asian currency crisis and the generalized ppp: evidence from the far east

ASIAN CURRENCY CRISIS AND THE GENERALIZED PPP 137[Asian Economic Journal 1998, Vol. 12 No. 3] 137[Asian Economic Journal 2005, Vol. 19 No. 2] 137

Asian Currency Crisis and the GeneralizedPPP: Evidence from the Far East

Taufiq ChoudhryUniversity of Bradford

The present paper investigates the effects of the Asian currency crisis of 1997–1998 on the generalized PPP between several real exchange rates of the Far Eastcountries. Monthly log of real exchange rates of the currencies of Thailand,Malaysia, Indonesia, the Philippines and South Korea vis-à-vis the US dollar andthe Japanese yen during 1990–2004 are applied in the investigation. Further testsare conducted between exchange rates vis-à-vis the Thai baht. Tests are conductedfor periods before and after the crisis. Results from the Johansen method ofmultivariate cointegration show a substantial change in the relationship betweenthese real exchange rates before and after the Asian currency crisis. This result isfound using rates based on three currencies: US dollar, yen and baht.

Keywords: real exchange rate, generalized purchasing power parity, Johansencointegration.

JEL classification codes: F40, F41.

I. Introduction

The real exchange rate is a price that represents the competitiveness of a domesticcountry relative to a foreign country. It guides the consumption and resourceallocation decisions across non-tradable and tradable goods, and also revealsnations’ comparative advantages (Lee et al., 2001). Several previous studiesclaim that during the post-World War II period real exchange rate is usuallynonstationary (non-mean reverting) in levels.1 A nonstationary real exchangerate in levels implies that absolute PPP does not hold. The generalized PPP(G-PPP) was developed by Enders and Hurn (1994) to explain the non-meanreverting behavior of the real exchange rates in the post-World War II period.According to Enders and Hurn (1994), bilateral real exchange rates are generallynonstationary because the fundamental macroeconomic variables that determine

1. See Rogoff (1996) and Enders and Hurn (1997) for citations of studies of the stochastic structureof real exchange rate.

ASIAN ECONOMIC JOURNAL 138

real exchange rates, such as real output levels and money supply, are nonstationaryin levels.

The G-PPP hypothesizes that real exchange rates will share common stochastictrends if the fundamental variables are sufficiently interrelated (Enders and Hurn,1994). Therefore, despite the fact that domestic economies differ, the trends inreal forcing variables affecting real exchange rates might be similar acrossdiverse countries (Enders and Hurn, 1997). According to Enders and Hurn (1997),the existence of a long-run equilibrium relationship between real exchange ratesmeans that a shock in any one real exchange rate will affect other bilateralexchange rates. If this is true, then most empirical exchange rate models aremisspecified. If a shock to any real exchange rate will affect the long-run valuesof other real rates, then incorporating the notion of a long-run equilibrium pathshould enhance structural models of exchange rate behavior.

The present paper investigates empirically, by means of the Johansenmultivariate cointegration method, the G-PPP between several Far East countries,before and after the Asian currency crisis of 1997–1998.2 To our knowledge, noother study investigates the affects of the Asian crisis on the G-PPP between theFar East countries. Ogawa and Kawasaki (2003) investigate the G-PPP betweenFar East countries only during the pre-crisis period. Therefore, the present paperinvestigates whether the Asian currency crisis of 1997–1998 had any affect onthe (potential) long-run relationship between the real exchange rates of the FarEast. According to Fujii (2002), as the period after the Asian crisis extends, it isimportant to examine how the economies have been adjusting in the aftermath inorder to better understand the full consequences of the crisis. Some economicrelationships might have been permanently transformed by the crisis, others mightbe relatively unaffected, especially those that rely upon arbitrage relationships.

II. Asian Currency Crisis of 1997–1998

Following years of stellar performance, Thailand, Malaysia, Indonesia, thePhilippines and South Korea experienced a plunge in the external valueof currencies and a sudden reversal of private capital flows from June 1997onward (Pesenti and Tille, 2000). Corsetti et al. (1998) assert that, reflecting themacroeconomic conditions in the East Asian region, national stock markets andcurrencies came under speculative pressures in early 1997. During the spring of1997, the Thai baht started experiencing severe speculative pressure. Thailandhad the weakest economic fundamentals in the region. Once the baht started todepreciate in July 1997, the currencies of Malaysia, Indonesia and the Philippines,with economic fundamentals and export structure similar to those of Thailand,also came under speculative pressure. By the end of October 1997, the baht haddepreciated relative to the US dollar by 55 percent below its January 1997 level;

2. Enders and Hurn (1994, 1997), Liang (1999) and Ogawa and Kawasaki (2003) also investigatethe G-PPP by means of the Johansen method of cointegration.

