Charalambos Pattichis
email: pattichis15@yahoo.com

The intertemporal budget constraint and current account sustainability in Cyprus: Evidence and policy implications

Keywords: Current account sustainability, Cyprus Intertemporal budget constraint, current account sustainability, Cyprus

The intertemporal budget constraint and current account sustainability in Cyprus: Evidence and policy implications

Introduction

The Republic of Cyprus, a full member of the European Union since May 2004, has decided to join the European Economic and Monetary Union (EMU) and adopt the euro as its currency. By giving up its own currency (the Cyprus pound) the country automatically loses the ability to devalue in order to tackle any possible future current account deficits. It is, therefore, important to ask whether Cyprus' current account balance is sustainable. If it is, then giving up its exchange rate policy may not be a serious cost of adopting the euro since the sustainability of the current account balance implies that this policy may not be required, ceteris paribus. However, if Cyprus' current account deficit is not sustainable, then the loss of exchange rate policy can impose serious economic costs on the country arising from the fact that Cyprus will need to use contractionary fiscal policy (which can harm economic growth) to tackle the growing external imbalances. The interested reader should consult De [START_REF] De Grauwe | The Economics of Monetary Union, 5 th Edition[END_REF] for a comprehensive analysis of the costs and benefits of adopting a common currency. Note that within a currency union a country, by definition, loses its monetary policy and, thus, has to rely on fiscal policy for macroeconomic management.

Historically, Cyprus has not been using its exchange rate policy for competitive devaluations to improve the current account. Instead, the aim of exchange rate policy in Cyprus has been the maintenance of price and general macroeconomic stability through the linkage of the Cyprus pound exchange rate to some currency anchor (see [START_REF] Kyriacou | The current framework of exchange rate policy in Cyprus and recent developments[END_REF] for a detailed description of exchange rate policy in Cyprus). An import-weighted currency basket served as the currency anchor from 1973 to 1984, when it was replaced with a trade-weighted currency basket until 1992. From 1992 onwards, the Cyprus pound at a central parity rate of CYP1=1.7086ECU with fluctuation bands of ± 2.25%. These bands were increased to ± 15% in January 2001 to enable the Central Bank of Cyprus to absorb any destabilizing capital movements that could arise from the process of capital account liberalization taking place in Cyprus [START_REF] Kyriacou | The current framework of exchange rate policy in Cyprus and recent developments[END_REF]). In May 2005, the Cyprus pound joined the European Exchange Rate Mechanism II (ERM II) at the abovementioned central parity rate and fluctuation bands.

Figure 1 presents the time series of exports of goods and services of Cyprus and its imports of goods and services (plus net transfer payments and net interest payments).

Both trade flows are in real terms. Figure 2 presents their difference (exports minus imports), i.e. the current account balance. Although real exports and imports seem to move together over time, since about 1999 the current account deficit has been increasing. The question, therefore, under investigation in this paper is whether Cyprus' current account deficit is sustainable.

To investigate this question, this paper follows the approach developed by [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] which tests for the existence of a long-run relationship between imports and exports. The rationale of such an approach is that if such a long-run relationship exists, the time series of imports and exports will never move too far apart from each other over time, implying that the current account balance is sustainable in the long-run. The next section presents the theoretical model and the testable hypotheses. Section 3 discusses the empirical results and the final section summarizes the conclusions and policy implications.

Theoretical underpinnings and empirical formulation

The approach of this paper follows closely [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF]. 1 The approach tests for the existence of a long-run relationship between imports and exports. The rationale of such an approach is straightforward. If such a long-run relationship exists, the time series of imports and exports will tend to move closely together over time, i.e. they will not permanently diverge from one another, implying that the current account balance is sustainable. The theoretical model that lends support to the existence of such a long-run relationship is presented in [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] and is briefly summarized below. [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] considers the case of a small open economy without a government sector that produces a single composite good. The representative consumer who lives in this economy is able to borrow and lend in international financial markets (at a given world interest rate) using one-period financial instruments. The agent's resources consist of endowments of output and redistributed profits and he uses them for consumption and savings to maximize intertemporal utility subject to these constraints. The current period budget constraint of this individual is given by:

C 0 = Y 0 + B 0 -I 0 -(1+r 0 ) B -1 (1)
where C 0 , Y 0 , and I 0 are current period consumption, output, and investment, respectively;

r 0 is the one-period world interest rate; B 0 is international borrowing, which could be (1992) showed that the IBC can be expressed as:

