capital market integration

7/28/2019 Capital Market Integration

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Journal of International Money and Finance .18 1999 267287

Capital market integration in the Pacific Basinregion: an impulse response analysis

Kate PhylaktisU

City Uni ersity Business School, Dept of Banking and Finance, Frobisher Crescent,

Barbican Centre, London EC2Y 8HB, UK

Abstract

We examine the extent of capital market integration in a group of Pacific Basin countries

following the deregulation of their markets, and explore whether the financial influence ofJapan in the region has overtaken that of the US. Looking at long-run comovements of realinterest rates through the use of cointegration, and using impulse response analysis toexamine the speed of adjustment of real interest rates to long-run equilibrium following ashock in one of the markets, which is another indicator of the degree of capital marketintegration, we find that these countries are closely linked with world financial markets andmore so with Japan than with US. 1999 Elsevier Science Ltd. All rights reserved.

JEL classifications: F21; F31; F32; F36

Keywords: Capital Market Integration; Impulse Response Analysis; Real Interest Rate

Parity; Pacific-Basin Capital Markets

1. Introduction

The objective of the paper is to examine the extent of capital market integrationin the Pacific Basin countries following the deregulation of their domestic financialmarkets and foreign exchange markets in recent years. This deregulation processhas varied across the countries both in terms of intensity and timing. For example,Hong Kong liberalised interest rates and international capital flows as early as

UTel.: q171 477 8735; fax: 171 477 8881; e-mail: [email protected]

0261-5606r99r$ - see front matter 1999 Elsevier Science Ltd. All rights reserved. .PII: S 0 2 6 1 - 5 6 0 6 9 8 0 0 0 4 6 - 1


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( )K. Phylaktis rJournal of International Money and Finance 18 1999 267287268

1973, while Taiwan and Korea took steps towards liberalisation as late as 1987 and1988, respectively.1

This general deregulation process has been accompanied by signs that Japan hasbeen increasing its financial influence in Asia and possibly overshadowing that of .the US. For example, Yuan 1986 in his examination of capital flows amongst the

.Pacific Basin economies finds that 65% of total net flows private and public in1972 and 55% in 19801982 came from Japan, while the corresponding figures forthe US were 31% and 41%, respectively. More recently, data on securities showthat Japan purchased net US$725 million worth of foreign securities from theNewly Industrialised countries in Asia on a cumulative basis during the periodfrom 1988 to 1991, while the US sold foreign securities on a net basis in thosecountries.2

.Tavlas and Ozeki 1992 examined the role of yen in the region and found thatAsian Central Banks in the course of the 1980s have increased their holdings ofyen from 13.9% of their foreign exchange reserve portfolios to 17.5%. The yen isalso being used more widely to invoice trade and finance in Asia. The percentageof southeast Asian imports denominated in yen increased from 2% in 1983 to19.4% in 1990. The countries that incurred large international debt in the 1970sand early 1980s subsequently shifted the composition away from dollar-de-nominated debt towards yen-denominated debt.3

In this paper, we examine these developments by looking at the effects that theymay have had on the degree of capital market integration between the financial

markets of Pacific Basin countries and the world financial markets, such as the USand Japan. We also examine whether the degree of capital market integration withJapan is greater than that with the US. We look at the covariability of real interestrates, which examines whether prices of financial assets in different countries movein conjunction, but not necessarily have the same level. Different interest-ratelevels may prevail because of different levels of transaction costs and taxes,including the implicit tax of foreign exchange controls and the political riskpremium relating to future controls.4

We also concentrate on a different indicator of capital market integration thancustomarily used in the literature. We look at the speed of adjustment of interestrates to re-establish long-run equilibrium following a shock. The various capitalmarket imperfections which drive a wedge between market rates of return alsoreduce the sensitivity of capital supply that would eliminate differentials beyond

1Other countries in our sample, such as Singapore, deregulated domestic financial markets and foreignexchange markets in 1978, while Malaysia although began liberalising exchange controls in 1973,completed the process in 1984. A free market interest rate regime was adopted in 1978. Japan began thereform of its domestic financial market in mid 1970s, while its foreign exchange market was liberalisedat the end of 1980.2 .See Chinn and Frankel 1994 .3 For example, in Malaysia in 1993 the yen denominated long-term debt was 37.5% compared to 25.1%

.in dollars see The World Bank Debt Tables, 199495 .4 . .See Phylaktis and Wood 1984 and Phylaktis 1988 for an explanation of the effects of capital controlson the international parity conditions.


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( )K. Phylaktis rJournal of International Money and Finance 18 1999 267287 269

the wedge. We examine how long it will take to eliminate a differential which isgreater than that defined by the wedge. The greater the degree of capital mobility

the faster the adjustment to long-run equalisation of real interest rates.Even in the case when capital controls are ineffective, i.e. do not interpose awedge between domestic and foreign real rates of return, the mere existence ofcontrols may give rise to a gradual adjustment to long-run equilibrium as investorsmay be facing increasing marginal costs of acquiring foreign assets or liabilities.This can be seen by considering popular channels through which controls can beevaded and capital can be transferred abroad, such as physical movements of cashand leads and lags. Physical movements of cash might involve increasing costs if thepotential fine is a proportion of the attempted transfer that is increasing with thesize of the transfer. Leads and lags could also involve increasing marginal costs

because importers and exporters can operate only with a limited flow their regular.trade transactions , and lengthening the delay with which they pay their foreign

bills becomes increasingly more difficult to justify to the authorities enforcing thecontrols.5

This indicator of capital market integration provides useful information to policymakers. The existence of a wedge between interest rates gives a country theopportunity to persue a stance of monetary policy which entails a different interest

.rate from the world interest rate under fixed exchange rates . If this stance is,however, changed entailing for example a bigger differential than the existing

.wedge the tightness of capital controls remaining the same , the length of time it

takes for the initial wedge to be re-established, indicates the breathing space thatthe monetary authorities have.