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similarly, the rupiah (Indonesia) fell by 54 percent, the ringgit (Malaysia) by34 percent, the peso (Philippines) by 33 percent, and the won (South Korea) by14 percent (Corsetti et al., 1998). Initially, South Korea was the least affected. Thiswas mainly because South Korea has the strongest economic fundamentals anddifferent export structure among the five countries under consideration (Corsettiet al., 1998). By the end of September 1997, the combined effective devaluationfor the currencies of the stated five countries had a strong negative impact onthe other countries in the region, such as Taiwan, Hong Kong and Singapore.By early 1998, the appreciation of the US dollar relative to the currencies ofThailand, Malaysia, the Philippines, South Korea and Indonesia reached 78, 52,52, 107 and 151 percent, respectively.

According to Corsetti et al. (1998), two main reasons have emerged for theAsian financial crisis of 1997–1998. First, sudden shifts in market expectationsand confidence were the main causes for the initial financial turmoil, its propaga-tion over time, and regional contagion. Based on Radelet and Sachs (1998),the crisis of 1997–1998 should not be attributed to deterioration in funda-mentals, but rather to panic on the part of domestic and international investors,reinforced to some extent by the failed policy response of the InternationalMonetary Fund and the international financial community. Second, the financialcrisis reflected structural and policy distortions in the countries of the region(Corsetti et al., 1998). The financial crisis was triggered by a fundamental imbal-ance and, once the crisis started, market overreaction and herding caused theplunge of exchange rates, asset prices and economic activity to be more severethan warranted by the initial weak economic conditions.3

Given the dramatic change in the nominal exchange rates of these countriesagainst the US dollar during the crisis, it is of empirical interest to see if there isa change in the G-PPP between the real exchange rates of these countries beforeand after the currency crisis. The Johansen method of multivariate cointegrationtest is applied to check for long-run stationary equilibrium relationships betweenthe real exchange rates. Three sets of tests are conducted. First, real exchangerates of the currencies of Thailand, Malaysia, Indonesia, the Philippines andSouth Korea vis-à-vis the US dollar are applied in the investigation. Further testsare conducted using the real rates of the stated currencies vis-à-vis the Japaneseyen. Both the dollar and the yen-based real rates are applied, because Ogawa andKawasaki (2003) indicate the importance of both currencies in the region. Lastly,real rates vis-à-vis the Thai baht are tested for the G-PPP before and after thecrisis. In this manner, the tests might show whether changing the base currencyof the real exchange rates changes the long-term relationship or not. Also, thethird test might determine whether G-PPP exists between exchange rates withoutthe presence of a major currency. The first two tests involve five real exchangerates and the last test includes four rates.

3. Further analysis of the Asian crisis may be found in Goldstein (1998) and Furman and Stiglitz(1998).


III. Generalized PPP

The G-PPP may be presented in the following form. For m countries in a worldof n countries, there exists a long-run equilibrium relationship between them − 1 bilateral real rates, such that:

R12t = α13R13t + α14R14t + α15R15t + . . . + α1mR1mt + εt, (1)

where R1it is the log of the bilateral real exchange rate in period t betweencountry i and country 1, α1i are the parameters of the cointegrating vectors, andεt is a stationary stochastic disturbance term. From Equation (1), when G-PPPholds, the real exchange rate between countries 1 and 2 can be expressed as aweighted average of the other real rates in the currency area (Liang, 1999).These weights not only reflect trade linkages, but also broader linkages such astechnology transfers, immigration and financial resource movements. Equation (1)becomes the strict (absolute) PPP relationship (between the currencies of coun-tries 1 and 2) if all the α1i are equal to zero.

The G-PPP has been interpreted in terms of an optimum currency area.Krugman and Obstfeld (2000, p. 629) define an optimum currency area as a‘group of regions with economies closely linked by trade in goods and servicesand by factor mobility’. An optimum currency area operates a single commoncurrency or group of currencies, which are irrevocably pegged to each other andhave unrestricted substitutability for all supply side transactions within the zone,while fluctuating together against the rest of the world. According to Mundell(1961), assuming rigid prices and wages in the short run without factor mobility,regions or nations constitute the domain of an optimal currency area if theyexperience the same types of real disturbances. Besides factor mobility, thevolume of intra-regional trade among members is also important (Trivisavavet,2001). Based on the Heckscher–Ohlin model, if two countries are major tradingpartners, some degree of factor price equalization may happen. Therefore, in asetting of more than two countries within a well-defined currency area, thefundamental variables should be sufficiently interrelated, so that the realexchange rates themselves will share a common stochastic trend(s). Hence, withina currency area, there should be at least one linear combination of the variousbilateral real exchange rates (Enders and Hurn, 1994).4

4. Enders and Hurn (1994) fail to find G-PPP between Germany, the UK and the USA, using Japanas the base country. They conclude that these three countries are not within a currency area. They areable to find G-PPP between these three countries and a few smaller Asian countries using Japan asthe base country. Enders and Hurn (1997) further present evidence of G-PPP between the Group ofSeven countries and Group of Seven countries based on the US dollar. Liang (1999) fails to findevidence of G-PPP between Hong Kong and China, but does find G-PPP when Japan is added to thegroup. All real rates are based on the US dollar. Ogawa and Kawasaki (2003) are able to findevidence of G-PPP between different combinations of real exchange rates from the Far East. Allstudies apply the Johansen method of cointegration.


