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Exchange rate determination and inflation in Southeast Asian countries
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ELSEVIER
Journal of DevelopmentEconomics Vol. 55 (1998) 421-437
JOURNAL OF Development ECONOMICS
Exchange rate determination and inflation in Southeast Asian countries Joseph D. Alba a, David H. Papell b,, a Department of Economics, Nanyang Technological Universi~, 2263, Singapore b Department of Economics, Universi~ of Houston, Houston, TX 77204, USA
Received 4 March 1994; revised 6 August 1996; accepted 31 December 1996
Abstract We estimate structural open economy models consistent with rational (RE) and theories consistent (TCE) expectations which incorporate cointegration to examine exchange rate determination and inflation in three southeast Asian countries. The RE model is rejected in favor of the TCE model. Our results show that while inflation in all three countries is affected by different external factors, Malaysia and Singapore avoided high inflation despite high levels of economic growth through 'tight' monetary policy. In contrast, the Philippines had high inflation, even with a stagnant economy, due to 'loose' monetary policy and the monetization of government debt. © 1998 Elsevier Science B.V. JEL classification: F41; O12; C32 Keywords: Exchange rates; Inflation;Cointegration;Southeast Asia
1. Introduction The system of floating exchange rates since 1973 has been characterized by high variability and large fluctuations in exchange rates. Much empirical work has been done to explain these variations in developed countries. ~ Since most
* Corresponding author. i These studies include Frankel (1979), Meese and Rogoff (1983) and Papell (1988). 0304-3878/98/$19.00 © 1998 Elsevier Science B.V. All rights reserved. PII S0304- 3878(98)00043-1
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developing countries peg their exchange rates to a major currency or a basket of currencies, very little attention has been given to exchange rate fluctuations in these countries. However, as Warner and Kreinin (1983) argue, even if developing countries peg their currencies to major currencies, their effective exchange rates will still vary so long as major currencies fluctuate against each other. If the effective exchange rates of developing countries vary because of the variation in major currencies, then they would exhibit the same characteristics as major currencies. Variation in effective exchange rates may also have effects on domestic inflation in developing countries. 2 The focus of the few studies that include exchange rate variation in developing countries is on its impact on domestic inflation. Lim (1987), McNelis (1987) and Morrison (1987) have used bilateral exchange rates, which have been occasionally devalued, as one of the determinants of inflation. Rana and Dowling (1985) and Bahmani-Oskooee and Malixi (1992) use a single-equation model with effective exchange rates as one of the variables that explain inflation. Rana and Dowling (1985) find that exchange rate changes and excess money supply do not have a significant effect in nine Asian countries. 3 Bahmani-Oskooee and Malixi (1992) also examine the determinants of inflation in four Asian and nine nonAsian developing countries. 4 They respecify the lag structure in the model 5 used by Rana and Dowling (1985) and show that effective exchange rate depreciation affects domestic inflation in nine of the 13 countries, including the Philippines and Thailand. Furthermore, excess money supply is found to have a positive significant effect on inflation except for the Philippines. This is contrary to the results of Otani (1975), who used an open economy macroeconomic model with an endogenous money supply specification to examine the impact of the money supply on inflation in the Philippines. The lack of robustness of Rana and Dowling's single-equation model may be due to misspecifications that can be corrected only by a structural model. 6 In this paper, we examine effective exchange rate determination and the factors that affect domestic prices in three southeast Asian countries from 1979 to 1995. Malaysia, Philippines and Singapore are chosen because differences in their
2 Edwards (1989) analyzes the behavior of real exchange rates in developing countries. 3 These countries are India, Indonesia, South Korea, Malaysia, Nepal, Philippines, Singapore, Taiwan, and Thailand. 4 The countries included in their study are Brazil, Dominican Republic, Egypt, Greece, India, South Korea, Mexico, Pakistan, Peru, Philippines, Portugal, Thailand, and Turkey. 5 Bahmani-Oskooeeand Malixi (1992) rely on the Almon lag specification instead of the simple lag and Koyck lag structures used by Rana and Dowling (1985). 6 Bahmani-Oskooeeand Malixi (1992) attribute the differences in their's and the results of Rana and Dowling (1985) on the lag structure specification.
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economic policies offer interesting comparisons on the varying results of such policies on the macroeconomy. 7 Singapore is one of the NICs in Asia which has experienced phenomenal growth since the 1970s. High growth is attributed to the shift of the economy to labor-intensive manufacturing from entreprt trade in the late 1960s 8 and the diversification of the economy into more capital-intensive manufacturing, tourism and financial and banking services starting in the late 1970s. 9 Despite Singapore's sustained growth, sound macroeconomic policies have kept inflation within the 0-10% range. Malaysia's growth of 8-10% since the late 1980s mimics Singapore's exceptional growth in the 1960s and early 1970s. The high growth rate coincides with the shift of the economy from primary to manufactured-good exports. As in Singapore, Malaysia maintained a low inflation within the range of 0-10%. The Philippines is a notable exception to the booming economies of southeast Asia. The country experienced both slow growth and high inflation in the 1980s, with the inflation rate reaching a peak of 49% in 1984. The Pacific Economic Development Report (1992-1993) identifies high inflation as one of the reasons the Philippines could not attract more foreign direct investment (FDI) as other countries in the region have done to aid in economic development. The three countries all manage their exchange rates. Balassa and Williamson (1987) describe the exchange rate policy of Singapore as a basket peg within a band. J0 The International Monetary Fund (1986) reports that the exchange rate of Malaysia is a basket peg while Fry (1986) characterizes the exchange rate policy in the Philippines as a managed float. We specify forward-looking price adjustment, following Papell (1997), in order to incorporate cointegration between the exchange rate and its fundamentals. The model is consistent with long-run equilibrium relationships such as purchasing power parity and monetary neutrality, while also tracking short-run movements in effective exchange rates. In contrast with Papell (1997), we employ an endogenous money supply specification and a more general price adjustment equation, which are more appropriate for countries with managed exchange rates, to examine the determinants of inflation. Two versions of the model are estimated. The first version is a constrained rational expectations (RE) model, with the constraints derived from the structural equations and the assumptions necessary to derive a unique rational expectations
7 The inclusion of Thailand and Indonesiaas well as the Newly IndustrializingCountries (NICs) of Hong Kong, South Korea and Taiwan would have made the comparisons even more interesting. However, quarterly real GDP and/or nominal GDP are not available for other east and southeast Asian countries. 8 Noland (1990), p. 33. 9 Wong(1990), p. 58. ]0 On the other hand, Moreno(1988) portrays Singapore'sexchange rate as a managed float.
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solution. The second version is a semi-constrained model, compatible with the concept, proposed by Frydman and Phelps (1990) and developed by Goldberg and Frydman (1993, 1996a, b), of theories consistent expectations (TCE). Although the rational expectations assumption, that economic agents have quantitative knowledge of the model, is clearly debatable, development of serious alternative hypotheses has proceeded slowly. In the TCE, the determinants of exchange rates and prices are qualitatively, but not quantitatively, consistent with the fully constrained rational expectations model. Both versions of the model are estimated for Malaysia, Philippines and Singapore using full system maximum likelihood methods with quarterly data from 1979 (I) to 1995 (II). Although the RE model has mostly significant results, with coefficients within the range of expected signs and magnitudes, the likelihood ratio test rejects the RE model in favor of the TCE model. The TCE model provides insight into understanding effective exchange rate dynamics and the determinants of domestic inflation in southeast Asian countries. While exchange rate expectations are affected mainly by the current exchange rate, effective exchange rate variation significantly affects inflation in all three countries. The other determinants of inflation vary among the different countries: Malaysia, with an average inflation rate of 3.7%, is affected by increases in foreign prices; Singapore, which had an average inflation rate of 2.9%, is affected by lower foreign interest rates; and the Philippines, with an average inflation rate of 15% in the 1980s, is affected by excess foreign demand and higher foreign interest rates. The positive impact of foreign interest rates may be due to the monetization of government debt. The RE model provides information about the three countries' monetary policies. The Philippines is shown to have a 'loose' or partly accommodating monetary policy while monetary policy for Malaysia and Singapore is 'tight' or stabilizing. The paper is organized as follows: Section 2 describes the model and its solution; Section 3 discusses the estimation results; Section 4 considers the policy implications; and the conclusions are in Section 5.
2. The model
We utilize a small open economy model, the demand side of which is based on Dornbusch (1976) and follows Papell (1997). Asset market equilibrium equates money supply and money demand, with money demand a function of domestic income and the domestic interest rate: mt -Pt
= Y, - a l l i t + ~ l t ,
(1)
where m t is the log of the domestic money supply, Pt is the log of the price level, is the log of domestic real income and i t is the domestic nominal interest rate.
Yt
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The disturbance terms (E's) are random variables which may be serially correlated. Equilibrium in the goods market is expressed as: Y, = a21( et + Pt * -- P,) + a22 Y, * - a23r, + e2,,
(2)
where e, is the log of the nominal effective exchange rate, j f Pt * is the log of foreign prices, Yt * is the log of foreign output, and r t is the domestic real interest rate. Uncovered interest rate parity equates the domestic nominal interest rate to the foreign interest rate (i t * ) plus the expected rate of depreciation ( ~ , + ~ - e , ) . Nonsystematic deviations from uncovered interest rate parity are represented by a random variable E3t: i,=i,*
+(et+l-et)
+E3t-
(3)
The real interest rate is defined as: r, = i , - ( / 3 , + , - - P t )
+'4,
(4)
where (/3r+ 1 - P t ) is the expected inflation rate. Most open economy price adjustment rules, such as in Dornbusch (1976) and Mussa (1981), are not consistent with a unit root in the price level. We specify price adjustment to be purely forward-looking, p,-p,_,
= a 3 , [ ( ~ , + / 3 , * ) - ( e , _ , + P t - , *)] + es,
(5)
where the rate of inflation responds to the rate of change in the equilibrium price level, the latter being defined as the price level consistent with purchasing power parity. This is a generalization of the rule in Papell (1997), which constrains a31 = 1. Mussa (1981) constrains a31 = 1 and includes a backward-looking term. l~_ For countries which allow their exchange rates to float, the expectation of exchange rate depreciation causes inflation, and it is appropriate to constrain a31 = 1. For the countries in this study, which either have a basket peg within a band or a managed float, that constraint is not appropriate and a3~ can even be negative. Credible government intervention to keep the exchange rate within a particular range would discourage economic agents from making frequent revisions in their exchange rate expectations. However, since exchange rates would still fluctuate within the limits set by the band in a basket peg or range in a managed float, a negative a31 in Eq. (5) reflects the positive impact of exchange rate depreciation on domestic prices.