0 1 lim t t n n n t B TA B µ µ = = + , (2)
where TA t = X t -M t (= Y t -C t -I t ) represents the trade balance in period t (i.e. income minus absorption), X t equals exports, M t equals imports, 0 =1/(1+r 0 ), and µ t is the discount factor defined as the product of the first t values of . If the last term in equation ( 2) is equal to zero the amount of borrowing (lending) in international markets is equal to the present value of all future trade surpluses (deficits). As [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] indicated, if this is not the case and B 0 is positive then the country is 'bubble-financing' its external debt, implying that the government will need to issue new debt. It is, therefore, critical to investigate whether the last term in equation ( 2) is equal to zero [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] and [START_REF] Hakkio | Is the budget deficit too large?[END_REF]). To derive a testable model, [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] makes a number of assumptions (including that the world interest rate is stationary and that imports and exports are nonstationary (I(1)) series) and derives the following model: the null hypothesis is that they are cointegrated and b=1. As [START_REF] Hakkio | Is the budget deficit too large?[END_REF] and [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] argued, the condition that exports and imports are cointegrated is necessary for the economy to satisfy its intertemporal budget constraint. However, the condition that b=1 is not. However, if b is less than one, the economy will fail to satisfy its budget constraint if trade flows are expressed relative to output since this would be inconsistent with a finite external debt-to-output ratio, and hence with the sustainability of the current account balance [START_REF] Hakkio | Is the budget deficit too large?[END_REF], [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF]).

X t = a + b M t + u t (3)
In the context of the government's IBC, [START_REF] Quintos | Sustainability of the deficit process with structural shifts[END_REF] has introduced 'strong' and 'weak' conditions for the sustainability of the budget deficit. Under the 'weak' condition she argued that cointegration between government revenues and expenditures is not a necessary but a sufficient condition for deficit sustainability, arguing that the necessary condition is that the debt grows more slowly than the borrowing rate. Following Quintos's work, [START_REF] Martin | US deficit sustainability: A new approach based on multiple endogenous breaks[END_REF] specifies that the government deficit is: (i) 'strongly' sustainable if and only if the I(1) processes of government revenues and expenditures are cointegrated and b=1; (ii) only 'weakly' sustainable if government revenues and 2 Although equation ( 3) is specified as in [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] the empirical analysis in section 3 suggests that X t should be treated as a 'long run forcing' variable explaining M t . Thus, the specification in section 3 becomes M t = a + b X t + u t and b refers to the coefficient of X t . This does not affect the two hypotheses that form the basis of the empirical analysis in this paper, which are: (a) Are M t and X t cointegrated, and (2) is b=1? For a similar approach see for example, [START_REF] Arize | Imports and exports in 50 countries: Tests of cointegration and structural breaks[END_REF] and [START_REF] Irandoust | Are imports and exports cointegrated? An international comparison[END_REF]. Assuming, realistically, that these results carry over to the case of current account sustainability, the present paper focuses on a test of 'strong' sustainability of Cyprus' current account.

In this paper, following [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] (see also, amongst others, [START_REF] Baharumshah | On the sustainability of current account deficits: evidence from four ASEAN countries[END_REF]), X t represents exports of goods and services, and M t measures imports of goods and services plus net transfer payments and net interest payments. As [START_REF] Hakkio | Is the budget deficit too large?[END_REF] and [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] have noted there are a number of ways to measure trade flows: in nominal terms, in real terms, and as a proportion of nominal or real Gross Domestic Product (GDP). [START_REF] Hakkio | Is the budget deficit too large?[END_REF] further argued that since in the theoretical model the interest rate is assumed stationary, nominal interest rates are ruled out since they are usually non-stationary. Thus, following [START_REF] Hakkio | Is the budget deficit too large?[END_REF] this paper focuses on real trade flows. Furthermore, since b being less than one implies that the economy will fail to satisfy its budget constraint if trade flows are expressed relative to output, the paper also analyzes the case of real trade flows as a proportion of real GDP [START_REF] Hakkio | Is the budget deficit too large?[END_REF]).