Of the five countries under study, four (Indonesia, Malaysia, the Philippinesand Thailand) belong to the ASEAN Bloc.5 The 1967 treaty of ASEAN statesthat the group would promote economic, social and cultural development of theregion through cooperation, and that the organization would be a forum forresolving economic and political differences. There has been unprecedentedregional growth in South-East Asia in the past 3 decades. Some studies claimthat ASEAN was a major factor in regional growth. This region of 500 millionpeople became one of the most dynamic in the world up to the Asian currencycrisis of 1997–1998. Potentially, ASEAN could provide a set-up for an optimumcurrency area among its members.

Bayoumi and Eichengreen (1994), based on the framework of Mundell (1961),conclude that Asia can be grouped into optimum currency areas: the NorthernAsian bloc (Japan, Korea and Taiwan) and the South-East Asian bloc (Hong Kong,Indonesia, Malaysia, Singapore and Thailand).6 Bayoumi et al. (2000) indicatea possible optimal currency area among the ASEAN countries of Malaysia,Indonesia, Singapore and Thailand. Trivisvavet (2001) also shows that Thailand,Malaysia, Singapore, the Philippines, Japan, South Korea and Hong Kong arepart of an optimum currency area before and after the Asian crisis. Ogawa andKawasaki (2003) show that South Korea, Singapore, Malaysia, Thailand, thePhilippines, Indonesia and China would be able to form a common currency, andthat a common currency basket consisting of both the US dollar and the Japaneseyen would be more applicable as an anchor currency than just the US dollar.7

According to Ogawa and Kawasaki (2003), the dollar peg alone is inadequate forthe Far East countries that have close economic links with not only the USA butalso Japan. A regional currency arrangement in the Far East would require thatregional currencies have stable linkages with each other, and should be stable inrelation to not only a single major currency, but a currency basket. Therefore,Ogawa and Kawasaki (2003) indicate the importance of both the US dollar andthe Japanese yen in a common currency basket for the Far East region. Giventhat both the dollar and the yen are important in the Far East region, the presentpaper applies both the dollar-based and the yen-based real exchange rates.

IV. Data

The data employed are monthly, ranging from January 1990 to September 2004.8

The currencies of Indonesia, Malaysia, the Philippines, South Korea and Thailand

5. The remaining members of the ASEAN Bloc are Brunei, Cambodia, Laos, Myanmar, Singaporeand Vietnam.6. Bayoumi and Eichengreen (1994) apply the Blanchard and Quah structural vector autoregressionapproach in their analysis. The structural vector autoregession method helps identify aggregatesupply and demand disturbances from time series data.7. Ito et al. (1998) also show that a currency basket peg system rather than a dollar peg systemwould have stabilized the trade balances in East Asian countries.8. The start of the period was constrained by the availability of the data.


are investigated. Given that all five countries are in the same region, and evidencealso indicates that these countries are somewhat politically and economicallyconnected and similar, it is of interest to study the G-PPP between them andthe potential effect of the Asian crisis on the G-PPP. Nominal exchange ratesof these countries currencies vis-à-vis the US dollar and the Japanese yenare applied.9 Further tests are conducted between Indonesia, Malaysia, thePhilippines and South Korea vis-à-vis the Thai baht. The data are furtherbroken into pre-crisis period (January 1990 to June 1997) and post-crisis period(July 1998 to September 2004).10 In this manner, change in the relationshipbefore and after the crisis is investigated. Tests are also conducted for the totalsample period.