H The n o m i n a l effective e x c h a n g e rate, rather than the bilateral rate, is used so as to h a v e a more general measure o f the e x c h a n g e rate. A n increase in e is a depreciation of the currency. 12 Aside f r o m e x o g e n o u s prices, these are the only specifications w h i c h we h a v e been able to derive that impose a unit root in the price level in the r e d u c e d f o r m of the model.
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The m o n e y supply is determined e n d o g e n o u s l y , and depends on domestic prices, domestic output, and the n o m i n a l exchange rate: 13 mt = wi Pt + w2 Yt q- w3et "-1-~.6t
(6)
The e n d o g e n o u s m o n e y supply specification is appropriate for countries, including those in this study, which actively m a n a g e their exchange rates. F o r e i g n prices, foreign output and foreign interest rate are considered exogen o u s and specified as unit root processes: P t * = P t - i * "~- U3t, Y, * = Y,- 1 * + u4t, and it* =it-l*
(7)
+uSt,
where the disturbance terms ( u ' s ) m a y be serially correlated. The exogeneity of the foreign variables can be justified by the small country assumption. Solving for et+ 1 from Eqs. ( 1 ) - ( 7 ) , we get: et+l = 8let + 62Pt + 63Pt *
+
64Yt
* "b
65i t *
"b
Ult ,
(8)
where b 0 = 1 - w2; b 1 = azl - az3a31; b 2 = ae3b 0 + a~l; b 3 = b 2 - boae3a31; 61 = (bob I + b 2 - w3)/b3; 6 2 = (1 -- boa21 - w j ) / b 3 ; 6 3 = boa21/b3; 6 4 = b 0 a22//b3; and 65 = - b 2 / b 3. Substituting Eqs. (7) and (8) into Eq. (5), we have: Pt = "Yler-1 + T2P~-I + "Y3P~-1 * + Y4Yt-1 * + Y s i t - i * + uzt,
(9)
where Yl = a31(61 - 1); Y2 = 1 + a31 62; "Y3 = a31 63; "Y4 = a31 64; and Y5 = a31 65. The exact specification of the disturbance terms ( u ' s ) is made both to eliminate serial correlation in the estimated residuals and for reasons of parsimony. T h e y are generated by an A R ( 2 ) p r o c e s s , u~t = Crjl u jr_ 1 + aj2 ul t-2 + ~Tj,, J = I . . . . . 5, where the ~7's are serially uncorrelated. The AR(2) is inverted to produce an A R M A ( 1 , 4 ) w h e n the model is estimated. The solution for the m o d e l involves the use o f the multivariate version o f the m e t h o d o f u n d e t e r m i n e d coefficients. 14 The variables are represented in infinite m o v i n g average form and represented as: 5 oc
Zkt = Z E I~kjiUjt-i '
k = 1 . . . . . 5.
(10)
j=li=O where Z t = ( Z I , , Z2,, Z3t, Z4,, Z s t ) ' = ( e t , Pt, Pt *, Yt *, i t * Y , u , = ( u j , , u3t, u4t, ust)', and the u n d e t e r m i n e d coefficients are the 7r's.
u2,,
~3 The propriety of the endogenous money supply specification was examined using the likelihood ratio test against the alternative formulation of money supply as 'exogenous' with w~, w2 and w3 equal to zero. We reject the 'exogenous' in favor of the endogenous specification at the 5% level of significance for all three countries. 14 Taylor (1986) describes the solution method.
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427
Eq. (10) is substituted into Eqs. (7)-(9) which result in the identities:
7r~ = 617r~0 + 1, 7Tljl=~I"tTIjO-r8 j , 77"211= )'I"B'I10, "B'2jI = )'I'Wlj0"4-)'j,j = 2 . . . . . 5, 7rjjo = 1, and
( 11 )
~ i / I = 1.
7rk)i+ I = QTrkji,
j = 1 . . . . . 5, i = 1 . . . . . ~,
(12)
where rr k = (Tr 1, 7r 2, ~'3, ~'4, 7r5)' and Q is a 5 × 5 matrix given by:
)'l i]
)'2 ~2 )'3 ~3 )'4 ~4 )'5 ~i 0 0 0
1 0 0
0 1 0
Further specification is required for a unique solution because each group of two equations in Eq. (11) has three unknowns, 7r~j~, 7rzj ~ and 7rj~0. The general solution to Eq. (12) is: IIkj i = ~_, ChDhkAih,
h=l
(13)
where the C ' s are arbitrary constants, the D ' s are the characteristic vectors and the A's are the characteristic roots. It can be shown that A~ = ( 6 ~ + 62a31) and )t z = 1. By Eq. (7), the other roots, )t3, A4 and A5, equal 1. The coefficient of A~ is set equal to zero to obtain a unique solution. This differs conceptually, although not algebraically, from the usual assumption that the conditionally expected time paths of the variables be stable. In this case, the assumption is that the conditionally expected paths follow unit root processes. The cointegrating vector enters into the reduced form as: e t = R 2 p t + R 3 p t * + R4Yt * + R s i t * ,
(14)
where the R ' s are expressed as combinations of characteristic roots and structural coefficients. The other components of the reduced form are the autoregressive and the moving average terms. Taking the infinite moving average representation implicit in Eqs. (7)-(9) and truncating it at the fourth order produces an ARMA(1,4) model, z, = A z t _ ,
+ B(L)v,,
(15)
where z, = ( e,, p,, Pt *' Yt *, it * ) " Ut =-" ( Ult' U2t' U3t' U4t' L!5t)t' and A and B are 5 × 5 matrices. A and B are nonlinear combinations of the a's, w's, 8 ' s and )"s. The v's, which are combinations of r/'s, are written so the matrix of the zero-lag
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coefficients is the identity matrix. Since A2, A3, A4, and A5 equal 1, the elements of the matrix A which determine domestic prices and the exogenous variables also equal 1, and, therefore, impose unit roots in the reduced form in domestic prices and the exogenous variables. We implement theories consistent expectations, where economic agents possess qualitative, but not quantitative knowledge of the economic model in reduced form, by estimating Eqs. (8) and (9) directly, rather than imposing the constraints from Eqs. (1)-(6). This allows us to construct nested tests of rational and theories consistent expectations. The TCE version of the model consists of Eq. (8), which describes the determinants of exchange rate expectations; Eq. (9), which shows the determinants of prices; and Eq. (7) which defines the foreign variables. The solution to the TCE is similar to the one outlined above except that no structural constraints are imposed on the 6's and 3,'s.
3. Estimation results The model is estimated for Malaysia, Philippines and Singapore for quarterly data starting in 1979 (I) to 1995 (II). The data are from the International Financial Statistics (IFS-CD, February 1996). The exchange rate used is the nominal effective exchange rate; the price level is the consumer price index (CPI); real income is GDP in 1990 prices; the interest rate is the money market rate for Malaysia, the 90-day treasury bill rate for the Philippines, and the three-month interbank rate for Singapore; and the money supply is M1. Foreign variables are weighted averages of each country's trading partners' prices, real income and interest rate. 15 The weights are based on the volume of trade. The model derived above assumes that all variables which enter the reduced form contain unit roots and, as specified in Eq. (14), the exchange rate is cointegrated with its fundamentals. We test these assumptions prior to estimating the model. 16 Since most of the variables are obviously trending, we performed tests which included both a constant and a time trend. Table 1 shows the results of the unit root tests. Using Augmented-Dickey-Fuller (ADF) tests, the unit root null cannot be rejected in favor of trend stationarity at the 5% level of significance in all of the 18 cases. One problem with the ADF test is that, since the null is a unit root, it may inappropriately fail to reject the unit root hypothesis in small samples. Kwiatkowski
15Eleven countries, accounting for approximately 75% of trade with Malaysia, Philippines and Singapore, are included in the calculation of the weights. 16In order to conservespace, we do not present details regarding these tests, which can be found in references listed in Tables 1 and 2.
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J.D. Alba, D.H. Papell / Journal qf Development Economics 55 (1998) 421-437 Table 1 Unit root tests Augmented Dickey-Fuller (ADF) test Variable
e p p* y* i* m
Malaysia
Philippines
Singapore
Dickey-Fuller t-statistic
Lag
Dickey-Fuller t-statistic
Lag
Dickey-Fuller t-statistic
Lag
-2.19 - 2.27 - 3.39 - 1.84 - 2.25 - 1.56
3 0 4 1 8 7
- 1.60 - 2.29 -2.89 -2.21 - 2.65 -2.52
3 3 7 3 7 0
-2.93 - 2.85 - 3.12 -2.22 - 2.27 - 1.55
6 6 8 3 8 3
KPSS test KPSS statistic Variable
Malaysia
Philippines
Singapore
e p p* y* i* m
0.172 0.173 0.168 0.198 0.089 0.307
0.243 (1.204 0.186 0.201 0.089 0.266
0.148 0.150 0.187 0.198 0.085 0.234
The critical values for the Dickey-Fuller t-statistics are -3.16, -3.49, and -4.13 and for the KPSS statistics are 0.119, 0.146, and 0.216 at the 10, 5, and 1% levels. For the ADF test, the unit root null is rejected if the value of the Dickey-Fuller t-statistic is less than the critical value. For the KPSS test, the null of trend stationarity is rejected if the value of the KPSS statistic is greater than the critical value. The critical values are from Mackinnon (1991) and Kwiatkowski et al. (1992). The lag length for the ADF tests is chosen by the criteria described in Campbell and Perron (1991).