Several studies have empirically employed the model, see for example, [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF],

Bahmani-Oskooee and Rhee (1997), [START_REF] Fountas | Are the US current account deficits really sustainable[END_REF], [START_REF] Apergis | Current account sustainability: the case of Greece[END_REF], [START_REF] Wu | Are current account deficits sustainable? Evidence from panel cointegration[END_REF], [START_REF] Arize | Imports and exports in 50 countries: Tests of cointegration and structural breaks[END_REF], [START_REF] Baharumshah | On the sustainability of current account deficits: evidence from four ASEAN countries[END_REF], and [START_REF] Irandoust | Are imports and exports cointegrated? An international comparison[END_REF]. To the best of our knowledge, the only study that has analyzed the case of Cyprus is [START_REF] Arize | Imports and exports in 50 countries: Tests of cointegration and structural breaks[END_REF]. However, [START_REF] Arize | Imports and exports in 50 countries: Tests of cointegration and structural breaks[END_REF]

Empirical results

Variable specification and data sources

All data used in this paper were taken from the International Financial Statistics IFS code 64). As in the case of exports, no data exist for a price index of imports. We have used the CPI as a deflator since it also includes the prices of imported goods. It should be noted that the CPI and GDP deflator series are very similar to one another.

Thus, in deriving measures of real imports and real exports, it makes no real difference if one deflates the nominal trade figures by the CPI or the GDP deflator.

Current Account balance (RCA): This is defined as real exports minus real imports.

Real exports as a proportion of real GDP (LRXGDP):

These are real exports (as defined above) expressed as a proportion of real GDP. The latter is nominal GDP (IFS code 99b) deflated by the GDP deflator.

Real imports as a proportion of real GDP (LRMGDP):

These are real imports (as defined above) expressed as a proportion of real GDP.

Unit root analysis

All variables used in the empirical analysis have been tested for unit roots using the Augmented Dickey-Fuller (ADF) test (see Dickey andFuller (1979, 1981)) and the DF-GLS test proposed by [START_REF] Elliott | Efficient tests for an autoregressive unit root[END_REF].

The ADF test results (with and without a trend included in the model) are presented in Table 1. The appropriate lag length of the autoregression was chosen based on the Akaike Information Criterion using a maximum lag order of two which was dictated by the small The DF-GLS results are presented in Table 2. The order of augmentation of the DF-GLS tests was based on two information criteria: the Modified Akaike Information Criterion (MAIC) (see [START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF]) and the Schwarz Information Criterion (SIC) using a maximum lag order of two. In those cases where the MAIC and the SIC selected different lag lengths with different conclusions about the time series properties of the variables, the paper reports results for both. In all such cases the SIC selected lag 1 and the MAIC selected lag 2. Table 2 also reports two critical values: the 5% critical values based on the response surface analysis of [START_REF] Cheung | Lag order and critical values of a modified Dickey-Fuller test[END_REF] and the 5% interpolated critical values based on [START_REF] Elliott | Efficient tests for an autoregressive unit root[END_REF].

[TAKE IN TABLES 1 AND 2 HERE]

The ADF test results reported in Table 1 suggest that the current account balance (RCA) is stationary. However, the DF-GLS test results are not so clear. The latter suggest that if inference is based on the model with lag 1 (selected by SIC) then RCA is indeed an I(0) variable regardless of whether or not a trend is included in the model. However, if inference is based on the model with lag 2 (selected by MAIC) then RCA appears to be an I(1) variable. STATA 10 also reports another information criterion: the Ng and Perron (1995) sequential-t method. Based on this information criterion (not reported in Table 2) the analysis suggests that RCA is an I(0) variable. Thus, overall the results suggest that 3 The econometric analysis in this paper was conducted using MICROFIT 4.0 [START_REF] Pesaran | Microfit 4.0: Interactive Econometric Analysis[END_REF] LRM appear to be I(1) variables at the 5% level under both the trend and no-trend cases.

Thus, we proceed to the cointegration analysis based on this conclusion.

The long-run relationship between LRM and LRX

The long-run relationship between LRM and LRX (both of which have been found to be I(1) variables based on the DF-GLS test) is investigated using the 'bounds' test. The 'bounds' test for the existence of a long-run level relationship between a dependent variable and a set of regressors advanced by [START_REF] Pesaran | Bounds testing approaches to the analysis of level relationships[END_REF] (see also [START_REF] Pesaran | Microfit 4.0: Interactive Econometric Analysis[END_REF]) is based on the F-statistic and is used to test the joint significance of the lagged levels of the included variables in an unrestricted error correction model (UECM). [START_REF] Pesaran | Bounds testing approaches to the analysis of level relationships[END_REF] show that the asymptotic 13 distribution of this F-statistic is non-standard, under the null hypothesis that there exists no long-run relationship between the variables, irrespective of whether the underlying regressors are I(0) or I(1).

