed., New

approaches to monetary economics (Cambridge University Press, Cambridge).

Barnett, W.A. and Y. Liu, 1995, The CAPM-extended Divisia monetary aggregate with exact tracking

under risk, Mimeo. (Washington University, St. Louis, MO) Jan.

Belongia, T. and J.A. Chalfant, 1989, The changing empirical definition of money: Some estimates

from a model of the demand for money substitutes, Journal of Political Economy 98, 387-397.

‘” This item is not seasonally adjusted, but is generally regarded as non-seasonal.

P. Spencer/European Economic Rel;iew 41 (1997) 1403-1419 1419

Cheung, Y.W. and KS. Lai, 1993, Finite sample sizes of Johansen’s likelihood ratio tests for

cointegration, Oxford Bulletin of Economics and Statistics 55, 313-328.

Chow, G.C. and A. Lin, 1987, Best linear unbiased interpolation of time series by related series,

Review of Economics and Statistics 53, 372-375.

Deaton, A. and J. Muellbauer, 1980, Economics and consumer behavior (Cambridge University Press,

Cambridge).

Deutsche Bundesbank, 1995a, The demand for money and currency substitution in Europe, Monthly

report (Deutsche Bundesbank, Frankfurt) Jan.

Deutsche Bundesbank, 1995b. The circulation of money abroad, Monthly report (Deutsche Bundes-

bank, Frankfurt) July.

Drake, L. and A. Mullineux, 1995, One Divisia money for Europe?, Discussion paper no. 95-04

(Department of Economics, University of Birmingham, Birmingham).

Fisher, P., S. Hudson and M. Pradhan, 1993, Divisia indices for money: An appraisal of theory and

practice, Working paper no 9 (Bank of England, London)

Girton, L. and D. Roper, 1981, Theory and implications of currency substitution, Journal of Money,

Credit, and Banking 13, 12-30.

Gorman, W.M., 1953, Community preference fields, Econometrica 21, 63-80.

Herrmann, H., H.-E. Reimers and K.-H. Toedter, 1994, Weighted monetary aggregates for Germany,

Mimeo. (Deutsche Bundesbank, Frankfurt).

Kremers, J. and T.D. Lane, 1991, Economic and monetary aggregation and the aggregate demand for

money in the EMS, International Monetary Fund Staff Papers 37, 777-805.

M&&non, R., 1982, Currency substitution and instability in the world Dollar standard, American

Economic Review 72, 320-333.

Monticelli, C. and M. Strauss-Kahn, 1991, European integration and the demand for broad money,

Mimeo. (Committee of Governors of EEC Central Banks).

Patterson, K.D., 1991, A non-parametric analysis of personal sector decisions on consumption, liquid

assets and leisure, Economic Journal 101, Sept., 1103-1116.

Spencer, P.D., 1994, Portfolio disequilibrium: Implications for the Divisia approach to monetary

aggregation, The Manchester School 62, June, 125-150.

Swafford, J.L. and G.A Whitney, 1987, Nonparametric tests of utility maximization and weak

separability for consumption, leisure and money, Review of Economics and Statistics 69, Aug.,

458-464.

Thomas, T., 1985, Portfolio theory and currency substitution, Journal of Money, Credit, and Banking 1,

347-357.

Varian, H.R., 1982, The nonparametric approach to demand analysis, Econometrica 50, 945-974.

Varian, H.R., 1983, Non-parametric tests of consumer behaviour, Review of Economic Studies 51,

99-110.

Date post:	31-Oct-2016
Category:	Documents
Upload:	peter-spencer
View:	212 times
Download:	0 times

Monetary integration and currency substitution in the EMS: The case for a European monetary...

Documents