et al. ( 1 9 9 2 ) h a v e r e c e n t l y d e v e l o p e d ( K P S S ) tests w h e r e the null is stationarity, a n d t h e s e r e s u l t s are also r e p o r t e d in T a b l e 1. T h e null o f t r e n d s t a t i o n a r i t y is r e j e c t e d at the 5 % level o f s i g n i f i c a n c e in 15 o u t o f t h e 18 cases, the e x c e p t i o n s b e i n g all t h r e e f o r e i g n i n t e r e s t rates. T h e c o m b i n a t i o n o f the A D F a n d K P S S tests g i v e s s u p p o r t for the a s s u m p t i o n o f u n i t roots in this data. T e s t s for c o i n t e g r a t i o n , w h i c h are r e p o r t e d in T a b l e 2, p r o d u c e m i x e d results. U s i n g E n g l e - G r a n g e r ( E G ) tests, w h i c h are A D F tests o n t h e r e s i d u a l s f r o m a c o i n t e g r a t i n g r e g r e s s i o n , t h e null h y p o t h e s i s o f n o c o i n t e g r a t i o n c a n n o t b e r e j e c t e d for a n y o f the t h r e e c o u n t r i e s . T h e s e tests, h o w e v e r , are s u b j e c t e d to the s m a l l s a m p l e p r o b l e m s d e s c r i b e d a b o v e . A s s h o w n b y H a k k i o a n d R u s h (1991), they are a l m o s t c o m p l e t e l y u n i n f o r m a t i v e w i t h s h o r t s p a n s o f d a t a a n d that t h e i r p o w e r c a n n o t b e i m p r o v e d b y s h i f t i n g to h i g h e r f r e q u e n c y d a t a to i n c r e a s e the n u m b e r o f observations. S h i n ( 1 9 9 4 ) h a s d e v e l o p e d tests, w h i c h are e s s e n t i a l l y K P S S tests o n the r e s i d u a l s o f a c o i n t e g r a t i n g r e g r e s s i o n , for w h i c h the null h y p o t h e s i s is c o i n t e g r a -
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Table 2 Cointegration tests Country
Malaysia Philippines Singapore
Engle-Granger test
KPSS test
Engle-Granger t-statistic
Lag
KPSS statistic
- 2.834 - 1.552 - 2.771
0 7 6
0.042 0.054 0.048
The critical values for the Engle-Granger t-statistic are - 4 . 8 9 , - 5 . 2 2 , and - 5 . 8 7 and for the KPSS statistic are 0.050, 0.061, and 0.087 with four explanatory variables at the 10, 5 and 1% levels. For the Engle-Granger test, the no cointegration null is rejected if the value of the Engle-Granger t-statistic is less than the critical value. For the KPSS test, the null of cointegration is rejected if the value of the KPSS statistic is greater than the critical value. The critical values are from Mackinnon (1991) and Shin (1994). The lag length for the Engle-Granger tests is chosen by the criteria described in Campbell and Perron (1991).
tion. Using the Shin tests, the null of cointegration between the exchange rate and its fundamentals cannot be rejected at the 10% level for any of the three countries. Combined with the rejection, using the KPSS test, of the unit root null for all three countries at the 5% level, this constitutes some evidence of cointegration considering the span of the data. In the theories consistent expectations framework of Goldberg and Frydman (1993), expectations functions may be periodically unstable, causing parameter instability in the reduced form. While testing for parameter stability when the date of the structural shift is unknown a priori, as in Andrews (1993), would be desirable, it is not possible with our small number of observations. We investigate the possibility of a shift in the trend function of the exchange rate. Using the tests developed by Vogelsang (1997), which allow for the presence of serial correlation and remain valid whether or not a unit root is present in the series, we cannot reject the no-trend-break null at the 5% level for Philippines and Singapore and at the 1% level for Malaysia. 17 Based on these results, we estimate the model in levels, imposing cointegration between the exchange rate and its fundamentals. We make this choice for two reasons. First, even though the sample period of 1979 to 1995 is very short, we find some evidence of cointegration. Second, the imposition of cointegration overcomes the problem of drawing conclusions from estimates in first-differences when the theory is in levels, is
17 The test is described and critical values provided in Vogelsang (1997). The values of the SupFt tests for exchange rate are 27.28 for Malaysia, 23.30 for Philippines and 21.88 for Singapore. The critical values are 22.60, 25.27 and 30.44 for significance levels of 10%, 5% and 1%, respectively. 18 Goldberg and Frydman (1996a) argue against the practice of first differencing after failing to reject the null of no cointegration.
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431
Following Phillips (1991), we use full system maximum likelihood methods to estimate the model. He shows that these methods are asymptotically optimal if they incorporate the correct prior restriction that n - r unit roots are present in the system, where n is the number of variables and r is the number of cointegrating vectors. If the unit roots are imposed rather than estimated, conventional asymptotic theory forms a valid basis of inference and standard likelihood ratio tests, which rely on tabulations of the chi-squared distribution, can be utilized. The reduced form of the model consists of the set of Eqs. (14) and (15), which are jointly estimated. With 5 variables and 1 cointegrating vector, we impose 4 unit roots. The RE model has 67 constrained parameters: 4 contemporaneous, which are elements of the cointegrating vector, 52 moving average, and 11 autoregressive. These are based on 18 estimated parameters--8 structural ( a ' s and w's), and 10 serial correlation (ct's). The TCE model is not bound by the 8 structural parameters. The 6 ' s in Eq. (8) and the y ' s in Eq. (9) are directly estimated together with the serial correlation parameters. The TCE model has 20 estimated parameters, two more than the RE model, The maximum likelihood estimates of the RE model's structural coefficients and its corresponding asymptotic t-values are shown in Table 3. The structural parameter estimates are of the expected sign, of reasonable magnitude and mostly significant. ~9 For all three countries, the (negative of the) interest rate semi-elasticity (alj) of money demand, the coefficients of the real exchange rate (a21) and the nominal interest rate (a23) in the output equation are positive and significant. The coefficient of foreign output (a22) is positive and significant for Malaysia and the Philippines but negative and significant for Singapore. The price adjustment coefficient (a3~) is negative and significant for all three countries. As discussed above, this result is consistent for the countries in this study, which have exchange rate regimes of either a basket peg within a band or a managed float. The price coefficient in the money supply equation (w 1) is negative and significant for Malaysia and Singapore but positive and significant for the Philippines. The real output coefficient (w 2) is positive and significant for Malaysia and the Philippines but negative and significant for Singapore. The effective exchange rate coefficient, w 3, is positive and significant for Malaysia and Philippines but negative and significant for Singapore. The results show that while monetary policy in the Philippines partly accommodated inflation, monetary policy in Malaysia and Singapore was 'tight' or anti-inflationary. Estimates of the TCE model are shown in Table 4. The parameter estimates for the TCE model are mostly significant. The coefficients of the 8's, except for 6~, are mostly significant but with small magnitudes. The coefficient of 61 is positive
~9Unless otherwise stated, we consider a critical value of 1.96 at the 5% level of significance.
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Table 3 Rational expectations model Parameter
a 1~ azj a22 az3 a3~ wj w2 w3 oq 1 a~2 Ol21 0/22 0/31 0'32 ol41 0/42 0/51 0/52
Malaysia
Philippines
Estimate
Asymptotic t-value
Estimate
Asymptotic t-value
Estimate
Asymptotic t-value
1.165 0.054 0.028 1.403 --0.460 --0.270 0.837 0.341 0.570 -0.117 0.426 - - 0.084 0.106 -0.032 0.234 - 0.001 0.196 -0.309
24.52 30.68 3.62 31.46 -- 13.17 -- 16.75 32.70 16.87 13.39 -41.62 3.62 - 0.66 1.18 -0.39 4.81 -0.14 41.73 -27.12
1.322 0.327 0.092 0.446 --0.258 0.361 0.518 0.563 0.834 -0.119 0.314 0.244 0.100 0.037 0.249 -0.031 0.239 -0.309
5.84 4.55 2.20 4.04 --4.45 4.27 4.52 7.24 6.16 -2.93 3.38 2.81 1.15 0.44 2.62 - 1.95 3.29 -4.66
0.958 0.059 -- 0.216 0.553 --0.549 --0.171 -- 0.039 --0.281 0.510 0.242 0.352 - 0.183 0.094 -0.030 0.232 -0.021 0.119 -0.269
5.52 10.08 -- 6.90 7.67 -- 10.72 --6.02 -- 7.36 --9.06 8.69 7.74 4.27 - 3.28 1.13 -0.41 8.40 -6.42 2.44 -2.12
Parameter values implied by the estimates Parameter Malaysia 31 32 33 34 3S 3'1 "Y2 Y3 Y4 Y5 R2 R3 R 4
Rs Log likelihood
and
Singapore
0.950 0.241 0.021 0.011 - 0.756 0.023 0.889 - 0.009 -- 0.005 0.348 4.856 0.416 0.215 - 15.228 931.89
Philippines
Singapore
0.745 0.302 0.099 0.028 - 0.965 0.065 0.922 - 0.026 - 0.007 0.249 1.186 0.388 0.109 - 3.789 855.14
significant for the three countries.
1.185 0.600 0.033 -0.121 - 0.829 - 0.102 0.671 - 0.018 0.067 0.455 - 3.240 -0.179 0.656 4.481 1002.83
Next period's
expected exchange
rate
( ~ , + l ) is m a i n l y i n f l u e n c e d b y c u r r e n t e x c h a n g e r a t e e t. T h i s r e s u l t is c o n s i s t e n t with
exchange
currencies and
rate
regimes
changes
that peg
the
domestic
only within a given band.
currency
to
Furthermore,
a basket 82
shows
of a
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J.D. AIba, D.H. Papell / Journal of Development Economics 55 (1998) 421-437 Table 4 Theories consistent expectations model Parameter
Malaysia
61 62
64
~5 ~1 TZ Y3 ~4 Y5 ~11 %2 a21 ~72 a3~ a32 ~41 a42 a51 a52
Asymptotic t-value
Estimate
Asymptotic t-value
Estimate
Asymptotic t-value
0.855 0.121 -0.073 --0.083 0.087 0,090 0.627 0.405 0.103 -0.274 1.186 0,308 0.145 0.192 0.059 0.074 0.325 0,173 0.237 -0.094
12.56 - 1.98 - 1.17 -- 1.92 1.09 2.06 2.64 2.38 1.48 -0.83 3.58 0.81 0.62 1.37 0.60 0.91 3.13 1.97 2.47 - 1.15
0,934 0.096 - 0.020 --0.042 -0.089 0.165 0,645 0.142 0.312 0.185 1.880 0,094 0.289 0.206 0.159 0.111 0.292 0.011 0.294 -0.071
35.45 2.95 -0.65 -- 2.19 - 3.93 5.14 9.15 0.84 4.01 3.51 3.94 3.77 3.10 2.37 1.67 1.27 2.93 1.40 5.13 - 2.34
0.906 -0,264 0,026 0.022 - 0.162 --0.082 0.718 0,033 0.015 -0.157 3.782 - 0.478 0.278 -0.100 -0.089 --0.532 0.168 - 0.058 - 0.031 -0.112
35,64 - 3,55 1.50 2.60 - 3.06 --4.55 11.66 0.97 0.20 - 2.91 3.00 - 2.72 2.74 - 1.28 - 1.08 -- 0.81 2.32 - 1.27 - 2.38 - 2.41/
Parameter values implied by the estimates Parameter Malaysia R2 R3 R4 R5 Log likelihood 2 (L T - LR)
Singapore
Estimate
--
63
Philippines
Philippines
Singapore
-0,835 - 0.508 -0.574 0.601 981.27
1.450 -0.307 -0.635 - 1.357 872.40
- 2.818 -0.279 0.237 - 1.724 1031.16
98.77
35.52
56.66
L,r and L R are the log likelihoods for the theories consistent expectations (T) and rational expectations (R) models. Then 2(L x - L R) is distributed chi-squared with two degrees of freedom. The relevant critical values of the chi-square distribution are 5.99 and 9.21 at the 5% and 1% levels of significance.
