exchange rate volatility and united kingdom trade: evidence from canada, japan and new zealand

Empir Econ (2008) 35:607–619DOI 10.1007/s00181-008-0185-2

ORIGINAL PAPER

Exchange rate volatility and United Kingdom trade:evidence from Canada, Japan and New Zealand

Taufiq Choudhry

Received: 15 February 2005 / Accepted: 15 November 2006 / Published online: 26 March 2008© Springer-Verlag 2008

Abstract This paper investigates the influence of exchange rate volatility on the realimports of the United Kingdom from Canada, Japan and New Zealand during the period1980–2003. The Johansen multivariate cointegration method and the constrained errorcorrection (general-to-specific) method are applied to study the relationship betweenreal imports and its determinants (including exchange rate volatility). Conditionalvariance from the GARCH(1,1) model is applied as exchange rate volatility. Bothnominal and real exchange rates are employed in the empirical study. Results indicatea significant effect of the exchange rate volatility on real imports. These exchange ratevolatility effects are mostly positive.

Keywords Real exports · Volatility · GARCH · Conditional variance · Cointegration

1 Introduction

One of the major concerns since the introduction of the flexible exchange rate has beenwhether the increase in exchange rate volatility (risk) has affected the internationaltrade flow. Higher exchange rate volatility leads to higher cost for risk-averse tradersand to less foreign trade (Arize et al. 2000). In other words, greater exchange riskincreases the riskiness of trade profits, leading risk-averse traders to reduce trade.Thus, a theoretical framework seems to indicate a negative relationship between

The author thanks an anonymous referee, the editor and Myles Wallace for several useful comments andsuggestions. Any remaining errors and omissions are the author’s responsibility alone.

T. Choudhry (B)School of Management, University of Southampton,Highfield, Southampton SO17 1BJ, UKe-mail: [email protected]

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international trade flow and exchange rate volatility. Akhtar and Hilton (1984), DeGrauwe (1988), Chowdhury (1993), Arize (1995, 1998) and Arize et al. (2000) pro-vide evidence that exchange rate volatility does reduce international trade flow. Forsome developed countries, currencies forward markets (and futures markets) can beused to reduce or hedge exchange rate risk (volatility), but it has been proved thatforward markets fail to completely eliminate exchange rate risk (Akhtar and Hilton1984; Arize et al. 2000). Moreover, any cost of forward hedging will reduce inter-national trade: importers who pay for the forward hedge will face higher prices forthe foreign goods, and exporters who incur these hedging costs will pass then alongas higher prices. The end result, in both instances, is a reduction of trade. Accordingto Klaassen (2006), this perception that greater exchange rate risk reduces trade hashelped motivate monetary unification in Europe, and is also related to currency marketintervention by central banks.

Some studies also promote a potential positive effect of the exchange rate volatilityon international trade. De Grauwe (1988) indicates that the response of the producerto an increase in exchange rate volatility depends upon the risk-averseness of the pro-ducer. When producers are extreme risk-averse, an increase in exchange rate volatilitywill raise the expected marginal utility of export revenue, as producers will want toexport more to avoid a drastic decline in their revenue stream. The opposite is truewhen producers are less risk-averse. Also according to Doyle (2001), the positiveeffect of the volatility could be due to the relatively high multinational ownership,which offers significant opportunities for diversification of exchange rate risk throughinternational intra-subsidiary changes in production and trade. Further analysis of apositive effect is provided in Sercu and Uppal (2003) and Franke (1991). Positiveeffect of the volatility on trade is shown by Asseery and Peel (1991), McKenzie andBrooks (1997), Doyle (2001), and Choudhry (2005).

According to Arize (1998), knowledge of the degree to which exchange rate vola-tility affects trade is important for the design of both exchange rate and trade policies.For example, if exchange rate volatility leads to an increase in imports, trade adjust-ment programmes that discourage import expansion could be unsuccessful if exchangerates are volatile. In addition, the intended effect of a trade liberalisation policy maybe doomed by a variable exchange rate and could precipitate a balance-of-paymentscrisis (Arize 1998; Arize et al. 2000).