In the case of two variables (LRM and LRX) the UECM takes the following form:

t t t i t n i i i t n i i t u LRX LRM LRX LRM LRM + + + + + = = = 1 1 0 1 0 (4)
where LRM and LRX are the first differences of the logarithms of LRM and LRX respectively, and u t is an error term. As [START_REF] Hendry | Dynamic specification[END_REF] They also provide two sets of asymptotic critical values (lower and upper critical bounds; hence the name for the test) for the two extreme cases: one that assumes that all regressors are I(0) and the other that all regressors are I(1). If the computed F-statistic exceeds the upper critical bound then the hypothesis of no long-run relationship is rejected, and such inference is made without knowing the order of integration of the regressors. In this way, the pre-testing problems and uncertainty involved in unit root tests are avoided. However, if the computed F-statistic is lower than the lower critical bound then the null hypothesis cannot be rejected. Finally, if the computed statistic falls 14 inside the critical bounds, inference is inconclusive and one needs to establish the order of integration of the underlying regressors before conclusive inference can be made.

The 'bounds' test is based on the assumption that the residuals in equation ( 4) are serially uncorrelated. It is, therefore, important that the lag order n be selected appropriately. It is also important to decide the appropriate specification of the deterministic components to be included in the model. To decide on the lag order, models were estimated for n=1,2,3 and the choice of the appropriate model was based on the AIC and SIC criteria. Both information criteria suggest that the appropriate model is n=1. The results of this model are presented in Table 3. It is clear that the model passes all the usual diagnostic tests including the cumulative sum (CUSUM) and cumulative sum of squares of recursive residuals (CUSUMSQ) tests for structural stability (not reported but available on request) (see [START_REF] Brown | Techniques for testing the constancy of regression relations over time[END_REF]). An inspection of the residuals suggests that there are no outliers. Only in one case the residual was slightly greater than two times the standard error of the regression and no residual was greater than three times the standard error of the regression. Thus, no dummy variables have been used.

[TAKE IN TABLE 3 HERE]

Equation (4) was also estimated with a time trend included in the model. The AIC and SIC indicated that, in this case as well, the appropriate lag order is n=1. The information criteria suggested that the model without a trend is slightly better. Since the results of the two models are very similar, the model with a deterministic trend is not reported here.

The model passes all the usual diagnostic tests (see Notes to Table 3 for details on these tests). The results suggest that the time trend is highly statistically insignificant. 4.

[TAKE IN TABLE 4 HERE]

Table 4 reports two sets of critical values: (a) those reported in [START_REF] Pesaran | Bounds testing approaches to the analysis of level relationships[END_REF] which relate to large samples; and (b) those reported in [START_REF] Narayan | The saving and investment nexus for China: evidence from cointegration tests[END_REF] which relate to small samples. It is clear from Table 4 that in all cases, the computed F statistic is greater than the upper critical bounds, suggesting that the null hypothesis cannot be accepted. Thus, the 'bounds' test results clearly suggest the existence of a long-run level relationship between LRM and LRX.

Unrestricted error correction models of the form given in equation (4) (with and without a deterministic trend) were also estimated with LRX as the dependent variable (results not reported but available on request). The 'bounds' tests in this case suggested the absence of a long-run relationship between the variables since the estimated F-statistics fell below the lower critical values (implying that cointegration is only present when LRM is the dependent variable [START_REF] Narayan | Estimating income and price elasticities of imports for Fiji in a cointegration framework[END_REF]). This suggests that LRX is a 'long-run forcing' variable explaining LRM [START_REF] Pesaran | Microfit 4.0: Interactive Econometric Analysis[END_REF]). Thus, in the remaining of the paper we will treat LRM as the dependent variable with long-run exports assumed given. This is not unrealistic for a small (resource-scarce) economy such as Cyprus, since an increase in exports will require higher imports of, for example, raw materials. For a similar approach see for example, [START_REF] Arize | Imports and exports in 50 countries: Tests of cointegration and structural breaks[END_REF] and [START_REF] Irandoust | Are imports and exports cointegrated? An international comparison[END_REF].

Having established the existence of a long-run relationship, the paper proceeds to estimate the long-run coefficients using the ARDL approach advocated by Pesaran and Shin (1999) (see also [START_REF] Pesaran | Microfit 4.0: Interactive Econometric Analysis[END_REF]). Pesaran and Shin (1999) have shown that after appropriate augmentation of the order of the ARDL model, the ARDL-based estimators of the long-run coefficients are super-consistent and valid inferences can be made using standard normal asymptotic theory. They have also shown that appropriate choice of the order of the ARDL model tackles any endogenous regressor problems and that their approach is applicable irrespective of whether the regressors are I(0), I(1) or mutually cointegrated.