The real exchange rate (Rt) in the present paper is defined as the nominalexchange rate deflated by a ratio of foreign and domestic price levels. Inlogarithmic form:

Rt = st + p*t − pt, (2)

where Rt is the log of real exchange rate, st is the log of the nominal spotexchange rate defined in local currency units per foreign currency unit, and pt

and p*t are the log of domestic and foreign price level, respectively. In thepresent paper, the USA and Japan are treated as the foreign countries. Whenthe real rates are created vis-à-vis the Thai baht, Thailand is the foreign country.The price index for all countries is presented by the CPI.11 All data are obtainedfrom DataStream. From Equation (2), it is clearly visible that a sudden anddramatic depreciation of the nominal exchange rate (st) as a result of the Asiancurrency crisis will have a dramatic effect on the real exchange rate (Rt). Hence,an observed upward movement of the real exchange rate indicates real deprecia-tion of the domestic currency. Figures 1–3 show the real rates based on the USdollar, the Japanese yen and the Thai baht during the total period, respectively.Most figures indicate a noticeable movement around the crisis period 1997–1998. In the case of the Thai baht-based real exchange rates, the movementaround 1997–1998 is downward, except for Indonesia. The upward movementof the real exchange rate of Indonesia against both the dollar and the yen isshown to be earlier than the movement in the other real rates. Given the suddenmovement in all the real rates around 1997–1998, it is of empirical interest tocheck for a possible change in the potential long-run relationships between thestated real exchange rates.

9. Most of the nominal exchange rates before the crisis were fixed against the dollar and othermajor currencies. After the crisis, some went to independently floating rates for a few years till 2001and then moved to a managed float.10. The actual period of the crisis (July 1997 to June 1998) is avoided because of the turbulentnature of the period, which might possibly bias the results. The length of the three sample periodsensures 177 observations during the total period, 90 observations during the pre-crisis period and87 observations during the post-crisis period.11. Lack of wholesale price index (WPI) for a few Asian countries prevented the use of WPI.

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V. Johansen Cointegration Tests and Results

Two or more nonstationary time series are cointegrated if a linear combinationof these is stationary. Cointegration tests in the present paper are conducted bymeans of the method developed by Johansen (1988) and Johansen and Juselius(1990).12 The Johansen method applies the maximum likelihood procedure todetermine the presence of cointegrating vectors in nonstationary time series.This method detects the number of cointegrating vectors and allows for tests ofhypotheses regarding elements of the cointegrating vector.13 This procedureprovides more robust results than other cointegration methods, especially whenmore than two variables are involved (Gonzalo, 1994). Furthermore, accordingto Gonzalo (1994), this procedure ensures that coefficient estimates are sym-metrically distributed and the median is unbiased, and the hypothesis tests maybe conducted using the standard asymptotic χ2-tests. This conclusion is alsovalid in the case of finite samples. The Johansen maximum likelihood approachsets up the nonstationary time series as the VAR:

Figure 3 Log of real exchange rates (Rt) vis-à-vis the Thai baht

12. This procedure provides more robust results when there are more than two variables (Gonzalo,1994). The Johansen procedure reveals overall the least size distortion (Haug, 1996) and is still morerobust than the other methods even when the errors are nonnormal (Gonzalo, 1994).13. More detailed analysis of the Johansen procedure is provided in Dickey and Rossana (1994)and Harris (1995).


∆Xt = C + i

K

=∑

1

Γi∆Xt−i + ΠXt−i + ηt ηt ~ niid(0,Y), (3)

where Xt is a vector of nonstationary (in levels) variables, ∆ implies first differ-ence, and C is the constant term. The information on the coefficient matrixbetween the levels of the series Π is decomposed as Π = αβ′, where the relevantelements of the α matrix are the adjustment coefficients, and the β matrix con-tains the cointegrating vectors. The constant term is included in order to capturethe trending characteristics of the time series involved.14 The Johansen methodprovides two different tests, the trace test and the maximum eigenvalue test,to determine the number of cointegrating vectors. If a nonzero vector(s) is indic-ated by these tests, a stationary long-run relationship(s) between the relevantvariables is implied. Osterwald-Lenum (1992) provides the appropriate criticalvalues required for these cointegration tests.

A likelihood ratio test and the Akaike Information Criterion are used to selectthe number of lags required in the cointegration test.15 Because there seems to bea linear trend in all the nonstationary series, cointegration tests are conductedwith the inclusion of a deterministic trend.16

Because cointegration tests require a certain stochastic structure of the timeseries involved, the first step in the estimation procedure is to determine if thevariables are stationary or nonstationary in levels. The present paper applies abattery of unit root tests: the KPSS test (Kwiatkowski et al., 1992), the non-linear KSS test (Kapetanios et al., 2003), the structural shift-oriented unit roottest (Perron, 1997), and the fractional unit root test, the GPH test (Geweke et al.,1983).17 All tests confirm all real rates are nonstationary in levels but are station-ary after first difference. This is true during all three periods.18 In other words,all series have one single unit root.19 Results from the unit root tests are notprovided, because they are quite standard and bulky, but they are available onrequest. Given these results, we proceed with the cointegration tests based on theassumption that all variables contain a unit root.