Many studies on capital market integration in the Pacific Basin region haveconcentrated on integration between Japan and the US see e.g. Otani and Tiwari,

.1981; Ito, 1988; and Bosner-Neal and Roley, 1994 . Recently, there has been a lotof interest in other Pacific Basin countries. For example, Bhoocha-Oom and

. Stansell 1990 look at interest rates adjusted and unadjusted for exchange rates. .changes between Hong Kong and Singapore vs. the US. Faruqee 1992 examines

the uncovered interest rate differential between Singapore, Korea and Thailand vs.the Japanese LIBOR taken to represent the world rate of interest.6 Dooley and

.Mathieson 1994 look at seven Pacific Basin countries vs. the US using ananalytical framework for interest rate determination, where the prevailing interest

rate represents a weighted average of open US interest rate adjusted for the.change in the exchange rate and closed economy rates that would have existed

7 .otherwise. Reisen and Yeches 1993 using the same framework examine Koreaand Taiwan in greater detail. The results of these studies support the view thatthere is substantial integration between domestic and international financial mar-

5 .Similar assumptions of increasing marginal costs are also made in Khan and Haque 1985 and Gros .1987 .6 The change in the exchange rate is assumed to be zero.7 .This is based on work done by Edwards and Khan 1985 for the case of Singapore and Colombia; and

.Haque and Montiel 1991 for 15 developing countries.


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kets in Japan, Hong Kong, Singapore and Malaysia, while the views are divided forKorea and Thailand. In Taiwan capital market integration with world financial

markets has been found to be limited.8

In the current paper, we present evidence which shows that there is substantialcapital market integration even in Korea and especially in Taiwan. That inferenceis based on the examination of real interest rates across countries. That encom-passes both financial integration as well as goods market integration as can be seen

.from the derivation of real interest rate parity RIP below. Assuming that theFisher Hypothesis holds for both countries, i and j in our case, we have

j j j .E r sE i yE , 1t t ,k t t ,k t t ,ki i i .E r sE i yE , 2t t ,k t t ,k t t ,k

where E is the expectations operator conditional on information given at time t,ti is the nominal k period interest rate, r is ex ante real interest rate and ist ,k t ,k t ,kthe inflation rate from time t to t qk. When one subtracts these equations andinvokes rational expectations, since ex ante real interest rates are not observed,one can decompose the resulting relationship as follows:

i i j i .r yr s i y i yfdt ,k t ,k t ,k t ,k t ,tqk

. j i .q fd yE s q E s y q t ,tqk t tqk t tqk t ,k t ,k

< . .q , E s 0, 3tqk t tqk t

where fd is the forward discount on the domestic currency, s is the log spott ,t qkexchange rate and is the inflation prediction error. The first term on the rightt qkhand side is the covered interest differential, the second represents the exchange

.risk premium and the third term the Purchasing Power Parity PPP . Thus, when .one examines RIP, one tests the joint hypothesis of covered interest parity CIP ,

PPP and the stationarity of the exchange risk premium. In the words of Frankel .1993 only CIP is an unalloyed criterion for capital mobility in the sense of thedegree of financial market integration across national boundaries. In our study,

however, forward markets were not well developed in all the countries and for thewhole sample period. We therefore proceeded to test for financial market integra-tion by examining real interest rates. We were encouraged to do so as earlier workhas shown that PPP holds for these Pacific Basin countries over the same sample

.period see e.g. Phylaktis and Kasimmatis, 1994a , thus implying that favourableresults to RIP will be indicative of financial market integration.

The paper contributes to the literature in the following ways. First, we use adifferent methodology to what has been used in the earlier studies on capitalmarket integration in the Pacific Basin countries. We use cointegration method-ology which examines whether real interest rates have a long-run relationship, but

8 .Recently Chinn and Maloney 1996 using a different method of measuring capital mobility based on aportfolio balance model, found evidence of greater degree of openness in Taiwan since early 1989.


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allows for movements in rates within a band defined by transaction costs, taxes andcapital controls. Second, we use the speed of adjustment of interest rates following

a shock in one of the markets as an additional indicator of capital marketintegration, by subjecting the system to an impulse response analysis. Finally, weuse a multivariate framework which takes into account the interactions of realinterest rates between the US and Japan in the examination of capital marketintegration between each one of these world markets and the Pacific Basincountries.

The paper is structured as follows. Section 2 explains methodological issues. Itdiscusses the real interest parity within the cointegration methodology. It explainsthe application of impulse response analysis in a cointegrated vector autoregressive

. .system VAR developed by Lutkepohl and Reimers 1992 and its relation to

capital market integration. It finally examines the application of multivariate .Granger Causality tests suggested by Dolado and Lutkepohl 1996 . Section 3

discusses the data and presents the empirical results. Finally, Section 4 summarisesthe main findings and policy implications.