positive and significant i m p a c t o f d o m e s t i c price on e x c h a n g e rate expectation o n l y f o r t h e P h i l i p p i n e s , a l t h o u g h t h e m a g n i t u d e o f t h e c o e f f i c i e n t is o n l y 0 . 0 9 . The
parameter
which
shows
the impact of the effective exchange
rate on
d o m e s t i c p r i c e s , Yl, is p o s i t i v e a n d s i g n i f i c a n t f o r M a l a y s i a a n d t h e P h i l i p p i n e s but negative and significant for Singapore, although the magnitude of the coeffic i e n t is s m a l l f o r M a l a y s i a a n d S i n g a p o r e . P r i c e p e r s i s t e n c e , 3~2, is p o s i t i v e a n d
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J.D. Alba, D.H. Papell / Journal of Development Economics 55 (1998) 421-437
significant for all three countries while the effect of foreign prices on inflation, 3'3, is positive and significant only for Malaysia. The excess foreign demand coefficient, 3'4, is positive and significant for the Philippines and positive but not significant for Malaysia and Singapore. The foreign interest rate coefficient, 3/5, is negative and significant for Singapore but positive and significant for the Philippines. We compare the RE model with the TCE model using the likelihood ratio test, and report the results in Table 4. According to the likelihood ratio test, 2 times the difference between the log likelihoods for the less constrained (TCE) and more constrained (RE) model is distributed chi-squared with 2 degrees of freedom. 20 The RE model can be rejected in favor of the TCE model at the 1% level of significance for all three countries.
4. Policy implications The evidence from the TCE model provides a number of policy implications. Inflation in Malaysia, Philippines and Singapore is significantly affected by movements in their effective exchange rates. This is consistent with the view of Warner and Kreinin (1983) that effective exchange rates still vary even for developing countries with a basket-peg or managed-float exchange-rate regime. While effective exchange rate depreciation significantly affects inflation in the Philippines and to a lesser extent, Malaysia, Singapore's has the reverse effect. The initial downward pressure on the domestic price of imports caused by the appreciation of the Singapore dollar may be offset by the eventual increase in price due to the stronger demand for imports. Singapore's imports, which range from luxury goods to basic necessities such as food and water, make up a large portion of Singapore's consumption. In addition, rising real income in Singapore since the mid-1960s also reinforce stronger demand for imports. Nevertheless, the magnitude of the coefficient, which shows the impact of exchange rates on inflation in Singapore, is small (0.08). Even though Singapore imports most of what its people need, foreign prices have no effect on inflation. Singapore avoided 'imported' inflation by deliberately allowing its exchange rate to sufficiently appreciate so domestic inflation is not affected by increasing foreign prices. On the other hand, higher foreign price is shown to affect inflation in Malaysia. Excess foreign demand significantly affects inflation only in the Philippines. Higher foreign interest rates, while negatively influencing inflation in Malaysia and Singapore, positively affects inflation in the Philippines. The positive impact
20 The degrees of freedom are based on the difference between the number of estimated parameters in the TCE (20) and RE (18) models.
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435
of excess foreign demand and foreign interest rate on Philippine inflation may be due to the heavy debt burden of the Philippines. When foreign demand is strong, resources may be diverted to exports to pay off foreign debt thereby causing shortages and higher domestic prices in the Philippines. On the other hand, higher foreign interest rates and larger interest payments on Philippine debt would not necessarily lead to inflation unless the central bank monetizes part of the debt. The three countries' monetary policies can be deduced from the constrained parameters of the money supply equation in the RE model. 2~ For the Philippines, the coefficients that relate the money supply to prices (w 1), domestic income (w 2 ), and the exchange rate (ws), are all positive and significant, indicating a 'loose' or partly accommodative monetary policy. Fry (1986) traces the excess money supply to the expansion of credit for government bail-out of failing private-sector enterprises. The inflation arising from the monetizing of govemment debt caused a depreciation of the exchange rate, which in tum produced even more inflation because half of Philippine imports are fuel, chemicals and engineering products required in domestic production. In contrast, Malaysia and Singapore are not burdened by debt and both countries have maintained 'tight' or anti-inflationary monetary policies.
5. Conclusions
We examine effective exchange rate fluctuations and determinants of inflation in Malaysia, Philippines and Singapore in the context of a structural open economy macro model which incorporates cointegration between the exchange rate and its fundamentals and an endogenous money supply specification. Two versions of the model are estimated. The first version is a constrained model consistent with rational expectations (RE) and the second version is a semi-constrained model compatible with theories consistent expectations (TCE). The RE model is strongly rejected in favor of the TCE model. We find, as expected in a basket-peg exchange-rate regime, that exchange rate expectations are mainly affected by the current exchange rate. The effective exchange rate variation significantly affects inflation in the three southeast Asian countries, although the magnitude of the coefficients is small for Malaysia and Singapore. This finding is similar to Bahmani-Oskooee and Malixi (1992) for the Philippines and Rana and Dowling (1985) for Singapore but contrary to the result of Rana and Dowling (1985) for Malaysia. 22 Rana and Dowling find effective exchange rate to negatively affect inflation in Malaysia and Singapore, although
2~The TCE model does not provideestimates of the money supply equation. 22 Malaysia and Singaporeare excluded in the Bahmani-Oskooeeand Malixi (1992) but included in Rana and Dowling (1985).
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their estimates are insignificant for Singapore and marginally significant for Malaysia. Our results also show that foreign prices affect inflation in Malaysia while foreign income has a positive effect on inflation in the Philippines. Foreign interest rate has a negative effect on inflation in Singapore but a positive effect in the Philippines. The positive effect of foreign interest rate on inflation in the Philippines m a y be due to the monetization of debt. Evidence of excess money supply is deduced from the RE model. This finding for the Philippines is consistent with the conclusion of Otani (1975), that excess money supply is one of the determinants o f inflation in the Philippines. Overall, the TCE model with cointegration gives insight on effective exchange rate fluctuations and determinants of domestic inflation consistent with the experiences o f Malaysia, Philippines and Singapore. Unlike Malaysia and Singapore, which have had high growth and stable prices, the Philippines experienced slow growth and high inflation during the 1980s. W h i l e inflation in all three southeast Asian countries are shown to be affected by different external factors, Malaysia and Singapore have avoided high inflation despite high levels of economic growth through 'tight' or anti-inflationary monetary policy, whereas, 'loose' or partly accommodative monetary policy in the Philippines resulted in high inflation.
References Andrews, D., 1993. Test for parameter instability and structural change with unknown change point. Econometrica 61, 821-856. Balassa, B., Williamson, J., 1987. Adjusting to Success: Balance of Payments Policies in East Asian NICs, Institute of International Economics, Washington, DC. Bahmani-Oskooee, M., Malixi, M., 1992. Inflationary effects of changes in effective exchange rates: LDCs experience. Appl. Econ. 24, 465-471. Campbell, J., Perron, P., 1991. Pitfalls and opportunities: What macroeconomists should know about unit roots. In: Blanchard, O., Fisher, S., (Eds.), NBER Macroeconomics Annual, 1991, 141-201. Dornbusch, R., 1976. Expectations and exchange rate dynamics. J. Political Econ. 84, 1161-1176. Edwards, S., 1989. Real Exchange Rates, Devaluation, and Adjustment, MIT Press, Cambridge, MA. Frankel, J., 1979. A theory of floating exchange rates based on real interest differentials. Am. Econ. Rev. 69, 601-622. Fry, M., 1986. Financial structure, financial regulation and financial reform in the Philippines and Thailand, 1960-1984. In: Cheng, H.-S., (Ed.), Financial Policy and Reform in Pacific Basin Countries, D.C. Heath and Company, Washington, DC. Frydman, R., Phelps, E., 1990. Pluralism of Theories Problems in Post-Rational-ExpectationsModeling, presented at the 1990 Sienna Summer Workshop in Expectations and Learning. Goldberg, M., Frydman, R., 1993. Theories Consistent Expectations and Exchange Rate Dynamics. In: Frisch, H., Worgoter, A. (Eds.), Open Economy Macroeconomics, Macmillan, 377-399. Goldberg, M., Frydman, R., 1996a. Empirical exchange rate models and shifts in the co-integrating vector. Struct. Change Econ. Dynamics 7, 55-78. Goldberg, M.D., Frydman, R., 1996b. Imperfect knowledge and behavior in the foreign exchange market. Econ. J. 106, 869-893.