14. As indicated by Harris (1995) and Johansen (1992), the choice of deterministic component inthe model has vital consequences for the asymptotic distribution of the rank test statistics. It is vitalin cointegration tests to determine the rank and the specification of the deterministic component ofthe model.15. According to Gonzalo (1994, p. 220), the cost of over-parametrizing by including more lags issmall in terms of efficiency, but this is not true if it is under-parametrized.16. We checked to determine the components, and results indicated the presence of a deterministictrend. These results are available on request.17. The null hypothesis in the two tests is different. In the KPSS, the null is the absence of a unitroot, while in the GPH test the null is the presence of a unit root.18. Based on the suggestion of a referee, the unit root tests were also conducted with structurebreak(s) because of the crisis. Results show that all real exchange rates are still nonstationary inlevels. These results are available on request.19. This result also indicates lack of absolute PPP between the US dollar and the Asian currenciesduring all three periods. A similar result is also obtained between these currencies and the Japaneseyen and the Thailand baht.


Tables 1–3 present the cointegration results from all three tests during thetotal period (January 1990–September 2004), pre-crisis period (January 1990–June 1997) and post-crisis period (July 1998–September 2004), respectively.20

Most researchers agree that the crisis period lasted from July 1997 to June 1998(Corsetti et al., 1998). Results from all cointegration tests during the total period

Table 1 Cointegration results for the total period (January 1990–September 2004)

Real exchange rates vis-à-vis the US dollar

Vectors Maximum eigenvalue test Trace test

r = 0 31.19 77.37*r ≤ 1 21.96 46.18r ≤ 2 12.14 24.22r ≤ 3 9.14 12.08r ≤ 4 2.94 2.94

Lags = 6, trace correlation = 0.484, autocorrelation test Lagrange multiplier (4), χ2(25) = 8.213,normality test, χ2(10) = 452.413†.

Real exchange rates vis-à-vis the Japanese yen


r = 0 44.81* 77.86*r ≤ 1 16.52 33.05r ≤ 2 8.68 16.54r ≤ 3 6.14 7.85r ≤ 4 1.71 1.71

Lags = 6, trace correlation = 0.428, autocorrelation test Lagrange multiplier (4), χ2(25) = 30.178,normality test, χ2(10) = 272.194†.

Real exchange rates vis-à-vis the Thai baht


r = 0 43.56* 57.43*r ≤ 1 6.55 13.87r ≤ 2 5.87 7.32r ≤ 3 1.46 1.46

Lags = 6, Trace correlation = 0.395, Autocorrelation test LM(4), χ2(16) = 12.212, normality testχ2(8) = 271.696†.Notes: * Indicates significance at the 1% level. † Indicates rejection of the null of normality at the

5% level.

20. All cointegration estimations and the unit root estimations are conducted by means of theRATS software.


Table 2 Cointegration results for the pre-crisis period (January 1990–June 1997)



r = 0 18.93 44.42r ≤ 1 13.19 25.49r ≤ 2 9.76 12.30r ≤ 3 1.29 2.54r ≤ 4 1.25 1.25

Lags = 2, trace correlation = 0.168, autocorrelation test Lagrange multiplier (4), χ2(25) = 29.031,normality test χ2(10) = 84.045†.



r = 0 20.98 48.25r ≤ 1 11.16 27.28r ≤ 2 9.47 16.12r ≤ 3 5.56 6.65r ≤ 4 1.09 1.09




r = 0 19.06 37.37r ≤ 1 9.75 18.32r ≤ 2 7.19 8.57r ≤ 3 1.38 1.38

Lags = 2, trace correlation = 0.144, autocorrelation test Lagrange multiplier (4), χ2(16) = 20.512,normality test χ2(8) = 31.68†.Note: † Indicates rejection of the null of normality at the 5% level.

(Table 1) indicate one significant vector at the 5 percent level or above. In otherwords, results show a long-run stationary equilibrium relationship betweenthe real exchange rates of five currencies vis-à-vis the US dollar during thetotal period. Similar results are found for rates vis-à-vis the Japanese yen, andvis-à-vis the Thai baht. As stated earlier, the cointegration results imply that ashock in one of the real exchange rates will affect other real exchange rates. Thecointegration result might require structural models of these real exchange rates’behavior to incorporate the long-run relationship(s) between the real rates. The


Table 3 Cointegration results for the post-crisis period (July 1998–September 2004)



r = 0 29.49 76.75*r ≤ 1 19.37 47.26r ≤ 2 18.43 27.89r ≤ 3 6.70 9.45r ≤ 4 2.76 2.76




r = 0 44.69* 91.56*r ≤ 1 20.61 46.88r ≤ 2 17.81 26.27r ≤ 3 7.49 8.46r ≤ 4 0.97 0.97




r = 0 29.17** 54.96*r ≤ 1 17.19 25.80r ≤ 2 6.73 8.60r ≤ 3 1.87 1.87

Lags = 4, trace correlation = 0.269, autocorrelation test Lagrange multiplier (4), χ2(16) = 18.287,normality test χ2(8) = 23.315†.Notes: * and ** indicate significance at the 1% and 5% level, respectively. † Indicates rejection of

the null of normality at the 5% level.