2. Methodological issues

2.1. Bi ariate and multi ariate cointegration

Real interest rate parity as a way of measuring the degree of integration betweentwo different financial markets has usually taken the form of estimating thefollowing regression

j 1 i .r s q r , 4t 0 t

where rj and ri are the real interest rates in countries j and i, respectively, andt ttesting for the joint hypothesis that s 0 and s 1.90 1

There are two weaknesses with the above test. First, it does not allow for any

capital market imperfections, such as transaction costs. Such costs even when theyare small can lead to estimates of and taking different values from the0 1expected ones of zero and one. Secondly, the usual regression results assume thatindividual real rates are stationary, which is not always the case.10 If the series arenon-stationary then the empirical estimates of the parameters and will be0 1

consistent but their estimated standard errors will not be consistent see Stock,.1987 .

.The use of cointegration technique, developed initially by Granger 1981 to

9Such an approach has been followed in earlier studies on real interest rate linkages applied to the US,Canada and some European countries see Mishkin, 1984a,b; Mark, 1985; Cumby and Mishkin,

.1986;and Merrick and Saunders, 1986 . The results are on the whole unfavourable to real interest rateequalisation.10 .See Mishkin 1995 for evidence on the non-stationary behaviour of real interest rates.


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explore the long-run relationship between two series, overcomes these problems.According to cointegration if two series, such as real interest rate in two different

markets, are non-stationary, but there exists some linear combination of themwhich is a stationary process, then the two rates are said to be cointegrated with a .. 11cointegrating parameter, see Eq. 4 . In the presence of transaction costs1

and non-synchronous trading, will be expected to be different from one. In the1case where costs are not proportional to interest rates, market efficiency impliesthat will be different from zero. Thus, even in the absence of capital controls0the joint hypothesis that s 0 and s 1 can be rejected because of transaction0 1costs and that will not imply that profitable arbitrage opportunities exist.

In a multivariate system the issue of cointegration can be given a different .perspective. Stock and Watson 1988 have developed a test for the existence of

common trends in a set of non-stationary variables. If a set of real rates have asingle common trend, that will mean that any single rate is representative of thegroup of rates examined indicating complete integration of the capital markets. Inother words, although each univariate series might contain a stochastic trend, in avector process these stochastic trends might be common to several of the variables.When the series contain the same stochastic trend then they are said to becointegrated.

. In implementing the Stock and Watson 1988 test as well as the tests for.cointegration in the bivariate case , we use the likelihood ratio test due to

. . ( )Johansen 1988 and Johansen and Juselius 1990 . Let Y ' r , r , r wheret P B C US J r is the real interest rate in the Pacific Basin country, r the real interest rateP B C USin the US and r the real interest rate in Japan; n the number of variables in theJsystem, three in this case. If Y is cointegrated, it can be generated by a vectort

.error correction model VECM :

ky1

.Y s q G Y q G Y q , 5t i tyi k ty1 tis1

where is a 3= 1 vector of drift, Gs are 3= 3 matrices of parameters, and is at3= 1 white noise vector. The Johansen trace test statistic of the null hypothesis

.that there are at most r cointegrating vectors 0 FrFn, and thus n yr commonstochastic trends is

n

.trace s yT In 1 y , 6 . iisrq1

where s are the n yr smallest squared canonical correlations of Y withi t y 1respect to Y corrected for lagged differences and T is the sample size actuallytused for estimation.

In the context of the current paper, if we cannot reject the hypothesis that there

11 .Such an approach has been applied to European countries by Goodwin and Grennes 1994 with morefavourable results to real interest rate parity.


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are two cointegrating vectors, that will imply that there is one common stochastictrend. This will suggest that any single rate will be representative of the rates

included in the group, which implies a considerable degree of integration amongstthe three markets.

2.2. Impulse response analysis

.We next subject our cointegrated vector autoregressive system VAR to impulseresponse analysis in order to examine how rapidly the real interest rate movementsin one market are transmitted to the other markets. As it has been explained thespeed of adjustment is an indicator of the degree of capital market integration.

.Lutkepohl and Reimers 1992 show that innovation accounting can be used to

obtain information concerning the interactions among the variables. They developthe asymptotic distribution of the impulse responses and forecast error variancecomponents of a Gaussian VAR process with cointegrated variables.12 They

.suggest that one estimates the undifferenced VAR of VECM of Eq. 5 .

.Y s qA Y q . . . qA Y q , 7t 1 ty1 p tyk t

where A are 3= 3 coefficient matrices in our case, by multivariate least squaresias the resulting estimator has an asymptotic normal distribution where a covari-ance matrix may be estimated by the usual formula for stationary processes.

.Eq. 7 can be transformed to a vector moving average representation givenbelow which can be used to examine the interactions between the three realinterest rates,

.Y s q . 8t i tyiis0

The coefficients of can be used to generate the effects of , and i P B C US J shocks on the entire time paths of r , r and r . The shocks can take the formP B C US Jof one standard error of each real interest rate. The accumulated effects of the

impulses in , and can be obtained by the appropriate summation of theP B C US Jcoefficients of the impulse response function. Since the VAR is underidentifiedCholeski decomposition is often used to orthogonalise the innovations. The resultsof this approach are not, however, invariant to the ordering of the variables in the

.VAR. In a recent paper Koop et al. 1996 have proposed an alternative approach,the generalised impulse response analysis which does not have that shortcoming.This is achieved by examining the shock in one of the variables, and integrating theeffects of other shocks using an assumed or historically observed distribution of theerrors.