The purpose of this paper is to investigate the effects of the exchange rate vola-tility on exports of Canada, Japan and New Zealand to the United Kingdom duringthe period 1980–2003. McKenzie (1999) has reviewed this literature and empha-ses a few key points. First, the need for care in specifying the technique by whichexchange rate volatility is measured, with increased attention to the application ofthe GARCH and related models. Second, application of data disaggregated by sector,market and time period. Third, there is a need to apply proper methods necessary tocorrect prospective problems of serial correlation and nonstationarity in time seriesdata. Not many previous papers investigate the effect of exchange rate volatility onUnited Kingdom international trade. Doyle (2001) presents a study of the effect of theexchange rate volatility on the imports from Ireland to the United Kingdom during1979–1992.

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Exchange rate volatility and United Kingdom trade 609

2 The structure of the model

This paper employs a model similar to the one used in Chowdhury (1993), Arize(1995, 1998), and Arize et al. (2000). The following relationship is tested to check forthe effects of exchange rate volatility on the United Kingdom real imports.

ln(Xt ) = δ1 ln(Yt ) + δ2 ln(Pt ) + δ3 ln(Vt ) + εt (1)

where ln(Xt ) is the log of real United Kingdom imports from Canada or Japan orNew Zealand, Yt is the log of real income of the United Kingdom, Pt is a measure ofrelative imports prices of Canada or Japan or New Zealand to the United Kingdom,Vt is the exchange rate volatility in log, and ε is the error term. Based on the stan-dard theory, the real income of the importing country should have a positive effect onthe import level (Bailey et al. 1986, 1987). Thus the coefficient of real income (δ1)

should be positive. According to Arize (1995) and Arize et al. (2000), the coefficientof the price ratio (δ2) should be negative. The relative price is the ratio of the importprices of Canada or Japan or New Zealand to the United Kingdom. Changes in theprice ratio represent changes in the terms of trade, reflecting the impacts of changesin nominal exchange rates, differing rates of inflation among countries, and changesin relative prices in each country between its non-traded goods and its exports (Baileyet al. 1986, 1987). As indicated by Bailey et al. (1986) and Arize (1995), the influ-ence of the exchange rate volatility (Vt ) on trade is uncertain. Investigation of the sizeand direction of the impact imposed by the exchange rate volatility (Vt ) on the UnitedKingdom import is the main theme of this study. The long-run relationship representedby equation 1 is investigated by means of the Johansen multivariate cointegration tests,and the constrained error correction model is applied to check for causality betweenthe variables.

Over the years, several different definitions of exchange rate volatility have beenapplied in the empirical tests. For example, the variance or standard deviation ofthe exchange rate and the vector autoregressive (VAR) methods are two of the com-mon means of measuring exchange rate volatility (see Bailey et al. (1986, 1987),Chowdhury (1993), and Arize et al. (2000)). But, according to Jansen (1989) and Korayand Lastrapes (1989), these methods have serious drawbacks. Given these drawbacksArize (1995) and McKenzie (1999), indicate that the exchange rate volatility may bemodelled by the Autoregressive Conditional Heteroscedastic (ARCH) model of Engle(1982) and the Generalised Autoregressive Conditional Heteroscedastic (GARCH)model of Bollerslev (1986).

In this paper, conditional variance of the first difference of the log of the exchangerate is applied as volatility. The conditional variance is estimated by means of theGARCH model. Asseery and Peel (1991) claim that from an optimising perspectiveuse, of the conditional variance from the GARCH model to proxy exchange rate vol-atility seems “economically relevant”. Kroner and Lastrapes (1993), Caporate andDoroodian (1994), Lee (1999) and Choudhry (2005) also apply the GARCH model’sestimated volatility of exchange rate in their study of international trade. These studiesapplying the GARCH model find varying effect in size and direction of the exchangerate volatility on trade.

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One of the main debates in this field is the use of the nominal or the real exchangerate. As stated by Bailey et al. (1986, 1987) and Akhtar and Hilton (1984), persuasivearguments can be made for the application of both exchange rates. This paper employsboth the nominal exchange rate and the real exchange rate in the empirical estimation.1

3 Data and empirical tests

The data from Canada, Japan and New Zealand applied are quarterly, ranging fromfirst quarter 1980 to first quarter 2003.2 The length of the data applied is based onavailability. The price indices are the import price indices for all four countries. Realincome of the United Kingdom is represented by the real personal income. The nomi-nal exchange rate applied is defined as foreign currency per United Kingdom sterling.Real exchange rate is created using these price indices and the nominal exchangerates.3 The data are obtained from DATASTREAM.