cointegration result might also imply the existence of an optimum currency areaamong these Far East countries during the total sample period. This result backsOgawa and Kawasaki’s (2003) view that a common currency basket consistingof both the US dollar and the Japanese yen will be more applicable as an anchorcurrency in the Far East region than just the US dollar. Also, cointegrationbetween all real exchange rates based on the Thai baht implies that the presenceof the US dollar or the Japanese yen is not necessary to find evidence of G-PPPand an optimum currency area between emerging countries. The diagnostic


tests fail to show significant serial correlation. Results indicate the presence ofnon-normal residuals; however, as indicated by Gonzalo (1994), the perform-ance of the Johansen method is still robust, even when the errors are non-normal.21 All the eigenvalues in all three tests are less than unity, implying thatthe system as a whole is stable.

Cointegration test results for the pre-crisis period (January 1990–June 1997)are provided in Table 2. In all three cointegration tests, both the trace test andthe eigenvalue test fail to indicate a significant vector. Therefore, no significantlong-term stationary relationship is found between the real exchange ratesduring the pre-crisis period in all three cases. All eigenvalues are less than unity.All diagnostic statistics are satisfactory. These results also indicate that thesecountries might not have been a part of the optimum currency area beforethe Asian currency crisis. Also a shock(s) in one of the real rates might not beaffecting other real rates, implying that standard structural models of realexchange rate might be adequate for this period.

Table 3 shows the results from the post-crisis period (July 1998–September2004). In all three cases, results indicate one significant vector at the 5 percentlevel or above. During the period after the crisis, there seems to be evidence ofan optimum currency area and transmission of shock(s) from one real exchangerate to others. Once again, all eigenvalues are less than unity. The diagnosticstatistics are, again, satisfactory. In these cointegration tests, the residuals arefound to be normal.

The cointegration results clearly show a considerable change in the long-runrelationship between the real exchanges rates before and after the Asiancurrency crisis of 1997–1998. Results present no evidence of G-PPP during theperiod before the currency crisis, but provide ample evidence of G-PPP after thecrisis period. This is not confined to one set of real rates, but found in all three,US dollar, yen and baht-based, real rates. The existence of a stationary long-runrelationship between the real rates after the crisis allows some interesting obser-vations to be made on the nature of modelling the real exchange rates amongthese countries during this period. The post-crisis results also imply that theseFar East countries might have been operating as an optimal currency area afterthe crisis, but not before the crisis. This result contradicts the result presented byTrivisvavet (2001), and provides support for Ogawa and Kawasaki (2003). Onceagain, the cointegration results show that a common currency basket of thedollar and the yen is more applicable than a single currency by itself.

The lack of G-PPP before the crisis and the presence of G-PPP (and theoptimal currency area) after the crisis might be a result of the higher level ofeconomic policy coordination and links in the exchange rate policies among

21. The autocorrelation is tested by means of the Lagrange multiplier type test for fourth orderautocorrelation (see Godfrey, 1988). The test of normality is based on a multivariate version of theunivariate Shenton–Bowman test (see Shenton and Bowman, 1977).


these countries after the crisis. As stated earlier, if fundamental macroeconomicfactors of different countries are interrelated, real exchange rates will sharecommon stochastic trends, therefore resulting in G-PPP. According to Hernandezand Montiel (2003), the Asian countries that were severely affected by thefinancial crisis made several changes to economic and exchange rate policiesduring the post-crisis period. These policy changes resulted in successfulmacroeconomic performance, stable growth, low inflation etc. during the post-crisis period. The policy changes also resulted in a link between the economicstructure and macroeconomic factors between these countries (Hernandez andMontiel, 2003). Tests using the total period also indicate a significant long-runequilibrium relationship between the rates in all three cases. This result might bedependent on the post-crisis period result.

VI. Normalized Equations and Long-Run Coefficients

The estimated cointegrating vector(s) is given economic meaning by normaliz-ing on the real exchange rate between the Thai baht and the US dollar, andbetween the Thai baht and the Japanese yen. In the third and last cointegrationtest, normalization is done on the real rate between the South Korean won andthe Thai baht. Any real exchange rate could have been applied to create thenormalized equations. The real rate based on baht/dollar and baht/yen are usedmainly because the baht was the first currency to be affected by the crisis. In thelast test, the won/baht is picked randomly. The normalized equations from thetotal period and the post-crisis period are provided in Table 4. No normalizedequations are provided from the pre-crisis period as no significant vectors werefound. The normalized vectors reflect the interrelationships among these realexchange rates. They can be interpreted as long-run elasticities. Using the χ2-test, all variables are tested for significance, as indicated by Johansen and Juselius(1990).