The issue of interest to us is to see how many months it takes for the impulse

12 .The asymptotic theory is derived from Johansen 1988; Johansen, 1991 maximum likelihood estima-tion procedure for such systems.


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responses to decay following a shock. The impulse responses should converge tozero because the system is stationary. In the multivariate system, however, of the

US, the Pacific Basin country and Japan, our interest is not when all threeresponses decay, but when the responses of the US and of the Pacific Basincountry decay on the one hand, and, when the responses of Japan and of thePacific Basin country decay on the other hand. If the speed of convergence of thefirst pair is faster than that of the second, then it can be said that the Pacific Basincountry is better integrated with US than with Japan. In examining the speed ofadjustment in this multivariate framework we take into account the interactionbetween the real interest rates in the US and Japan.

2.3. Multi ariate Granger causality tests

Apart from the examination of the long-run co-movements of real interest rates,we explore the short-run dynamics by performing multivariate Granger-causalitytests for cointegrating systems, by applying the methodology suggested by Dolado

. .and Lutkepohl 1996 . Dolado and Lutkepohl 1996 propose a method which leadsto Wald tests with standard asymptotic 2-distributions and which avoids possiblepretest biases associated with the usual procedure of estimating a first order

.differenced VAR if variables are known to be I 1 with no cointegration, and anerror correction model if they are known to be cointegrated. As they point out the

testing procedure on the estimation of unit roots, cointegration rank and cointe-grating vectors has unknown properties, leaving open the possibility of severedistortions in the inference procedure.

Their method is performed directly on the least squares estimators of thecoefficients of the VAR process specified in levels of the variables.13 The proce-dure is based on the argument that the non-standard asymptotic properties of theWald test on the coefficients of cointegrated VAR systems are due to thesingularity of the asymptotic distribution of the least square estimators. Theirsuggested method gets rid of the singularity by fitting a VAR process whose orderexceeds the true order. They show that this device leads to a non-singulardistribution of the relevant coefficients.

The method involves the following steps. First, one finds the lag structure of the . .VAR by testing a VAR k against a VAR k q 1 , k G 1 using the standard Wald

. .test. Secondly, if the true data generating process is a VAR k , a VAR k q 1 isfitted and standard Wald tests are applied on the first k VAR coefficient matrix.

In the context of our paper, the above method implies fitting the VAR depicted .in Eq. 7 for each Pacific Basin country. The expanded version of the VAR is

13 It should be noted that although the variables are allowed to be potentially cointegrated, it is notassumed that the cointegration structure of the system under investigation is known. Thereforepreliminary unit root tests are not necessary and, the testing procedure is robust to the integration andcointegration properties of the process.


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given below

. . .A L A L A LAr r 11 12 1310P B C P B C ,ty1 P B C . . . .A A L A L A Ls q q , 9r r 20 21 22 23US US ,ty1 US

r r A . . .J J,ty1 JA L A L A L30 31 32 33

where A are the parameters representing intercept terms and A the polynomi-i0 i jals in the lag operator L. We select the lag structure using the Wald test, and thenre-estimate the VAR adding one extra lag. Since each equation has the same laglength, we estimate the three equations using OLS as the estimates are consistent

.and asymptotically efficient e.g. Enders, 1995 . We perform Granger causality tests

to examine whether the past values of the US or Japanese rates have been drivingthe Pacific Basin country rate, by using the Wald tests once again on the initialVAR coefficients. For example, to test whether r Granger causes r , we testUS P B C

.the restriction that A L s 0.12

3. Empirical results

3.1. Data

Six Pacific Basin countries were selected for the empirical analysis: Singapore,Malaysia, Hong Kong, Korea, Taiwan and Japan. The sample period varies foreach country according to the availability of data. For Singapore the sample periodis August 1973September 1993; for Malaysia, January 1982September 1993; forTaiwan and Korea, February 1972September 1993; for Hong Kong, January1976September 1993; and for Japan, January 1974September 1993. We haveused end of month data apart from the case of Korea, where we have used end ofquarter. The money market interest rates used were as follows: 90-day TreasuryBill rate for the US; the three month Gensaki rate for Japan;14 the three monthregulated deposit rate for Hong Kong; and the three month interbank rate forSingapore and Malaysia. For Taiwan and Korea, we have used short-term curbrates. The curb market is an unofficial, largely unregulated financial marketinvolving small borrowers and lenders.15 We have used these rates because thedomestic financial markets were highly regulated even during the 1980s.16,17

14Gensaki transactions consist of the resale or repurchase of bonds at a fixed price after a fixed period.They are short-term capital transactions using bonds as collateral. Prior to February 1977 we have usedthe 60-day Gensaki rate.15 The size of the markets remains substantial but has fallen over the years. For example, in the mid1970s the aggregate size of the curb market in Taiwan was as large as that of all financial institutions

put together. In 1986, according to flow-of-funds accounts for private business enterprises, the ratio of .curb market to total bank borrowing was 48% see Fry, 1990 . The curb rate in Taiwan is the average

loan rate against post-dated checks in Kaoshiung city; and in Korea it is a monthly rate which isprovided quarterly.