As required by cointegration tests, first the stochastic structure (presence of unitroots) of each series is determined. There are several different types of unit root testsin the literature. This paper applies the augmented Dickey–Fuller test (Dickey andPantula 1987). These tests show that all series have one unit root, thus they are station-ary after first difference but are non-stationary in levels. Since these results are quitestandard, they are not provided but are available request.

Two or more nonstationary time series are cointegrated if a linear combination ofthese is stationary. Cointegration tests in this paper are conducted by means of themethod developed by Johansen and Juselius (1990).4 The Johansen method appliesthe maximum likelihood procedure to determine the presence of cointegrating vectorsin nonstationary time series. This method detects the number of cointegrating vec-tors, and allows for tests of hypotheses regarding elements of the cointegrating vector.Detailed description and analysis of the Johansen method are provided in many articlesand books, and thus are not given here.

Table 1 present the cointegration results. For all three countries, two relationshipsare tested for possible cointegration. The first and second tests are conducted withthe nominal exchange rate volatility and the real exchange rate volatility in the VAR,

1 The GARCH results indicate volatility clustering in all six exchange rate series. These results are availableon request.2 All three countries are involved in trade with the United Kingdom on a large scale. United Kingdom isCanada’s third largest export market. During 2001, total bilateral trade between the two countries amountedto 16.6 billion Canadian dollars. Five billion Canadian dollars worth of goods was exported to the UnitedKingdom by Canada. Total bilateral trade between the United Kingdom and Japan during 2000 was about2.2 trillion yen. Total Japanese exports to the United Kingdom during 2000 were about 1.7 trillion yen.Similarly, United Kingdom is New Zealand’s fourth largest export market. Total bilateral trade during 1996was over 1 billion pounds, and total exports to the United Kingdom were equal to 632 million pounds.3 The real exchange rate is defined as log of (ex)*(PUK/PF), where ex is the nominal exchange rate betweenUnited Kingdom pound and other currencies, PUK is the price index of the United Kingdom, and PF is theprice index of Canada or Japan or New Zealand.4 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 more robust thanthe other methods even when the errors are non-normal (Gonzalo 1994).

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Table 1 Cointegration results

Vectors r = 0 r ≤ 1 r ≤ 2 r ≤ 3

Nominal exchange rate volatility in the VAR

Canada

Max eigenvalue test 32.03 16.71 2.90 0.67

Trace Test 52.31a 20.28 3.57 0.67

Eigenvalues 0.3022 0.1712 0.0321 0.0075

Lags = 4, Trace correlation = 0.483, Autocorrelation LM(4), χ2(16) = 20.016, Normality χ2(8) = 82.127∗

Japan

Max eigenvalue test 52.15a 16.90 9.51 1.24

Trace test 79.80a 27.65 10.75 1.24

Eigenvalues 0.4434 0.1730 0.1013 0.0138


New Zealand


Trace test 58.04a 28.01 8.45 0.19

Eigenvalues 0.2864 0.1973 0.0887 0.0022


Real exchange rate volatility in the VAR

Canada

Max eigenvalue test 43.34a 14.56 10.30 0.68

Trace test 68.88a 25.54 10.98 0.68

Eigenvalues 0.3923 0.1541 0.1116 0.0078


Japan


Trace test 38.82 21.61 8.10 2.30

Eigenvalues 0.1723 0.1380 0.0617 0.0249


New Zealand


Trace test 46.05 22.82 8.25 0.40

Eigenvalues 0.2253 0.1479 0.0827 0.0044


a,b Significance at the 1% and 5% level, respectively* Rejection of the null at the 5% level

respectively. Both tests involving Canada show only one significant cointegrating vec-tor at the 5% or above level. Thus results indicate a long-run equilibrium relationshipbetween the Canadian real imports to the United Kingdom, United Kingdom realincome, the price ratio between the two countries, and exchange rate volatility (bothnominal and real). Four and six lags are applied in the VAR in the first and second

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tests, respectively. The diagnostic tests fail to show any significant serial correlation.Results indicate the presence of non-normal residuals but, as indicated by Gonzalo(1994) Johansen method still provides the most robust results than the other methodseven when the errors are non-normal.