During the total period in the first test involving real exchange rates based onthe US dollar, all variables are significant at the 5 percent level or above. Allcoefficients are larger than unity, implying a large size affect. For example,results show that a 1 percent rise (fall) in the Malaysia/USA real exchange ratewill induce a 1.96 percent fall (rise) in the real exchange rate of Thailand/USA.

In the second test based on the Japanese yen, only the Korean exchange rateis not significant. Comparing the coefficients between the dollar-based rates andthe yen-based rates, the signs on the significant coefficients are the same, exceptfor the case of Korea. Therefore, the analysis of the coefficients in the secondtest (yen-based real rates) is quite similar to the US dollar-based real rates coeffi-cients. The size of the coefficients in the yen-based test is much smaller than theones in the US dollar-based tests. According to Ender and Hurn (1994), if thereal exchange rates are only influenced by real output processes of the variousnations, the normalized vector(s) will be smaller the more similar are a country’saggregate demand parameters. This result is understandable, given that Japan is

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Table 4 Normalized equations and coefficients

Thailand Korea Malaysia Indonesia The Philippines

Total period (January 1990–September 2004)Exchange rates vis-à-vis the US dollar 1.00* −1.302* −1.958* 1.646* 1.582*

(6.00) (10.35) (6.63) (16.61) (12.48)Exchange rates vis-à-vis the Japanese yen 1.00* 0.327 −0.530* 0.319* 0.671*

(19.13) (2.23) (4.96) (16.46) (12.90)Exchanges rates vis-à-vis the Thai baht — 1.00* 1.334* −0.828* −2.068*

— (6.58) (5.71) (23.14) (32.85)Post-crisis period (July 1998–September 2004)

Exchange rates vis-à-vis the US dollar 1.00* 3.612 6.462* 0.929 −9.941*(0.06) (0.68) (9.37) (0.43) (4.55)

Exchange rates vis-à-vis the Japanese yen 1.00* −0.599 0.191 0.444* 0.039(11.99) (1.88) (0.55) (8.61) (0.03)

Exchange rates vis-à-vis the Thai baht — 1.00* 0.566* −0.129* −0.990*— (14.83) (10.56) (8.84) (14.02)

Notes: * Indicates significantly different from zero at the 1% level. χ2 statistics in parentheses. —, no data.


also part of the Far East. The coefficients in the baht-based test are all signi-ficant. In absolute value, the size of the coefficient is relatively large.

Normalized equations from the post-crisis period are also provided in Table 4.Once again, the vectors are normalized using the baht/dollar, baht/yen and won/baht exchange rates. In the US dollar-based tests, only the rates from Malaysiaand the Philippines are significant. The size of the significant coefficients inabsolute value is quite high. Using the Japanese yen-based rates, only the Koreanand Indonesian real rates are significant. The last test between the rates basedon the Thai baht shows that all variables are significant. Because of a lack ofcointegration in the pre-crisis period, it is not possible to provide a comparisonof the coefficients between the two periods (pre- and post-crisis).

VII. Speed of Adjustment and Weak Exogeneity

The Johansen maximum likelihood approach also estimates the speed of adjust-ment to disequilibrium of each variable in the VAR. Each coefficient of rate ofadjustment associated with each nonstationary variable in the VAR indicates thespeed at which the variable in question adjusts towards the single long-runcointegration relationship. In other words, the speed of adjustment coefficientsshow how fast a change in (baht/dollar) real exchange rate or any other variablein the VAR responds to the disequilibrium changes represented by thecointegration vector. Therefore, in the present paper, the speed of adjustmentshows how quickly any deviation from G-PPP tends to correct itself. Accordingto Harris (1995, p. 98), if a certain variable adjustment coefficient is insignific-antly different from zero, then the variable is known to be weakly exogenous.In such a case, the equation for the weakly exogenous variable contains noinformation about the long-run coefficients because the cointegration relation-ship does not enter into this equation and can be excluded from the left-handside of the VAR.

Table 5 presents the speed of adjustment coefficients. These coefficients areonly presented for the total period and the post-crisis period. No speed coeffi-cients are provided for the pre-crisis period as no long-run equilibrium relation-ship is found between the real exchange rates. Therefore, again, no comparisonof the speed of adjustment between the pre-crisis and the post-crisis period canbe provided. The largest coefficient (0.538) is found in the case of real exchangeof the rupiah expressed against the yen during the post-crisis period. This resultimplies that rupiah/yen real exchange adjusts at the rate of 54 percent per monthtoward the long-run equilibrium. The smallest significant coefficient (0.003) isfound for the real rates of ringgit expressed against the dollar during the post-crisis period. Therefore, this exchange rate only adjusts at the speed of0.3 percent per month towards the long-run equilibrium. During the post-crisisperiod the significant speed of adjustment is relatively high. Some of the adjust-ment coefficients are not found to be significant, indicating that few of the realrates are weakly exogenous.