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In order to examine the effects of deregulation we divided the sample into twosub-periods. The first period ends in December 1980 and represents the period of

financial regulation. The second sub-period covers January 1981

October 1993 andrepresents the post-liberalisation period. As it has already been mentioned subs-tantial deregulation took place in many of the countries under consideration in thelate 1970s and early 1980s. The exact date of the division of the sample period waschosen because of the drastic shift in foreign exchange control policy that tookplace in Japan. The Foreign Exchange and Foreign Exchange Trade Law waspassed which freed most capital flows. In addition, restrictions on non-residentsGensaki transactions were completely eliminated. This relaxation of foreign ex-change controls makes possible the examination of capital market integrationbetween Japan and the Pacific Basin countries during the 1980s.18

.Following Cumby and Mishkin 1986 , the ex post real interest rate on a j periodfinancial instrument held until maturity was defined from the Fisher Condition as

.r s i y , 10t ,k t ,k t ,k

where r is the real return at time t earned from holding the asset for k periods;t ,ki is the nominal k period interest rate; and is the rate of inflation from t tot ,k t ,kt qk.19

3.2. Bi ariate and multi ariate cointegration results

Before testing for cointegration we tested for unit roots in the real interest ratesfor the two sub-periods, whenever that was possible. The results are not presentedbut can be made available by the author. We used the Augmented Dickey Fuller

.test with and without trend as recommended by Engle and Granger 1987 andfound that the null hypothesis of a unit root for the first difference can be rejectedfor all real interest rates. On the other hand, the null hypothesis of a unit root inlevels was accepted in all cases.20 Thus, like most financial series, these real

.interest rates are I 1 , which means that first differencing is required to achievestationarity.

16 Taiwan completely liberalised interest rates in 1989, while Korea started to lift regulations on interestrates in 1989 and completed the process in 1993.17 Data on prices refer to consumer price index and were taken from the International FinancialStatistics published by the International Monetary Fund, except for Taiwan where prices were takenfrom Monthly Statistics of the Republic of China. Data on interest rates for Malaysia, Hong Kong andSingapore were provided by Nomura Bank; for Korea by the International Monetary Fund which wereobtained from the Korean Development Institute; for Taiwan they were taken from the FinancialStatistics Monthly, Taiwan District, Republic of China; and for the US and Japan from Datastream.18 It should be noted that the US had no foreign exchange controls during the whole sample period, sothat the current exercise on the degree of capital market integration reflects developments in the

Pacific Basin countries.19All returns and inflation are continuously compounded.20 Lags were added in order to induce whiteness of the residuals. In the case of pre-liberalisationsub-period for Singapore and post-liberalisation sub-period for Taiwan the null hypothesis for unit rootcould not be rejected at high lags only.


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We proceeded to test for cointegration. We present first the bivariate cointegra-tion results between the real interest rate of each Pacific Basin country vs. US and

vs. Japan. We use the Johansen trace statistic which has been corrected for small . 21 . .sample bias see Reimers, 1992 . Thus, we use Tynk in Eq. 6 instead of T. .The lag length is chosen by applying the Schwarz information criterion SIC on

. .the undifferenced VAR developed by Schwarz 1978 . Reimers 1992 finds thatthe SIC does well in selecting the lag length.

The results vs. US are shown in Panel A of Table 1. The hypothesis of zero .cointegrating vectors H : rs 0 , against the alternative of one or more cointegrat-0

ing vectors, is easily rejected in every case except in the first sub-period in Japan, atthe 5% level of significance. On the other hand, the hypothesis of at most one

.cointegrating vector H : rF 1 is not rejected.0

The results vs. Japan are reported in Panel B of Table 1. Once again the resultsare favourable to real interest rate parity. When the statistic is corrected for

.sample size, the hypothesis of at most one cointegrating vector H : rF 1 is not0 .rejected, whilst the hypothesis of zero cointegrating vectors H : rs 0 is rejected0

in every case at the 5% level of significance.We have proceeded to the multivariate cointegration tests for real interest rates

in each Pacific Basin country, the US and Japan. Table 2 reports the results ofcalculating the Johansen Maximum likelihood-ratio test statistics to define thedimensionality of the common stochastic trend process. Using a 5% significancelevel, we cannot reject the null hypothesis that there are zero cointegrating vectors

in the first sub-period. This is not surprising as we have found in the bivariate casethat the real interest rates of Japan and the US are not cointegrated.

On the other hand, the results for the second sub-period show the hypothesisthat there are two cointegrating vectors in the full three-dimensional system cannotbe rejected in every case, apart from the group which includes Korea. This impliesthat there is one common stochastic trend and that any single rate is representativeof the three rates, which indicates a considerable degree of integration amongstthese markets. In the case of Korea, the hypothesis that only one cointegratingvector exists cannot be rejected, so that there are two common stochastic trends

and a lower degree of capital market integration.We performed the same exercise for the second sub-period but included all thecountries in one group. Namely, we included the US, Japan, Malaysia, Singapore,Taiwan and Hong Kong. We left Korea out because the frequency of the data isquarterly. The results, also presented in Table 2, show that we cannot reject thehypothesis that five stochastic trends are present, or that there is one singlestochastic common trend, confirming the existence of integration amongst thecapital markets of the Pacific Basin countries, the US and Japan.

In general the results presented so far lend substantial support to real interestrate parity within the framework of cointegration. The results, however, do not

21 The trace test appears to be more robust to non-normality of errors compared to the maximal .eigenvalue see Cheung and Lai 1993 for Monte Carlo results on this issue.