Results using the Japanese and New Zealand data are also shown in Table 1. Bothcountries present similar results. Using the nominal exchange rate volatility in the caseof both countries, one significant vector is indicated at the 5% level or above. Thus,results show a significant long-run equilibrium relationship between real imports fromJapan and New Zealand to the United Kingdom and its determinants (including thenominal exchange rate volatility). In both tests, 4 lags are applied in the VAR. Nosignificant serial correlation is indicated, but the residuals are non-normal again inboth tests.

Using the real exchange rate volatility, both Japan and New Zealand fail to indicateany significant vector(s). These results fail to show any significant long-run rela-tionship between real imports (from Japan and New Zealand) and its determinants,when real exchange rate volatility is included. The diagnostic statistics are againsatisfactory.

4 Normalized equations and long-run elasticities

The estimated cointegrating vectors are given economic meaning by means of nor-malizing on the real imports. The normalized equations are obtained by dividing eachcointegrating vector by the negative of the cointegrating vector on real imports. Theresulting coefficients represent the long-run elasticities. The significance of the vari-ables is tested by means of the likelihood ratio test (chi-square).

Table 2 presents the normalized equations from all tests. No normalized equationsare provided for Japan and New Zealand when real exchange rate volatility is includedin the VAR. This is because no significant vector(s) were found in these tests. Thereal income consistently imposes an expected significant positive effect on the real

Table 2 Normalized equations and long-run elasticities

Country Log of import Log of income Log of price ratio Log of volatility

Nominal exchange rate volatility in the VAR

Canada 1.00a(13.21) 2.005a(9.19) 2.414a(10.00) 2.913a(28.88)

Japan 1.00b(4.37) 2.530a(4.88) 3.657a(8.43) 1.394a(37.98)

New Zealand 1.00a(5.93) 0.047(0.09) 1.234a(10.33) −0.023(0.00)

Real exchange rate volatility in the VAR

Canada 1.00a(6.89) 0.986a(7.12) 1.947b(4.59) 2.229a(23.50)

Japan _ _ _ _

New Zealand _ _ _ _

a,b Significance at the 1 and 5% level, respectivelyData in the parentheses are the χ2 statistics (log likelihood ratio)

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imports in all tests, except in the case of New Zealand where it is insignificant.5 Theincome coefficient is relatively large in most tests, indicating a large elastic effect onthe United Kingdom imports. Choudhry (2005), Bahmani-Oskooee and Kara (2005),Arize et al. (2000), and Arize (1995, 1998) also show highly elastic income effect ontrade. For example, in the first Canadian result (with the nominal exchange rate vola-tility in the VAR), a 1% increase in United Kingdom income increases the Canadianexports to the United Kingdom by 2%.

The price ratio coefficient is positive and significant in all four tests. These coef-ficients are also large, and thus the effect is highly elastic. In the Japanese test, 1%jump in the prices ratio increases the export from Japan to the United Kingdom by3.7%. Finding a positive effect of the price ratio is not unusual. Arize et al. (2000),Bahmani-Oskooee and Kara (2005) and Choudhry (2005) also provide some evidenceof the positive highly elastic effect of the price ratio. According to Arize et al. (2000)the positive effect of the price ratios may-be due to three reasons: first is the use ofunit-value indices and they are only accurate if the composition of the unit remainsthe same; second some countries may be operating limit pricing to deter entry whilecompeting in other areas; and third, it can be the result of poor quality data.

The exchange rate volatility imposes a significant effect in three of the four tests.In the case of New Zealand, nominal exchange rate volatility is insignificant. In thethree significant cases, exchange rate volatility imposes a large size positive (direct)effect on the real imports of the United Kingdom. For example, in the first Canadiantest, a 1% increase in volatility increases the United Kingdom import from Canadaby 2.9%. Results presented in this paper fail to show a negative (inverse) effect ofthe exchange rate volatility on trade. Similar to the findings of some other empiricalstudies our results are consistent with the theoretical arguments that increased volatil-ity can increase trade. Figure 1 presents the log of actual value of Canadian export tothe United Kingdom and the fitted value of the export based on the normalized coef-ficients from the nominal exchange rate volatility cointegration test. The two linesseem to move together up and down and relatively closely. The volatile nature of thefitted line may be due to the volatility of the nominal exchange rate.6 Other significantcointegration tests provide similar graphs of actual and fitted values of exports. Thesegraphs are available on request.