Table 5 Speed of adjustment

Total period

Baht/dollar Won/dollar Ringgit/dollar Rupiah/dollar Peso/dollar

−0.011 −0.032* 0.019** 0.108* 0.030*(−0.743) (−2.628) (2.175) (4.215) (2.306)

baht/yen Won/yen Ringgit/yen Rupiah/yen Peso/yen

−0.007 −0.093** 0.005 0.397* 0.086(−0.158) (−1.908) (0.115) (4.446) (1.494)

Won/baht Ringgit/baht Rupiah/baht Peso/baht

−0.032*** 0.002 −0.196* −0.039**(−1.783) (0.110) (−5.923) (−2.273)

Post-crisis period

Baht/dollar Won/dollar Ringgit/dollar Rupiah/dollar Peso/dollar

−0.012** 0.015** 0.003** 0.010 −0.004(−1.948) (2.027) (2.427) (0.577) (−0.543)

baht/yen Won/yen Ringgit/yen Rupiah/yen Peso/yen

−0.090 −0.021 −0.110*** 0.538* 0.007(−1.297) (−0.331) (−1.588) (3.707) (0.077)

Won/baht Ringgit/baht Rupiah/baht Peso/baht

−0.054 0.045 −0.385* −0.124*(−1.038) (1.010) (−3.546) (−3.024)

Notes: *, **, and *** indicate significantly different from zero at the 1%, 5% and 10% level,respectively. Asymptotic t-statistics in parentheses.

VIII. Conclusion

The Asian currency crisis of 1997–1998 resulted in a massive depreciation ofthe currencies of the Far East countries against the US dollar and other majorcurrencies. The present paper investigates the long-run relationship between thelog of real exchange rates of the currencies of Thailand, Malaysia, Indonesia, thePhilippines and South Korea vis-à-vis the US dollar before and after the cur-rency crisis. In other words, the paper investigates the G-PPP between the statedreal exchange rates before and after the crisis. The G-PPP has been interpreted


in terms of optimum currency area. Also, presence of G-PPP implies that ashock(s) in one real exchange rate might affect other rates and, therefore,might require the incorporation of long-run real exchange rate relationshipsinto the structural model of real exchange rates. The empirical investigation isconducted using monthly data ranging from January 1990 to September 2004,and the Johansen method of multivariate cointegration test. Further tests areconducted using the Japanese yen-based and Thai baht-based real exchangerates. These further tests are conducted to check whether changing the basecurrency changes the results. These further tests also check to see if a commoncurrency basket consisting of both the US dollar and the Japanese yen will bemore applicable as an anchor currency in the region than just the US dollar orthe Japanese yen.

Tests are conducted for three periods: total period (January 1990 to September2004), pre-crisis period (January 1990 to June 1997) and post-crisis period (July1998 to September 2004). In this manner, the change in the relationship betweenreal exchange rates from the pre- to the post-crisis periods is investigated. Theturbulent crisis period is avoided in the test, so as not to influence the results.Results from the pre-crisis period (January 1990 to June 1997) fail to indicatethe presence of G-PPP between these rates. This result is found using all threecurrency-based real exchange rates. Opposite results are obtained from the post-crisis period (July 1998 to August 2004). Using all three currency-based realexchange rates, cointegration between the real exchange rates is confirmed.In each case, one stationary long-run relationship is confirmed. In other words,results provide evidence of G-PPP between the real exchanges of the five FarEast countries, regardless of the base currency, after the Asian crisis. Evidenceof G-PPP during the post-crisis period might be a result of the increased linkin the economic and exchange rate policies among these countries during thepost-crisis period. The cointegration results also imply there might not havebeen an optimum currency area between these countries during the period beforethe Asian crisis, but after the crisis there is evidence of an optimum currencyarea among the countries. This result also indicates that both the US dollar andthe Japanese yen are important in the region and, therefore, a common currencybasket consisting of both the currencies, as an optimum currency, might be moreapplicable than just the US dollar or the Japanese yen. This period’s results alsoimply a transmission of shock(s) between the real exchange rates. This trans-mission might require the inclusion of the long-run relationship(s) betweenthese real rates in the structural model of real exchange rate. Results from thecointegration tests also show a long-run stationary relationship between the realrates during the total period. This result is the same in all three tests.

The speed of adjustment coefficients found to be relatively high in some casesimplies a fast adjustment to the long-run equilibrium. Some of the real exchangerates are found to be weakly exogenous. No adjustment coefficients are providedfor the pre-crisis period because of a lack of any significant cointegrating vectors.Results provided in the present paper advocate further research in this field.


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asian currency crisis and the generalized ppp: evidence from the far east

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