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Table 1Bivariate cointegration tests for real interest rates

Johansen test statistics

H : rs 0 H : rF 10 0

Panel A: comparisons with US

SingaporeAugust 1973December 1980 26.443 7.623January 1981September 1993 48.430 3.830

JapanJanuary 1974December 1980 18.594 5.596January 1981September 1993 20.987 5.022

KoreaJanuary 1972April 1980 22.691 7.094January 1981March 1993 24.811 3.948

MalaysiaFebuary 1982September 1993 30.685 6.835

Hong KongJanuary 1976September 1993 51.839 7.089January 1981September 1993 24.247 4.669

TaiwanJanuary 1972December 1980 21.175 4.885January 1981September 1993 28.270 3.279

Panel B: Comparisons with Japan

SingaporeAugust 1973December 1980 29.747 7.546January 1981September 1993 20.504 5.307

Korea

January 1972

March 1980 20.871 6.726January 1981March 1993 21.716 6.996

MalaysiaFebuary 1982September 1993 27.920 5.709

Hong KongJanuary 1976September 1993 23.418 7.163January 1981September 1993 21.272 5.307

Taiwan

January 1972

December 1980 25.026 7.472January 1981September 1993 27.198 7.171

Notes. If r denotes the number of significant vectors, then the Johansen trace statistics test thehypotheses of at most one and zero cointegrating vectors, respectively. The statistics include a finite

.sample correction see Reimers, 1992 . The 5% critical value for H : rF 1 is 9.243 and H : rs 0 is0 0 .19.964 Osterwald-Lenum, 1992 .


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Table 2Multivariate cointegration tests of real interest rates

Johansen test statistics

H : rs 0 H : rF 1 H : rF 20 1 0

Panel A. Group: Pacific Basin country, US and Japan

First sub-periodSingapore 32.03 9.27 3.55Korea 27.48 14.87 4.17Taiwan 26.57 11.25 3.71

Second sub-perioda aSingapore 53.94 21.96 3.41a

Korea 36.77 15.65 3.43a aMalaysia 41.80 21.39 3.00a aHong Kong 54.39 24.45 6.63a aTaiwan 61.04 25.27 9.13

Group B: US, Japan, Malaysia, Singapore, Hong Kong and Taiwan

95% Quantilea

H : rs 0 182.341 102.1390a

H : rF 1 126.233 76.0690a

H : rF 2 84.945 53.1160a

H : rF 3 49.515 34.9100a

H : rF 4 24.617 19.9640H : rF 5 6.691 9.2430

Notes. For the precise sample periods for Panel A see Table 1. The sample period for Panel B isFebuary 1982September 1993. If r denotes the numbers of significant cointegration vectors, then theJohansen statistics test the hypotheses of at most two, one, and zero cointegrating vectors using the

.trace. The statistics include a finite sample correction see Reimers, 1992 . The 5% critical value for H :0 .rF 2 is 9.24, for H : rF 1 is 19.96 and H : rs 0 is 34.91 Osterwald-Lenum, 1992 .0 0

a Denotes significance at the 5%.

answer the issues relating to the change in the degree of capital market integrationin post-liberalisation times, and whether Japan has come to dominate the region.

3.3. Impulse response analysis

Answers to the above questions are provided by using innovation accounting,and in particular impulse response analysis, to examine how long it takes for thereal interest rate parity to be re-established following a one standard deviationshock in one of the rates. The greater the speed of adjustment the greater thecapital market integration.

The lag structure of the VAR has been selected using SIC the same criterion.used in the cointegration exercise . It is well known that SIC is a strongly

consistent lag order selection criterion suited for the analysis of finite-lag orderVAR models. Furthermore, various studies have concluded that the SIC performs

.best in small samples see e.g. Lutkepohl, 1991 , which is the case in our study.Before proceeding to the impulse response analysis we examined the non-diagonal


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error variance matrix for each VAR and found, using the log-likelihood ratiostatistic, that in all the cases we could not reject the null hypothesis that the shocks

in different real interest rates are contemporaneously uncorrelated see Appendix. .A . In such a situation, Pesaran and Shin 1998 show that the orthogonalisedimpulse responses and the generalised impulse responses, which are invariant tothe ordering of the variables in the VAR, coincide. Thus, we performed theexercise using Choleski decomposition to orthogonalise the innovations, knowingthat the results will not be sensitive to the ordering of the variables. We explorethe interactions of r , r and r due to a shock originating in the US i.e. anP B C US J

.impulse in in Eq. 8 . We assume that the ordering of variables is r , rUS US P BCand r . Thus, the real interest rate in the US and the Pacific Basin country isJpermitted to have an instantaneous effect on the rate in Japan, but the latter can

only have a lagged impact on the other rates.The results of the first subperiod are given in Table 3. Since we could not reject

the null hypothesis that there are zero cointegrating vectors in the multivariatecointegration tests for that period, the analysis uses a bivariate framework. Thefirst column shows the number of months it takes the real interest rate in eachPacific Basin country and that of the US to converge to their long-run equilibriumfollowing an impulse in the US real interest rate; the second column shows thenumber of months it takes the real interest rate of each Pacific Basin country toconverge with that of Japan following an impulse in the Japanese real interestrate.22 It can be seen that Singapore was equally integrated with the US and Japan,

the interest rates taking 26 months to converge. In contrast, Taiwan and Koreawere both better integrated with the US, the convergence taking 22 and 21 months,respectively. In the case of Taiwan the convergence with the Japanese interestrates was very slow, taking approximately 55 months.