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Hakkio, C., Rush, M., 1991. Cointegration: how short is the long run?. J Int. Money and Finance 10, 571-581. International Monetary Fund, Annual Report on Exchange Arrangements and Exchange Restrictions, 1986, Washington, DC. Kwiatkowski, D., Phillips, P., Schmidt, P., Shin, Y., 1992. Testing the null hypothesis of stationarity against the alternative of a unit root: how sure are we that economic time series are non-stationary?. J. Econometrics 54, 159-178. Lira, J., 1987. The new structuralist critique of the monetarist theory of inflation: the case of the Philippines. J. Dev. Econ. 17, 175-199. Mackinnon, J., 1991. Critical Values for Cointegration Tests. In: Engle, R., Granger, C. (Eds.), Readings in Cointegration, Oxford Univ. Press, Oxford. McNelis, P.D., 1987. Indexing, exchange rate policy and inflationary feedback effects in Latin America. World Dev. 15, 1107-1117. Meese, R., Rogoff, K., 1983. Empirical exchange rate models of the seventies. J. Int. Econ. 14, 3-24. Morrison, R., 1987. Inflation in Portugal, 1953-1980: an econometric analysis. Kyklos 40, 217-219. Moreno, R., 1988. Exchange Rate and Monetary Policy in Singapore and Hong Kong. In: Chen, H.-S., Glick, R. (Eds.), Monetary Policy Changes in Pacific Basin Countries, Kluwer Academic Publishers, Boston. Mussa, M., 1981. Sticky prices and disequilibrium adjustment in a rational model of the inflationary process. Am. Econ. Rev. 71, 1020-1027. Noland, M., 1990. Pacific basin developing countries: Prospects for the future, Institute of Institutional Economics, Washington, DC. Otani, I., 1975. Inflation in an open economy: a case study of the Philippines. IMF Staff Papers 22, 750-774. Papell, D., 1988. Expectations and exchange rate dynamics after a decade of floating. J. Int. Econ. 25, 303-317. Papell, D., 1997. Cointegration and exchange rate dynamics. J. Int. Money and Finance. 16, 445-459. Phillips, P., 1991. Optimal inference in cointegrated systems. Econometrica 59, 283-306. Rana, P., Dowling, J., 1985. Inflationary effects of small but continuous changes in effective exchange rates: nine Asian LDCs. Rev. Econ. Stat. 67, 496-500. Shin, Y., 1994. A residual-based test of the null of cointegration against the alternative of no cointegration. Econometric Theory 10, 91 - 115. Taylor, J., 1986. New econometric approaches to stabilization policy in stochastic models of macroeconomic fluctuations. In: Griliches, Z., Intrilligator, M. (Eds,), Handbook of Econometrics, NorthHolland, Amsterdam. Vogelsang, T., 1997. Wald-type tests for detecting shifts in the trend function of a dynamic time series. Warner, D,, Kreinin, M., 1983, Determinants of international trade flows. Rev. Econ. Stat. 65, 96-140. Wong, J., 1990. Singapore's experience of industrial restructuring, lessons for the other Asian NICs. In: Naya, S., Takayama, A. (Eds.), Economic Development in East and Southeast Asia, Institute of Southeast Asian Studies, Singapore.
ij ~
ELSEVIER
Journal of DevelopmentEconomics Vol. 55 (1998) 421-437
JOURNAL OF Development ECONOMICS
Exchange rate determination and inflation in Southeast Asian countries Joseph D. Alba a, David H. Papell b,, a Department of Economics, Nanyang Technological Universi~, 2263, Singapore b Department of Economics, Universi~ of Houston, Houston, TX 77204, USA
Received 4 March 1994; revised 6 August 1996; accepted 31 December 1996
Abstract We estimate structural open economy models consistent with rational (RE) and theories consistent (TCE) expectations which incorporate cointegration to examine exchange rate determination and inflation in three southeast Asian countries. The RE model is rejected in favor of the TCE model. Our results show that while inflation in all three countries is affected by different external factors, Malaysia and Singapore avoided high inflation despite high levels of economic growth through 'tight' monetary policy. In contrast, the Philippines had high inflation, even with a stagnant economy, due to 'loose' monetary policy and the monetization of government debt. © 1998 Elsevier Science B.V. JEL classification: F41; O12; C32 Keywords: Exchange rates; Inflation;Cointegration;Southeast Asia
1. Introduction The system of floating exchange rates since 1973 has been characterized by high variability and large fluctuations in exchange rates. Much empirical work has been done to explain these variations in developed countries. ~ Since most
* Corresponding author. i These studies include Frankel (1979), Meese and Rogoff (1983) and Papell (1988). 0304-3878/98/$19.00 © 1998 Elsevier Science B.V. All rights reserved. PII S0304- 3878(98)00043-1
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developing countries peg their exchange rates to a major currency or a basket of currencies, very little attention has been given to exchange rate fluctuations in these countries. However, as Warner and Kreinin (1983) argue, even if developing countries peg their currencies to major currencies, their effective exchange rates will still vary so long as major currencies fluctuate against each other. If the effective exchange rates of developing countries vary because of the variation in major currencies, then they would exhibit the same characteristics as major currencies. Variation in effective exchange rates may also have effects on domestic inflation in developing countries. 2 The focus of the few studies that include exchange rate variation in developing countries is on its impact on domestic inflation. Lim (1987), McNelis (1987) and Morrison (1987) have used bilateral exchange rates, which have been occasionally devalued, as one of the determinants of inflation. Rana and Dowling (1985) and Bahmani-Oskooee and Malixi (1992) use a single-equation model with effective exchange rates as one of the variables that explain inflation. Rana and Dowling (1985) find that exchange rate changes and excess money supply do not have a significant effect in nine Asian countries. 3 Bahmani-Oskooee and Malixi (1992) also examine the determinants of inflation in four Asian and nine nonAsian developing countries. 4 They respecify the lag structure in the model 5 used by Rana and Dowling (1985) and show that effective exchange rate depreciation affects domestic inflation in nine of the 13 countries, including the Philippines and Thailand. Furthermore, excess money supply is found to have a positive significant effect on inflation except for the Philippines. This is contrary to the results of Otani (1975), who used an open economy macroeconomic model with an endogenous money supply specification to examine the impact of the money supply on inflation in the Philippines. The lack of robustness of Rana and Dowling's single-equation model may be due to misspecifications that can be corrected only by a structural model. 6 In this paper, we examine effective exchange rate determination and the factors that affect domestic prices in three southeast Asian countries from 1979 to 1995. Malaysia, Philippines and Singapore are chosen because differences in their
2 Edwards (1989) analyzes the behavior of real exchange rates in developing countries. 3 These countries are India, Indonesia, South Korea, Malaysia, Nepal, Philippines, Singapore, Taiwan, and Thailand. 4 The countries included in their study are Brazil, Dominican Republic, Egypt, Greece, India, South Korea, Mexico, Pakistan, Peru, Philippines, Portugal, Thailand, and Turkey. 5 Bahmani-Oskooeeand Malixi (1992) rely on the Almon lag specification instead of the simple lag and Koyck lag structures used by Rana and Dowling (1985). 6 Bahmani-Oskooeeand Malixi (1992) attribute the differences in their's and the results of Rana and Dowling (1985) on the lag structure specification.
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423
economic policies offer interesting comparisons on the varying results of such policies on the macroeconomy. 7 Singapore is one of the NICs in Asia which has experienced phenomenal growth since the 1970s. High growth is attributed to the shift of the economy to labor-intensive manufacturing from entreprt trade in the late 1960s 8 and the diversification of the economy into more capital-intensive manufacturing, tourism and financial and banking services starting in the late 1970s. 9 Despite Singapore's sustained growth, sound macroeconomic policies have kept inflation within the 0-10% range. Malaysia's growth of 8-10% since the late 1980s mimics Singapore's exceptional growth in the 1960s and early 1970s. The high growth rate coincides with the shift of the economy from primary to manufactured-good exports. As in Singapore, Malaysia maintained a low inflation within the range of 0-10%. The Philippines is a notable exception to the booming economies of southeast Asia. The country experienced both slow growth and high inflation in the 1980s, with the inflation rate reaching a peak of 49% in 1984. The Pacific Economic Development Report (1992-1993) identifies high inflation as one of the reasons the Philippines could not attract more foreign direct investment (FDI) as other countries in the region have done to aid in economic development. The three countries all manage their exchange rates. Balassa and Williamson (1987) describe the exchange rate policy of Singapore as a basket peg within a band. J0 The International Monetary Fund (1986) reports that the exchange rate of Malaysia is a basket peg while Fry (1986) characterizes the exchange rate policy in the Philippines as a managed float. We specify forward-looking price adjustment, following Papell (1997), in order to incorporate cointegration between the exchange rate and its fundamentals. The model is consistent with long-run equilibrium relationships such as purchasing power parity and monetary neutrality, while also tracking short-run movements in effective exchange rates. In contrast with Papell (1997), we employ an endogenous money supply specification and a more general price adjustment equation, which are more appropriate for countries with managed exchange rates, to examine the determinants of inflation. Two versions of the model are estimated. The first version is a constrained rational expectations (RE) model, with the constraints derived from the structural equations and the assumptions necessary to derive a unique rational expectations
7 The inclusion of Thailand and Indonesiaas well as the Newly IndustrializingCountries (NICs) of Hong Kong, South Korea and Taiwan would have made the comparisons even more interesting. However, quarterly real GDP and/or nominal GDP are not available for other east and southeast Asian countries. 8 Noland (1990), p. 33. 9 Wong(1990), p. 58. ]0 On the other hand, Moreno(1988) portrays Singapore'sexchange rate as a managed float.
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solution. The second version is a semi-constrained model, compatible with the concept, proposed by Frydman and Phelps (1990) and developed by Goldberg and Frydman (1993, 1996a, b), of theories consistent expectations (TCE). Although the rational expectations assumption, that economic agents have quantitative knowledge of the model, is clearly debatable, development of serious alternative hypotheses has proceeded slowly. In the TCE, the determinants of exchange rates and prices are qualitatively, but not quantitatively, consistent with the fully constrained rational expectations model. Both versions of the model are estimated for Malaysia, Philippines and Singapore using full system maximum likelihood methods with quarterly data from 1979 (I) to 1995 (II). Although the RE model has mostly significant results, with coefficients within the range of expected signs and magnitudes, the likelihood ratio test rejects the RE model in favor of the TCE model. The TCE model provides insight into understanding effective exchange rate dynamics and the determinants of domestic inflation in southeast Asian countries. While exchange rate expectations are affected mainly by the current exchange rate, effective exchange rate variation significantly affects inflation in all three countries. The other determinants of inflation vary among the different countries: Malaysia, with an average inflation rate of 3.7%, is affected by increases in foreign prices; Singapore, which had an average inflation rate of 2.9%, is affected by lower foreign interest rates; and the Philippines, with an average inflation rate of 15% in the 1980s, is affected by excess foreign demand and higher foreign interest rates. The positive impact of foreign interest rates may be due to the monetization of government debt. The RE model provides information about the three countries' monetary policies. The Philippines is shown to have a 'loose' or partly accommodating monetary policy while monetary policy for Malaysia and Singapore is 'tight' or stabilizing. The paper is organized as follows: Section 2 describes the model and its solution; Section 3 discusses the estimation results; Section 4 considers the policy implications; and the conclusions are in Section 5.
2. The model
We utilize a small open economy model, the demand side of which is based on Dornbusch (1976) and follows Papell (1997). Asset market equilibrium equates money supply and money demand, with money demand a function of domestic income and the domestic interest rate: mt -Pt
= Y, - a l l i t + ~ l t ,
(1)
where m t is the log of the domestic money supply, Pt is the log of the price level, is the log of domestic real income and i t is the domestic nominal interest rate.