5 Test of causality between real export and its determinants

Cointegration also implies that the transitory components of the series can be given adynamic error correction representation, i.e. a constrained error correction model canbe applied that captures the short-run dynamic adjustment of cointegration variables.The constrained error correction model allows for a causal linkage between two or

5 The New Zealand cointegration relationship was also tested without the insignificant variables, incomeand nominal exchange rate volatility in the VAR. The result indicated a lack of cointegration between theremaining two variables. This implies the need for the income and exchange rate volatility in the VAR. Thisresult is available on request.6 This overshooting and undershooting of the fitted values was expected due to the inclusion of the exchangerate volatility. But the long run movement of the two series is similar and together implying a decent coin-tegrating relationship.

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ACTUAL FITTED

1980 1982 1984 1986 1988 1990 1992 1994 1996 1998 2000 20021.2

1.4

1.6

1.8

2.0

2.2

2.4

2.6

2.8

Fig. 1 Actual and fitted values of Canadian export

more variables stemming from a common trend or equilibrium relationship. As longas two or more variables are cointegrated, causality must exist in at least one direc-tion. The methodology applied in this paper follows the Hendry (1987) “general-to-specific” paradigm. In the present context, the following representation is implied:

�Zt = C + A(L)�Zt + θ1 ECt−1 + εt (2)

where:C = vector of constant termsA(L) = matrix of finite order of lag polynomialsZ = log (X, Y, P, V )θ1 = vector of coefficientsECt−1 = ln(X)t−1 − δ∗

1 ln Yt−1 − δ∗2 ln Pt−1 − δ∗

3 ln Vt−1where δ∗

1 , δ∗2 , and δ∗

3 are the estimates of δ1, δ2 and δ3Within a constrained error correction model, causality may arise from two sources.

The second term on the right-hand side of Eq. 2 represents the short-term dynamicinteraction between real exports and their determinants, and the conventional testsof causality may be based on the significance of these terms. The disequilibriumadjustment of each variable towards its long-run equilibrium value is then captured bythe error correction term, ECt−1, with the coefficient of this term in each individualequation depending on the speed of adjustment of the variable towards its long-runequilibrium value; the coefficient (θ1) represents the speed of adjustment towards thelong-run equilibrium.7 If this coefficient is insignificant, then the dependent variable

7 With the cointegrating vector normalized on real exports, in which real exports the dependent variableis, the associated element of θ1 represents the speed of adjustment directly. In the remaining equations,the corresponding elements of θ1 represent the ratio of speed of adjustment of the relevant variables to thevalue of its associated coefficient in the cointegration relationship.

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Table 3 Error correction test results, Canada: nominal exchange rate volatility

Lags Constant EC �Imports �Income �PR �EXV Diagnostics

0 −0.0428 _ _ _ _(−0.910)

1 −0.0134c _ _ 0.7662a _ R2 = 0.175(−1.830) (2.718)

2 _ _ _ 0.0707b DW = 2.08(2.199)

3 _ _ 0.9888a 0.0804a JB = 4.725(3.743) (2.673)

4 _ _ _ _ LM χ2(6) = 6.831

5 _ _ _ _ ARCH χ2(1) = 0.001

6 _ _ _ _ ARCH χ2(3) = 2.475

7 _ _ _ _ RESET test F(1, 82) = 0.189

8 _ _ _ _ Chow test F(6, 81) = 0.929

a,b,c Significance at the 1, 5 and 10% level, respectivelyDW Durbin–Watson, JB Jarque–Bera, LM serial correlation test, Chow Test structural stability test, Reset TestRamsey’s specification error statistics, ARCH autoregressive conditional heteroscedasticity of the residuals

does not adjust to correct departures from equilibrium. In these models, a variableis econometrically exogenous only if the lagged changes in the dependent variablesprovide explanatory power.

Initially, in zero to eight lags of the first difference of each variable in Eq. 2,a constant term (C) and one lagged error correction term (EC) generated from theJohansen method are applied.8 Then, as required by the general-to-specific method,the dimensions of the parameter space are reduced to final parsimonious specifica-tions by eliminating insignificant coefficients or imposing statistically insignificantcoefficients. Under the general-to-specific approach, diagnostic tests of the statisticaladequacy of the model come first, with an examination of inferences for the theorydrawn from the model left until after a statistically adequate model has been found(Brooks 2002).