The results of the second sub-period are reported in Table 4. The speed ofadjustment is faster in the second sub-period compared to the first, vs. both the USand Japan, in the case of Singapore and Taiwan. In Taiwan, the improvement in

Table 3

Impulse response analysis for bivariate cointegrating systems Pacific Basin country and the USfollowing an impulse in the US real interest rate; Pacific Basin country and Japan following an impulse

.in the Japanese real interest rate : number of months for the real interest rates to converge

Impulse in the US real Impulse in the Japanese . .interest rates months real interest months

Singapore 26 26Korea 21 24

. .7 quarters 8 quartersTaiwan 22 55

Notes. The analysis refers to the first sub-period, details of which are given in Table 1. The data in thecase of Korea are quarterly and the speed is transferred into months by multiplying by 3.

22 The speed of adjustment refers to the impulse responses decaying up to 0.1.


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Table 4 .Impulse response analysis for cointegrating systems: Pacific-Basin country, US and Japan : number of

. .months for the real interest rate of each Pacific Basin country to converge with i US, and with ii

Japan, following an impulse in the US real interest rate

Singapore Korea Malaysia Hong Kong Taiwan . . . . .months months months months months

US 24 30 19 27 18 .10 quarters

Japan 18 24 .8 quarters 13 21 17

Notes. The order of variables is the real interest rate of US, Pacific Basin country and Japan. Theanalysis refers to the second sub-period, details of which are given in Table 1. The data in the case of

Korea are quarterly and the speed is transferred into months by multiplying by 3.

the speed of adjustment is incredible, especially vs. Japan 17 months reduced from.55 months . Korea seems to be less integrated with the US in the second

subperiod.One feature stands out clearly. All the countries in the sample are better

integrated with Japan than with the US in the second sub-period. This finding isvery interesting, as it confirms the other indicators which show that Japans

influence is dominating the region. Furthermore, Malaysia is the most integratedcountry with Japan, followed by Taiwan and Singapore.

3.4. Multi ariate Granger causality tests

In this last set of tests we look at the short-run dynamics of the system of realinterest rates by performing multivariate Granger-causality tests. Our objective inthis exercise is to find out whether it is the real interest rate of the US or of Japanwhich drives the interest rates in the Pacific Basin countries.

.Applying the methodology suggested by Dolado and Lutkepohl 1996 and

outlined in Section 2.3, we test whether past values of the US real interest ratesGranger cause rates in the Pacific Basin country and vice versa i.e. with reference

. .to Eq. 9 we test the restriction A L s 0 using the Wald test. We also test for12the reverse causality from the Pacific Basin country to the US, i.e. whether

.A L s 0. We repeat the exercise for Granger causality between Japan and the21 .Pacific Basin countries. We test whether A L s 0 for causality running from13

.Japan to the Pacific Basin country; and A L s 0 for reverse causality.31The results between the US and the Pacific Basin country are shown in Panel A

of Table 5 and cover the second sub-period. At the 5% level of significance we find

US real interest rates to Granger cause rates in four of the five countries, theexception being Taiwan. On the other hand, there is no causality from the PacificBasin country to the US. These results show clearly that US real interest ratesdrive the interest rates in Singapore, Korea, Malaysia and Hong Kong.


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Table 5Granger causality tests using a multivariate approach between each Pacific Basin country, US andJapan

. .Market A L s 0 A L s 012 21

Panel A: Wald tests of causality between US and each Pacific Basin countryaSingapore 32.586 12.475

y2 . .0.007= 10 0.131aKorea 18.603 1.267

y1 . .0.004= 10 0.866aMalaysia 17.939 9.746

. .0.003 0.082aHong Kong 25.209 12.356

y1 . .0.007= 10 0.091

Taiwan 9.333 10.101 . .0.407 0.343

. .Market A L s 0 A L s 013 31Panel B: Wald tests of causality between Japan and each Pacific Basin country

aSingapore 16.034 11.219 . .0.041 0.189

aKorea 19.595 6.444y1 . .0.006= 10 0.168

a aMalaysia 20.245 11.592 . .0.001 0.040

a a

Hong Kong 14.829 18.575 . .0.038 0.009aTaiwan 17.517 6.319

. .0.041 0.708

Notes. The tests cover the second sub-period details of which are given Table 1. For an explanation of . athe restrictions see Eq. 5 . Figures in parentheses are P-values. denotes significance at the 5% level.

In Panel B of Table 5, we present the results of the Granger causality testsbetween Japan and each Pacific Basin country. It can be seen that real interestrates in Japan Granger cause interest rates in all the countries including Taiwan.

At the same time, rates in Malaysia and Hong Kong affect Japanese rates.Taking together the results of both Panels, one can make the following points

with regard to the interactions between the real rates in the Pacific Basin region,the US and Japan. First, the results show that both the US and Japanese ratesdrive interest rates in Hong Kong, Singapore, Malaysia and Korea. Secondly, theresults show the importance of Japan in the case of Taiwan, as US rates do notGranger cause rates in Taiwan. Finally, the results show that there are close linksbetween Japan and Malaysia, and between Japan and Hong Kong, as the directionof causality is a two way one in each case.

4. Summary and conclusions

In this paper, we have examined the extent of capital market integration in a


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group of Pacific Basin countries. Our main concern was to see whether there hasbeen an increase in the degree of capital market integration with world markets,

such as the US and Japan, following the deregulation of the regions financialmarkets. Furthermore, whether the degree of capital market integration with Japanhas been greater compared to that of the US, i.e. whether a Yen block has beencreated in the area.