Yt
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425
The disturbance terms (E's) are random variables which may be serially correlated. Equilibrium in the goods market is expressed as: Y, = a21( et + Pt * -- P,) + a22 Y, * - a23r, + e2,,
(2)
where e, is the log of the nominal effective exchange rate, j f Pt * is the log of foreign prices, Yt * is the log of foreign output, and r t is the domestic real interest rate. Uncovered interest rate parity equates the domestic nominal interest rate to the foreign interest rate (i t * ) plus the expected rate of depreciation ( ~ , + ~ - e , ) . Nonsystematic deviations from uncovered interest rate parity are represented by a random variable E3t: i,=i,*
+(et+l-et)
+E3t-
(3)
The real interest rate is defined as: r, = i , - ( / 3 , + , - - P t )
+'4,
(4)
where (/3r+ 1 - P t ) is the expected inflation rate. Most open economy price adjustment rules, such as in Dornbusch (1976) and Mussa (1981), are not consistent with a unit root in the price level. We specify price adjustment to be purely forward-looking, p,-p,_,
= a 3 , [ ( ~ , + / 3 , * ) - ( e , _ , + P t - , *)] + es,
(5)
where the rate of inflation responds to the rate of change in the equilibrium price level, the latter being defined as the price level consistent with purchasing power parity. This is a generalization of the rule in Papell (1997), which constrains a31 = 1. Mussa (1981) constrains a31 = 1 and includes a backward-looking term. l~_ For countries which allow their exchange rates to float, the expectation of exchange rate depreciation causes inflation, and it is appropriate to constrain a31 = 1. For the countries in this study, which either have a basket peg within a band or a managed float, that constraint is not appropriate and a3~ can even be negative. Credible government intervention to keep the exchange rate within a particular range would discourage economic agents from making frequent revisions in their exchange rate expectations. However, since exchange rates would still fluctuate within the limits set by the band in a basket peg or range in a managed float, a negative a31 in Eq. (5) reflects the positive impact of exchange rate depreciation on domestic prices.
H The n o m i n a l effective e x c h a n g e rate, rather than the bilateral rate, is used so as to h a v e a more general measure o f the e x c h a n g e rate. A n increase in e is a depreciation of the currency. 12 Aside f r o m e x o g e n o u s prices, these are the only specifications w h i c h we h a v e been able to derive that impose a unit root in the price level in the r e d u c e d f o r m of the model.
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The m o n e y supply is determined e n d o g e n o u s l y , and depends on domestic prices, domestic output, and the n o m i n a l exchange rate: 13 mt = wi Pt + w2 Yt q- w3et "-1-~.6t
(6)
The e n d o g e n o u s m o n e y supply specification is appropriate for countries, including those in this study, which actively m a n a g e their exchange rates. F o r e i g n prices, foreign output and foreign interest rate are considered exogen o u s and specified as unit root processes: P t * = P t - i * "~- U3t, Y, * = Y,- 1 * + u4t, and it* =it-l*
(7)
+uSt,
where the disturbance terms ( u ' s ) m a y be serially correlated. The exogeneity of the foreign variables can be justified by the small country assumption. Solving for et+ 1 from Eqs. ( 1 ) - ( 7 ) , we get: et+l = 8let + 62Pt + 63Pt *
+
64Yt
* "b
65i t *
"b
Ult ,
(8)
where b 0 = 1 - w2; b 1 = azl - az3a31; b 2 = ae3b 0 + a~l; b 3 = b 2 - boae3a31; 61 = (bob I + b 2 - w3)/b3; 6 2 = (1 -- boa21 - w j ) / b 3 ; 6 3 = boa21/b3; 6 4 = b 0 a22//b3; and 65 = - b 2 / b 3. Substituting Eqs. (7) and (8) into Eq. (5), we have: Pt = "Yler-1 + T2P~-I + "Y3P~-1 * + Y4Yt-1 * + Y s i t - i * + uzt,
(9)
where Yl = a31(61 - 1); Y2 = 1 + a31 62; "Y3 = a31 63; "Y4 = a31 64; and Y5 = a31 65. The exact specification of the disturbance terms ( u ' s ) is made both to eliminate serial correlation in the estimated residuals and for reasons of parsimony. T h e y are generated by an A R ( 2 ) p r o c e s s , u~t = Crjl u jr_ 1 + aj2 ul t-2 + ~Tj,, J = I . . . . . 5, where the ~7's are serially uncorrelated. The AR(2) is inverted to produce an A R M A ( 1 , 4 ) w h e n the model is estimated. The solution for the m o d e l involves the use o f the multivariate version o f the m e t h o d o f u n d e t e r m i n e d coefficients. 14 The variables are represented in infinite m o v i n g average form and represented as: 5 oc
Zkt = Z E I~kjiUjt-i '
k = 1 . . . . . 5.
(10)
j=li=O where Z t = ( Z I , , Z2,, Z3t, Z4,, Z s t ) ' = ( e t , Pt, Pt *, Yt *, i t * Y , u , = ( u j , , u3t, u4t, ust)', and the u n d e t e r m i n e d coefficients are the 7r's.
u2,,
~3 The propriety of the endogenous money supply specification was examined using the likelihood ratio test against the alternative formulation of money supply as 'exogenous' with w~, w2 and w3 equal to zero. We reject the 'exogenous' in favor of the endogenous specification at the 5% level of significance for all three countries. 14 Taylor (1986) describes the solution method.
J.D. Alba, D.H. Papell / Journal of Development Economics 55 (1998) 421-437
427
Eq. (10) is substituted into Eqs. (7)-(9) which result in the identities:
7r~ = 617r~0 + 1, 7Tljl=~I"tTIjO-r8 j , 77"211= )'I"B'I10, "B'2jI = )'I'Wlj0"4-)'j,j = 2 . . . . . 5, 7rjjo = 1, and
( 11 )
~ i / I = 1.
7rk)i+ I = QTrkji,
j = 1 . . . . . 5, i = 1 . . . . . ~,
(12)
where rr k = (Tr 1, 7r 2, ~'3, ~'4, 7r5)' and Q is a 5 × 5 matrix given by:
)'l i]
)'2 ~2 )'3 ~3 )'4 ~4 )'5 ~i 0 0 0
1 0 0
0 1 0
Further specification is required for a unique solution because each group of two equations in Eq. (11) has three unknowns, 7r~j~, 7rzj ~ and 7rj~0. The general solution to Eq. (12) is: IIkj i = ~_, ChDhkAih,
h=l
(13)
where the C ' s are arbitrary constants, the D ' s are the characteristic vectors and the A's are the characteristic roots. It can be shown that A~ = ( 6 ~ + 62a31) and )t z = 1. By Eq. (7), the other roots, )t3, A4 and A5, equal 1. The coefficient of A~ is set equal to zero to obtain a unique solution. This differs conceptually, although not algebraically, from the usual assumption that the conditionally expected time paths of the variables be stable. In this case, the assumption is that the conditionally expected paths follow unit root processes. The cointegrating vector enters into the reduced form as: e t = R 2 p t + R 3 p t * + R4Yt * + R s i t * ,
(14)
where the R ' s are expressed as combinations of characteristic roots and structural coefficients. The other components of the reduced form are the autoregressive and the moving average terms. Taking the infinite moving average representation implicit in Eqs. (7)-(9) and truncating it at the fourth order produces an ARMA(1,4) model, z, = A z t _ ,
+ B(L)v,,
(15)
where z, = ( e,, p,, Pt *' Yt *, it * ) " Ut =-" ( Ult' U2t' U3t' U4t' L!5t)t' and A and B are 5 × 5 matrices. A and B are nonlinear combinations of the a's, w's, 8 ' s and )"s. The v's, which are combinations of r/'s, are written so the matrix of the zero-lag
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coefficients is the identity matrix. Since A2, A3, A4, and A5 equal 1, the elements of the matrix A which determine domestic prices and the exogenous variables also equal 1, and, therefore, impose unit roots in the reduced form in domestic prices and the exogenous variables. We implement theories consistent expectations, where economic agents possess qualitative, but not quantitative knowledge of the economic model in reduced form, by estimating Eqs. (8) and (9) directly, rather than imposing the constraints from Eqs. (1)-(6). This allows us to construct nested tests of rational and theories consistent expectations. The TCE version of the model consists of Eq. (8), which describes the determinants of exchange rate expectations; Eq. (9), which shows the determinants of prices; and Eq. (7) which defines the foreign variables. The solution to the TCE is similar to the one outlined above except that no structural constraints are imposed on the 6's and 3,'s.
3. Estimation results The model is estimated for Malaysia, Philippines and Singapore for quarterly data starting in 1979 (I) to 1995 (II). The data are from the International Financial Statistics (IFS-CD, February 1996). The exchange rate used is the nominal effective exchange rate; the price level is the consumer price index (CPI); real income is GDP in 1990 prices; the interest rate is the money market rate for Malaysia, the 90-day treasury bill rate for the Philippines, and the three-month interbank rate for Singapore; and the money supply is M1. Foreign variables are weighted averages of each country's trading partners' prices, real income and interest rate. 15 The weights are based on the volume of trade. The model derived above assumes that all variables which enter the reduced form contain unit roots and, as specified in Eq. (14), the exchange rate is cointegrated with its fundamentals. We test these assumptions prior to estimating the model. 16 Since most of the variables are obviously trending, we performed tests which included both a constant and a time trend. Table 1 shows the results of the unit root tests. Using Augmented-Dickey-Fuller (ADF) tests, the unit root null cannot be rejected in favor of trend stationarity at the 5% level of significance in all of the 18 cases. One problem with the ADF test is that, since the null is a unit root, it may inappropriately fail to reject the unit root hypothesis in small samples. Kwiatkowski
15Eleven countries, accounting for approximately 75% of trade with Malaysia, Philippines and Singapore, are included in the calculation of the weights. 16In order to conservespace, we do not present details regarding these tests, which can be found in references listed in Tables 1 and 2.