Tables 3, 4, 5 and 6 present the error correction results. In order to save space,only results with the real import (first difference) as the dependent variable are pro-vided. Remaining results are available on request. Tables 3 and 4 present the Canadianresults with the nominal exchange rate volatility and real exchange rate volatility,respectively. In both tests, the one-lagged error term (ECt−1) is significant, and hasthe proper negative sign. Significance of the error term implies that overlooking thecointegration relationship between the variables would have introduced misspecifica-tion in the underlying dynamic structure (Arize et al. 2000). The significance of theerror term implies causality from all four independent variables to the real imports inthe long run. In Table 3, the size of the coefficient on the lagged error term (−0.0134)

8 In the case of the dependent variable (change in the real export), one to eight lags are applied in theright-hand side of the equation.

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Table 4 Error correction test results, Canada: real exchange rate volatility


0 0.3638b _ _ _ _(2.229)

1 −0.0451b _ _ _ _ R2 = 0.122(−2.185)

2 _ _ _ _ DW = 2.13

3 _ _ 0.6507a 0.1662a JB = 1.513(2.300) (2.815)

4 _ _ _ _ LM χ2(6) = 4.225

5 _ _ 0.2878b _ ARCH χ2(1) = 0.001(2.074)

6 _ _ _ _ ARCH χ2(3) = 0.238

7 _ _ _ _ RESET test F(1, 81) = 2.818

8 _ _ _ _ Chow test F(5, 83) = 0.410

a,b,c Significance at the 1, 5 and 10% level, respectivelyDW Durbin–Watson, JB Jarque–Bera, LM Serial correlation test, Chow Test structural stability test,Reset Test Ramsey’s specification error statistics, ARCH autoregressive conditional heteroscedasticity of theresiduals

Table 5 Error correction test results, Japan: nominal exchange rate volatility


0 −0.031 _ _ _ 0.0407b

(−0.500) (2.167)

1 −0.0035b _ _ −0.4371a _ R2 = 0.124(−2.513) (−2.653)

2 _ _ _ _ DW = 1.85

3 _ _ −0.3225c _ JB = 2.833(−1.912)

4 _ _ _ _ LM χ2(6) = 4.885

5 _ _ _ _ ARCH χ2(1) = 0.003

6 0.1772c _ _ _ ARCH χ2(3) = 1.721(1.8608)

7 _ _ _ _ RESET test F(1, 82) = 0.372

8 _ _ _ _ Chow test F(6, 81) = 0.659

a,b,c Significance at the 1, 5 and 10% level, respectivelyDW Durbin–Watson, JB Jarque–Bera, LM serial correlation test, Chow Test structural stability test, Reset TestRamsey’s specification error statistics, ARCH autoregressive conditional heteroscedasticity of the residuals

indicates that 1.34% of the adjustment of real imports towards the long-run equilib-rium takes place per quarter. According to the dynamics of the equations, relativeprices and exchange rate volatility have significant short-run effects on real imports,in addition to the long-run effects. The short-run effect of the nominal exchange rate

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Table 6 Error correction test results, New Zealand: nominal exchange rate volatility


0 0.4364a _ _ _ 0.0407b

(3.654) (2.167)

1 −0.4409a −0.4366a _ _ _ R2 = 0.455(−3.758) (−3.417)

2 −0.2856b _ 1.1860b _ DW = 2.08(−2.284) (2.502)

3 −0.1942c _ _ _ JB = 0.647(−1.954)

4 _ _ 0.9952b _ LM χ2(6) = 4.369(1.998)

5 _ _ _ −0.7181c ARCH χ2(1) = 0.00001(−1.985)