We examined these issues by looking at the equalisation of real interest ratesusing cointegration methodology which tests for a long-run relationship betweeninterest rates within a band defined by transaction costs, taxes and capital controls.We have examined another indicator of integration which relates to the speed ofadjustment of real rates to re-establish long-run equilibrium following a shock inone of the rates using innovation accounting analysis. Our final exercise concen-

trated on exploring the short-run dynamics through multivariate Granger-causalitytests. In general, the use of multivariate approach in this paper is superior to thebivariate framework used in other studies, as the process takes into account theinteraction of interest rates between the US and Japan in examining capital marketintegration between each one of these world markets and the Pacific Basincountries.

The following conclusions have been derived from our analysis:First, the results from the cointegration exercise both bivariate and multivari-

.ate provide supportive evidence for real interest rate parity and capital market

integration.Second, the results using the speed of adjustment of real interest rates followinga shock as an indicator of capital market integration show that the degree ofintegration has increased in post-liberalisation period in Singapore and especiallyin Taiwan. Korea, the other country for which data exist for the pre-liberalisationperiod, seems to be less integrated with the US in the second sub-period.

An interesting result which emerges from the analysis is the greater capitalmarket integration with Japan than with the US in the second sub-period. Itconfirms the other indicators which show that Japans influence is dominating theregion. The increased effects of Japan through time could be due not only to the

financial liberalisation in the Pacific Basin countries but in Japan itself. This result,which is based on multivariate approach, is in contrast to the bivariate analysis

.examined in Phylaktis 1997 , which used a different method to estimate the speedof adjustment within the cointegration methodology and showed that capitalmarket integration was greater with the US than with Japan. Thus, taking intoaccount the interaction of interest rates between the US and Japan in theexamination of capital market integration in the Pacific basin countries revealedthe increasing financial influence of Japan.

Finally, the results of the multivariate Granger-causality tests show that both

Japanese and US interest rates have been driving the rates in Singapore, HongKong, Malaysia and Korea, while only Japanese interest rates have been drivingrates in Taiwan. In Malaysia and Hong Kong, however, there is also reversecausation from both of these countries to Japan. Thus, the results of this exercise


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point also to the fact that some of the countries in the Pacific Basin region havedeveloped close links with Japan and in some of them, such as Taiwan, shocks

originating in the US are transmitted via Japan.In this paper, we have established that there are extensive real interest ratelinkages with world financial markets even in countries like Taiwan and to a lesserdegree Korea, where there are still extensive foreign exchange controls. Importantfactors which might have helped investors to move funds across borders are themisinvoicing of exports and imports and leads and lags.23 The open character ofthese economies in terms of exports and imports, together with the low protection-ism in current account transactions, have created many such opportunities for

.evading controls. Mathieson and Rojas-Suarez 1993 have also found in theirstudy, on the experience of industrialised and developing countries with capital

controls, that controls were most effective when they were combined with traderestrictions. Furthermore, the existence of illicit trade in drugs and other goods insome of the Pacific Basin countries has provided additional channels throughwhich funds could be moved across borders.24

The substantial degree of capital market integration in the Pacific Basin Regionfound in the paper, both in terms of the comovement of real interest rates and interms of the speed of adjustment of interest rates following a shock, indicates thatthe effectiveness of monetary policy to affect these economies might be limited inthe long run.25

Acknowledgements

Comments from two anonymous referees are gratefully acknowledged. Theauthor wishes to thank Hashem Pesaran and the participants of the 1997 Econo-metric Society European Meeting in Toulouse, France, for helpful suggestions. Thepaper was written while the author was a visiting consultant at the ResearchDepartment of the International Monetary Fund.

Appendix A:

Examination of the non-diagonal errorariance matrix for each VAR

We test the hypothesis that the shocks in the real interest rates r , r and r ,P B C US Jare not contemporaneously correlated by using the log-likelihood ratio statistic

23 .Kamin 1993 finds under-invoicing and over-invoicing of trade transactions has often been a keyvehicle for large scale capital flows in developing countries with extensive capital controls.24 The existence of a substantial black market for dollars as manifested by the high and variable black

.market premium see Phylaktis and Kasimmatis, 1994b corroborates the development of such channels.25 This is subject to the condition that the interest rates used in the analysis are representative of the

.economy-wide interest rates facing most firms and consumers see Chinn and Dooley, 1995 .


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. .LR H rH s y2 LL yLL0 1 U R

where LL and LL are the maximised values of the log-likelihood functionU R . .under H the unrestricted model and under H the restricted model , respec-1 0

tively. LR is asymptotically distributed as a 2 variate with three degrees offreedom. The 95 and 99% critical values are 7.815 and 11.345, respectively. In the

. .first sub-period where there is a bivariate VAR of i r ,r and ii r ,r , theBP C US P B C J2 variate has two degrees of freedom. The 95 and 99% critical values are 5.991and 9.210, respectively.

First Period:

Bivariate VAR models of r and r ; and r and r .P B C US P B C J

Singapore SingaporeUS Japan

LR 3.442 0.060

Korea KoreaUS Japan

LR 0.368 5.262

Korea TaiwanUS Japan

LR 0.060 9.200

Second Period:

Trivariate VAR models of r ,r and r .P BC US J

Singapore Korea Malaysia Hong Kong TaiwanUS US US US USJapan Japan Japan Japan Japan

LR 7.868 6.365 5.341 2.815 10.419

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