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J.D. Alba, D.H. Papell / Journal qf Development Economics 55 (1998) 421-437 Table 1 Unit root tests Augmented Dickey-Fuller (ADF) test Variable
e p p* y* i* m
Malaysia
Philippines
Singapore
Dickey-Fuller t-statistic
Lag
Dickey-Fuller t-statistic
Lag
Dickey-Fuller t-statistic
Lag
-2.19 - 2.27 - 3.39 - 1.84 - 2.25 - 1.56
3 0 4 1 8 7
- 1.60 - 2.29 -2.89 -2.21 - 2.65 -2.52
3 3 7 3 7 0
-2.93 - 2.85 - 3.12 -2.22 - 2.27 - 1.55
6 6 8 3 8 3
KPSS test KPSS statistic Variable
Malaysia
Philippines
Singapore
e p p* y* i* m
0.172 0.173 0.168 0.198 0.089 0.307
0.243 (1.204 0.186 0.201 0.089 0.266
0.148 0.150 0.187 0.198 0.085 0.234
The critical values for the Dickey-Fuller t-statistics are -3.16, -3.49, and -4.13 and for the KPSS statistics are 0.119, 0.146, and 0.216 at the 10, 5, and 1% levels. For the ADF test, the unit root null is rejected if the value of the Dickey-Fuller t-statistic is less than the critical value. For the KPSS test, the null of trend stationarity is rejected if the value of the KPSS statistic is greater than the critical value. The critical values are from Mackinnon (1991) and Kwiatkowski et al. (1992). The lag length for the ADF tests is chosen by the criteria described in Campbell and Perron (1991).
et al. ( 1 9 9 2 ) h a v e r e c e n t l y d e v e l o p e d ( K P S S ) tests w h e r e the null is stationarity, a n d t h e s e r e s u l t s are also r e p o r t e d in T a b l e 1. T h e null o f t r e n d s t a t i o n a r i t y is r e j e c t e d at the 5 % level o f s i g n i f i c a n c e in 15 o u t o f t h e 18 cases, the e x c e p t i o n s b e i n g all t h r e e f o r e i g n i n t e r e s t rates. T h e c o m b i n a t i o n o f the A D F a n d K P S S tests g i v e s s u p p o r t for the a s s u m p t i o n o f u n i t roots in this data. T e s t s for c o i n t e g r a t i o n , w h i c h are r e p o r t e d in T a b l e 2, p r o d u c e m i x e d results. U s i n g E n g l e - G r a n g e r ( E G ) tests, w h i c h are A D F tests o n t h e r e s i d u a l s f r o m a c o i n t e g r a t i n g r e g r e s s i o n , t h e null h y p o t h e s i s o f n o c o i n t e g r a t i o n c a n n o t b e r e j e c t e d for a n y o f the t h r e e c o u n t r i e s . T h e s e tests, h o w e v e r , are s u b j e c t e d to the s m a l l s a m p l e p r o b l e m s d e s c r i b e d a b o v e . A s s h o w n b y H a k k i o a n d R u s h (1991), they are a l m o s t c o m p l e t e l y u n i n f o r m a t i v e w i t h s h o r t s p a n s o f d a t a a n d that t h e i r p o w e r c a n n o t b e i m p r o v e d b y s h i f t i n g to h i g h e r f r e q u e n c y d a t a to i n c r e a s e the n u m b e r o f observations. S h i n ( 1 9 9 4 ) h a s d e v e l o p e d tests, w h i c h are e s s e n t i a l l y K P S S tests o n the r e s i d u a l s o f a c o i n t e g r a t i n g r e g r e s s i o n , for w h i c h the null h y p o t h e s i s is c o i n t e g r a -
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Table 2 Cointegration tests Country
Malaysia Philippines Singapore
Engle-Granger test
KPSS test
Engle-Granger t-statistic
Lag
KPSS statistic
- 2.834 - 1.552 - 2.771
0 7 6
0.042 0.054 0.048
The critical values for the Engle-Granger t-statistic are - 4 . 8 9 , - 5 . 2 2 , and - 5 . 8 7 and for the KPSS statistic are 0.050, 0.061, and 0.087 with four explanatory variables at the 10, 5 and 1% levels. For the Engle-Granger test, the no cointegration null is rejected if the value of the Engle-Granger t-statistic is less than the critical value. For the KPSS test, the null of cointegration is rejected if the value of the KPSS statistic is greater than the critical value. The critical values are from Mackinnon (1991) and Shin (1994). The lag length for the Engle-Granger tests is chosen by the criteria described in Campbell and Perron (1991).
tion. Using the Shin tests, the null of cointegration between the exchange rate and its fundamentals cannot be rejected at the 10% level for any of the three countries. Combined with the rejection, using the KPSS test, of the unit root null for all three countries at the 5% level, this constitutes some evidence of cointegration considering the span of the data. In the theories consistent expectations framework of Goldberg and Frydman (1993), expectations functions may be periodically unstable, causing parameter instability in the reduced form. While testing for parameter stability when the date of the structural shift is unknown a priori, as in Andrews (1993), would be desirable, it is not possible with our small number of observations. We investigate the possibility of a shift in the trend function of the exchange rate. Using the tests developed by Vogelsang (1997), which allow for the presence of serial correlation and remain valid whether or not a unit root is present in the series, we cannot reject the no-trend-break null at the 5% level for Philippines and Singapore and at the 1% level for Malaysia. 17 Based on these results, we estimate the model in levels, imposing cointegration between the exchange rate and its fundamentals. We make this choice for two reasons. First, even though the sample period of 1979 to 1995 is very short, we find some evidence of cointegration. Second, the imposition of cointegration overcomes the problem of drawing conclusions from estimates in first-differences when the theory is in levels, is
17 The test is described and critical values provided in Vogelsang (1997). The values of the SupFt tests for exchange rate are 27.28 for Malaysia, 23.30 for Philippines and 21.88 for Singapore. The critical values are 22.60, 25.27 and 30.44 for significance levels of 10%, 5% and 1%, respectively. 18 Goldberg and Frydman (1996a) argue against the practice of first differencing after failing to reject the null of no cointegration.
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431
Following Phillips (1991), we use full system maximum likelihood methods to estimate the model. He shows that these methods are asymptotically optimal if they incorporate the correct prior restriction that n - r unit roots are present in the system, where n is the number of variables and r is the number of cointegrating vectors. If the unit roots are imposed rather than estimated, conventional asymptotic theory forms a valid basis of inference and standard likelihood ratio tests, which rely on tabulations of the chi-squared distribution, can be utilized. The reduced form of the model consists of the set of Eqs. (14) and (15), which are jointly estimated. With 5 variables and 1 cointegrating vector, we impose 4 unit roots. The RE model has 67 constrained parameters: 4 contemporaneous, which are elements of the cointegrating vector, 52 moving average, and 11 autoregressive. These are based on 18 estimated parameters--8 structural ( a ' s and w's), and 10 serial correlation (ct's). The TCE model is not bound by the 8 structural parameters. The 6 ' s in Eq. (8) and the y ' s in Eq. (9) are directly estimated together with the serial correlation parameters. The TCE model has 20 estimated parameters, two more than the RE model, The maximum likelihood estimates of the RE model's structural coefficients and its corresponding asymptotic t-values are shown in Table 3. The structural parameter estimates are of the expected sign, of reasonable magnitude and mostly significant. ~9 For all three countries, the (negative of the) interest rate semi-elasticity (alj) of money demand, the coefficients of the real exchange rate (a21) and the nominal interest rate (a23) in the output equation are positive and significant. The coefficient of foreign output (a22) is positive and significant for Malaysia and the Philippines but negative and significant for Singapore. The price adjustment coefficient (a3~) is negative and significant for all three countries. As discussed above, this result is consistent for the countries in this study, which have exchange rate regimes of either a basket peg within a band or a managed float. The price coefficient in the money supply equation (w 1) is negative and significant for Malaysia and Singapore but positive and significant for the Philippines. The real output coefficient (w 2) is positive and significant for Malaysia and the Philippines but negative and significant for Singapore. The effective exchange rate coefficient, w 3, is positive and significant for Malaysia and Philippines but negative and significant for Singapore. The results show that while monetary policy in the Philippines partly accommodated inflation, monetary policy in Malaysia and Singapore was 'tight' or anti-inflationary. Estimates of the TCE model are shown in Table 4. The parameter estimates for the TCE model are mostly significant. The coefficients of the 8's, except for 6~, are mostly significant but with small magnitudes. The coefficient of 61 is positive
~9Unless otherwise stated, we consider a critical value of 1.96 at the 5% level of significance.
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432
Table 3 Rational expectations model Parameter
a 1~ azj a22 az3 a3~ wj w2 w3 oq 1 a~2 Ol21 0/22 0/31 0'32 ol41 0/42 0/51 0/52
Malaysia
Philippines
Estimate
Asymptotic t-value
Estimate
Asymptotic t-value
Estimate
Asymptotic t-value
1.165 0.054 0.028 1.403 --0.460 --0.270 0.837 0.341 0.570 -0.117 0.426 - - 0.084 0.106 -0.032 0.234 - 0.001 0.196 -0.309
24.52 30.68 3.62 31.46 -- 13.17 -- 16.75 32.70 16.87 13.39 -41.62 3.62 - 0.66 1.18 -0.39 4.81 -0.14 41.73 -27.12
1.322 0.327 0.092 0.446 --0.258 0.361 0.518 0.563 0.834 -0.119 0.314 0.244 0.100 0.037 0.249 -0.031 0.239 -0.309
5.84 4.55 2.20 4.04 --4.45 4.27 4.52 7.24 6.16 -2.93 3.38 2.81 1.15 0.44 2.62 - 1.95 3.29 -4.66
0.958 0.059 -- 0.216 0.553 --0.549 --0.171 -- 0.039 --0.281 0.510 0.242 0.352 - 0.183 0.094 -0.030 0.232 -0.021 0.119 -0.269
5.52 10.08 -- 6.90 7.67 -- 10.72 --6.02 -- 7.36 --9.06 8.69 7.74 4.27 - 3.28 1.13 -0.41 8.40 -6.42 2.44 -2.12
Parameter values implied by the estimates Parameter Malaysia 31 32 33 34 3S 3'1 "Y2 Y3 Y4 Y5 R2 R3 R 4
Rs Log likelihood
and
Singapore
0.950 0.241 0.021 0.011 - 0.756 0.023 0.889 - 0.009 -- 0.005 0.348 4.856 0.416 0.215 - 15.228 931.89
Philippines
Singapore
0.745 0.302 0.099 0.028 - 0.965 0.065 0.922 - 0.026 - 0.007 0.249 1.186 0.388 0.109 - 3.789 855.14
significant for the three countries.