6 _ _ _ _ ARCH χ2(3) = 0.152

7 _ _ _ _ RESET test F(1, 78) = 2.405

8 _ _ _ _ Chow test F(7, 76) = 0.904

a,b,c Significance at the 1, 5and 10% level, respectivelyDW Durbin–Watson, JB Jarque–Bera, LM Serial correlation test, Chow Test structural stability test,Reset Test Ramsey’s specification error statistics, ARCH autoregressive conditional heteroscedasticity of theresiduals

volatility is positive (using lags 3 and 4), but small in size (0.07 and 0.08). In Table 4,using the real exchange rate volatility, the speed of adjustment of real import towardsthe long-run equilibrium is 4.51% per quarter. This adjustment rate is relatively fasterthan the one using the nominal exchange rate volatility. Using the real exchange ratevolatility (Table 4) once again, no significant short-run effect is found of the realincome and real imports. There is evidence of short-run effect imposed by changesin exchange rate volatility and price ratios. Results from both tests also indicate thatreal imports of the United Kingdom from Canada are not econometrically exogenous.The range of the coefficient of correlation (R2) is between 0.12 and 0.18. In addition,in both tests, diagnostic statistics for serial correlation, abnormal residuals, structureinstability, heteroscedasticity and linearity assumption are satisfactory.

Tables 5 and 6 present the Japanese and the New Zealand results using nominalexchange rate volatility, respectively. Once again, both the error terms are negativeand significant. In the case of Japan, the speed of adjustment (−0.0035) is quite slow,at 0.35% per quarter, while in the case of New Zealand, the speed of adjustment(−0.4409) is quite high, at 44% per quarter. Results from the Japan show a short-runeffect on real imports is imposed by lagged real imports, price ratio, and exchangerate volatility. The nominal exchange rate volatility imposes a significant positive, butsmall size, short-run effect on real imports. In the case of New Zealand, also, short-runeffects are imposed by lags of real imports, price ratio, and exchange rate volatility.For New Zealand, there is some evidence of inverse short-run effect of exchange ratevolatility on real imports. Again, results from both tests indicate that real imports of

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the United Kingdom from Japan and New Zealand are not econometrically exoge-nous. The R2 is 0.124 in the case of Japan, and 0.455 for New Zealand. Once again,in both tests, all diagnostic statistics are satisfactory. No error correction tests usingreal exchange rate volatility are done for Japan and New Zealand. This is due to thelack of cointegration between the variables when real exchange rate volatility, usingthe Japanese and New Zealand currencies against the United Kingdom pound, is inthe VAR.

6 Conclusion and implications

One of the major concerns since the introduction of the flexible exchange rate has beenwhether the increase in exchange rate volatility (risk) has affected the internationaltrade flow. This paper investigates the effects of the exchange rate volatility on theUnited Kingdom imports from Canada, Japan and New Zealand during 1980–2003.Conditional variance of the first difference of the log of the exchange rate estimatedfrom a univariate GARCH(1,1) is applied as volatility. The paper then applies the mul-tivariate cointegration method and constrained error correction (general to specific)models to study the relationship between real imports and their determinants, that is,real income, export price ratio and exchange rate volatility. Both the volatility of thenominal and the real exchanges rates between the United Kingdom pound and thecurrencies of Canada, Japan and New Zealand are employed.

Results obtained indicate a stationary long-run equilibrium relationship betweenreal imports and their determinants for all three countries, using the nominal exchangerates’ volatility. Using the real exchange rate volatility, only the Canadian tests resultsindicate a stationary long-run equilibrium relationship. Normalized equations indicatethat in the majority of the relationships, exchange rate volatility imposes a positiveeffect on real exports. This result may imply that exchange rate variability measured byvolatility enhances international trade flows from Canada, Japan and New Zealand tothe United Kingdom. Similar to the findings of some other empirical studies our resultsare consistent with the theoretical arguments that increased volatility can increasetrade. Further analysis is conducted by means of Hendry (1987) general-to-specificerror correction (causality) tests. Error corrections also show that causality does existfrom the exchange rate volatility to real imports. This is true, using both the nominaland the real exchange rate volatility. Also, causality from volatility to real importsis also found in the short-run. In most tests, the short-run effect of volatility on realimports is also positive.

The results presented suggest that exchange rate volatility considerations are impor-tant for modelling United Kingdom import behaviour from Canada, Japan and NewZealand. Any trade adjustment programmes by the United Kingdom that discourageimport expansion could be unsuccessful if exchange rates are volatile. If policy mak-ers ignore the stability of the nominal and real exchange rates between the UnitedKingdom pound and Canada/Japan/New Zealand currencies, policy actions aimedat stabilising these import markets are likely to generate uncertain results. Resultspresented in this paper advocate further research in this field, using data from othercountries.

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