1.185 0.600 0.033 -0.121 - 0.829 - 0.102 0.671 - 0.018 0.067 0.455 - 3.240 -0.179 0.656 4.481 1002.83
Next period's
expected exchange
rate
( ~ , + l ) is m a i n l y i n f l u e n c e d b y c u r r e n t e x c h a n g e r a t e e t. T h i s r e s u l t is c o n s i s t e n t with
exchange
currencies and
rate
regimes
changes
that peg
the
domestic
only within a given band.
currency
to
Furthermore,
a basket 82
shows
of a
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J.D. AIba, D.H. Papell / Journal of Development Economics 55 (1998) 421-437 Table 4 Theories consistent expectations model Parameter
Malaysia
61 62
64
~5 ~1 TZ Y3 ~4 Y5 ~11 %2 a21 ~72 a3~ a32 ~41 a42 a51 a52
Asymptotic t-value
Estimate
Asymptotic t-value
Estimate
Asymptotic t-value
0.855 0.121 -0.073 --0.083 0.087 0,090 0.627 0.405 0.103 -0.274 1.186 0,308 0.145 0.192 0.059 0.074 0.325 0,173 0.237 -0.094
12.56 - 1.98 - 1.17 -- 1.92 1.09 2.06 2.64 2.38 1.48 -0.83 3.58 0.81 0.62 1.37 0.60 0.91 3.13 1.97 2.47 - 1.15
0,934 0.096 - 0.020 --0.042 -0.089 0.165 0,645 0.142 0.312 0.185 1.880 0,094 0.289 0.206 0.159 0.111 0.292 0.011 0.294 -0.071
35.45 2.95 -0.65 -- 2.19 - 3.93 5.14 9.15 0.84 4.01 3.51 3.94 3.77 3.10 2.37 1.67 1.27 2.93 1.40 5.13 - 2.34
0.906 -0,264 0,026 0.022 - 0.162 --0.082 0.718 0,033 0.015 -0.157 3.782 - 0.478 0.278 -0.100 -0.089 --0.532 0.168 - 0.058 - 0.031 -0.112
35,64 - 3,55 1.50 2.60 - 3.06 --4.55 11.66 0.97 0.20 - 2.91 3.00 - 2.72 2.74 - 1.28 - 1.08 -- 0.81 2.32 - 1.27 - 2.38 - 2.41/
Parameter values implied by the estimates Parameter Malaysia R2 R3 R4 R5 Log likelihood 2 (L T - LR)
Singapore
Estimate
--
63
Philippines
Philippines
Singapore
-0,835 - 0.508 -0.574 0.601 981.27
1.450 -0.307 -0.635 - 1.357 872.40
- 2.818 -0.279 0.237 - 1.724 1031.16
98.77
35.52
56.66
L,r and L R are the log likelihoods for the theories consistent expectations (T) and rational expectations (R) models. Then 2(L x - L R) is distributed chi-squared with two degrees of freedom. The relevant critical values of the chi-square distribution are 5.99 and 9.21 at the 5% and 1% levels of significance.
positive and significant i m p a c t o f d o m e s t i c price on e x c h a n g e rate expectation o n l y f o r t h e P h i l i p p i n e s , a l t h o u g h t h e m a g n i t u d e o f t h e c o e f f i c i e n t is o n l y 0 . 0 9 . The
parameter
which
shows
the impact of the effective exchange
rate on
d o m e s t i c p r i c e s , Yl, is p o s i t i v e a n d s i g n i f i c a n t f o r M a l a y s i a a n d t h e P h i l i p p i n e s but negative and significant for Singapore, although the magnitude of the coeffic i e n t is s m a l l f o r M a l a y s i a a n d S i n g a p o r e . P r i c e p e r s i s t e n c e , 3~2, is p o s i t i v e a n d
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significant for all three countries while the effect of foreign prices on inflation, 3'3, is positive and significant only for Malaysia. The excess foreign demand coefficient, 3'4, is positive and significant for the Philippines and positive but not significant for Malaysia and Singapore. The foreign interest rate coefficient, 3/5, is negative and significant for Singapore but positive and significant for the Philippines. We compare the RE model with the TCE model using the likelihood ratio test, and report the results in Table 4. According to the likelihood ratio test, 2 times the difference between the log likelihoods for the less constrained (TCE) and more constrained (RE) model is distributed chi-squared with 2 degrees of freedom. 20 The RE model can be rejected in favor of the TCE model at the 1% level of significance for all three countries.
4. Policy implications The evidence from the TCE model provides a number of policy implications. Inflation in Malaysia, Philippines and Singapore is significantly affected by movements in their effective exchange rates. This is consistent with the view of Warner and Kreinin (1983) that effective exchange rates still vary even for developing countries with a basket-peg or managed-float exchange-rate regime. While effective exchange rate depreciation significantly affects inflation in the Philippines and to a lesser extent, Malaysia, Singapore's has the reverse effect. The initial downward pressure on the domestic price of imports caused by the appreciation of the Singapore dollar may be offset by the eventual increase in price due to the stronger demand for imports. Singapore's imports, which range from luxury goods to basic necessities such as food and water, make up a large portion of Singapore's consumption. In addition, rising real income in Singapore since the mid-1960s also reinforce stronger demand for imports. Nevertheless, the magnitude of the coefficient, which shows the impact of exchange rates on inflation in Singapore, is small (0.08). Even though Singapore imports most of what its people need, foreign prices have no effect on inflation. Singapore avoided 'imported' inflation by deliberately allowing its exchange rate to sufficiently appreciate so domestic inflation is not affected by increasing foreign prices. On the other hand, higher foreign price is shown to affect inflation in Malaysia. Excess foreign demand significantly affects inflation only in the Philippines. Higher foreign interest rates, while negatively influencing inflation in Malaysia and Singapore, positively affects inflation in the Philippines. The positive impact
20 The degrees of freedom are based on the difference between the number of estimated parameters in the TCE (20) and RE (18) models.
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435
of excess foreign demand and foreign interest rate on Philippine inflation may be due to the heavy debt burden of the Philippines. When foreign demand is strong, resources may be diverted to exports to pay off foreign debt thereby causing shortages and higher domestic prices in the Philippines. On the other hand, higher foreign interest rates and larger interest payments on Philippine debt would not necessarily lead to inflation unless the central bank monetizes part of the debt. The three countries' monetary policies can be deduced from the constrained parameters of the money supply equation in the RE model. 2~ For the Philippines, the coefficients that relate the money supply to prices (w 1), domestic income (w 2 ), and the exchange rate (ws), are all positive and significant, indicating a 'loose' or partly accommodative monetary policy. Fry (1986) traces the excess money supply to the expansion of credit for government bail-out of failing private-sector enterprises. The inflation arising from the monetizing of govemment debt caused a depreciation of the exchange rate, which in tum produced even more inflation because half of Philippine imports are fuel, chemicals and engineering products required in domestic production. In contrast, Malaysia and Singapore are not burdened by debt and both countries have maintained 'tight' or anti-inflationary monetary policies.
5. Conclusions
We examine effective exchange rate fluctuations and determinants of inflation in Malaysia, Philippines and Singapore in the context of a structural open economy macro model which incorporates cointegration between the exchange rate and its fundamentals and an endogenous money supply specification. Two versions of the model are estimated. The first version is a constrained model consistent with rational expectations (RE) and the second version is a semi-constrained model compatible with theories consistent expectations (TCE). The RE model is strongly rejected in favor of the TCE model. We find, as expected in a basket-peg exchange-rate regime, that exchange rate expectations are mainly affected by the current exchange rate. The effective exchange rate variation significantly affects inflation in the three southeast Asian countries, although the magnitude of the coefficients is small for Malaysia and Singapore. This finding is similar to Bahmani-Oskooee and Malixi (1992) for the Philippines and Rana and Dowling (1985) for Singapore but contrary to the result of Rana and Dowling (1985) for Malaysia. 22 Rana and Dowling find effective exchange rate to negatively affect inflation in Malaysia and Singapore, although
2~The TCE model does not provideestimates of the money supply equation. 22 Malaysia and Singaporeare excluded in the Bahmani-Oskooeeand Malixi (1992) but included in Rana and Dowling (1985).
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their estimates are insignificant for Singapore and marginally significant for Malaysia. Our results also show that foreign prices affect inflation in Malaysia while foreign income has a positive effect on inflation in the Philippines. Foreign interest rate has a negative effect on inflation in Singapore but a positive effect in the Philippines. The positive effect of foreign interest rate on inflation in the Philippines m a y be due to the monetization of debt. Evidence of excess money supply is deduced from the RE model. This finding for the Philippines is consistent with the conclusion of Otani (1975), that excess money supply is one of the determinants o f inflation in the Philippines. Overall, the TCE model with cointegration gives insight on effective exchange rate fluctuations and determinants of domestic inflation consistent with the experiences o f Malaysia, Philippines and Singapore. Unlike Malaysia and Singapore, which have had high growth and stable prices, the Philippines experienced slow growth and high inflation during the 1980s. W h i l e inflation in all three southeast Asian countries are shown to be affected by different external factors, Malaysia and Singapore have avoided high inflation despite high levels of economic growth through 'tight' or anti-inflationary monetary policy, whereas, 'loose' or partly accommodative monetary policy in the Philippines resulted in high inflation.
References Andrews, D., 1993. Test for parameter instability and structural change with unknown change point. Econometrica 61, 821-856. Balassa, B., Williamson, J., 1987. Adjusting to Success: Balance of Payments Policies in East Asian NICs, Institute of International Economics, Washington, DC. Bahmani-Oskooee, M., Malixi, M., 1992. Inflationary effects of changes in effective exchange rates: LDCs experience. Appl. Econ. 24, 465-471. Campbell, J., Perron, P., 1991. Pitfalls and opportunities: What macroeconomists should know about unit roots. In: Blanchard, O., Fisher, S., (Eds.), NBER Macroeconomics Annual, 1991, 141-201. Dornbusch, R., 1976. Expectations and exchange rate dynamics. J. Political Econ. 84, 1161-1176. Edwards, S., 1989. Real Exchange Rates, Devaluation, and Adjustment, MIT Press, Cambridge, MA. Frankel, J., 1979. A theory of floating exchange rates based on real interest differentials. Am. Econ. Rev. 69, 601-622. Fry, M., 1986. Financial structure, financial regulation and financial reform in the Philippines and Thailand, 1960-1984. In: Cheng, H.-S., (Ed.), Financial Policy and Reform in Pacific Basin Countries, D.C. Heath and Company, Washington, DC. Frydman, R., Phelps, E., 1990. Pluralism of Theories Problems in Post-Rational-ExpectationsModeling, presented at the 1990 Sienna Summer Workshop in Expectations and Learning. Goldberg, M., Frydman, R., 1993. Theories Consistent Expectations and Exchange Rate Dynamics. In: Frisch, H., Worgoter, A. (Eds.), Open Economy Macroeconomics, Macmillan, 377-399. Goldberg, M., Frydman, R., 1996a. Empirical exchange rate models and shifts in the co-integrating vector. Struct. Change Econ. Dynamics 7, 55-78. Goldberg, M.D., Frydman, R., 1996b. Imperfect knowledge and behavior in the foreign exchange market. Econ. J. 106, 869-893.
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