In the case of two variables in levels (LRM and LRX) the ARDL model takes the following form:

t i t p i i i t p i i t LRX LRM LRM + + + = = = 0 1 (5)
where t is an error term. The appropriate model is selected on the basis of information criteria and the long-run coefficient of LRX (b) is derived as:

1 i i i i b = .
Using a maximum lag order of p=2, the AIC selected an ARDL (2,2) model. Thus, the coefficient of LRX is statistically significant and almost equal to 1 in magnitude. 4 A Wald test of the hypothesis that b=1 is accepted by the data ( 2(1)=0.245

[p-value=0.620]). These results suggest that Cyprus is not in violation of its intertemporal budget constraint. 5

Furthermore, the results suggest that the coefficient of the lagged error correction term (-1.04; t-ratio: -4.92) in the error correction representation of the model is correctly signed and statistically significant, thus confirming the conclusion that a long run relationship exists between the variables (see [START_REF] Engle | Co-integration and error correction: Representation, estimation and testing[END_REF]). Its magnitude suggests high 4 It should be noted that an almost identical long run model is derived if the order of the ARDL model is selected on the basis of the SIC. Furthermore, an identical long-run coefficient of LRX is derived using the results of the UECM presented in Table 3. Based on equation ( 4) the long-run coefficient in this case is calculated as: -( / ). This is not surprising since, as argued before, the UECM is a simple reparameterisation of an ARDL model. 5 The Johansen procedure has also been used as a robustness check of the results. The Johansen results confirm that LRX and LRM are cointegrated and the resulting cointegrating vector is very similar to the one reported in the main body of the article.

Furthermore, the hypothesis that b=1 was accepted by the data.

The long-run relationship between LRMGDP and LRXGDP

The unit root test results presented in Tables 1 and2 suggest that both LRXGDP and LRMGDP are I(0) variables. The stationarity of these variables implies that the 'spurious' regression problem identified by [START_REF] Granger | Spurious regressions in econometrics[END_REF] is not applicable here. However, it should be noted that in the theoretical framework used to derive the testable model, [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] assumes that exports and imports are I(1) series, and a country satisfies its IBC if these nonstationary series are cointegrated. Since both our variables are I(0), the concept of cointegration is not applicable and the relevance of the theoretical model questionable. Thus, the approach here is to simply estimate the long-run relationship and test whether b is statistically significant and equal to one (and ensure that the residuals of our model have the desirable properties). Therefore, the analysis in this section should be treated with caution.

To derive the long-run relationship between the variables, we estimated a simple static regression by ordinary least squares of the form:

LRMGDP t = a + b LRXGDP t + u t
where u t is an error term. The estimated equation is given as (t-ratios in brackets):

LRMGDP t = 0.016 + 0.931 LRXGDP t (0.1) (4.07) (A trend was also included in the model but since it was statistically insignificant it was subsequently dropped from the specification.) The static model passes all the usual diagnostic tests at the 5% level, except the test for Normality (results not reported but stability at the 5% level. An ADF unit root test of the residuals of the static equation confirms that they are stationary (estimated ADF(of lag 1) statistic=-4.7; critical value: -3.58). A Wald test of the hypothesis that b=1 was accepted by the data ( 2 (1)=0.09 [p-value=0.763]). These results further suggest that Cyprus' current account balance is sustainable.

To further check the robustness of the results, we have estimated an ARDL(2,2) model for LRMGDP (lag order selected by the AIC using a maximum lag order of two). This model passes all the usual diagnostic tests (including the CUSUM and CUSUMSQ tests for structural stability) and the results are presented in Table 5.

[TAKE IN TABLE 5 HERE]

The long-run relationship derived from the ARDL model is given as (t-ratios in brackets): LRMGDP t = 0.165 + 1.138 LRXGDP t (0.9) (4.27)

A Wald test of the hypothesis that b=1 was accepted by the data ( 2 (1)=0.27 [p-value=0.604]). Thus, these results further support the conclusions of the previous models.

These conclusions and their implications are summarized in the next section.

Conclusions and policy implications

Cyprus has decided to join the European monetary union and adopt the euro as its currency. One of the costs of joining a monetary union is the loss of exchange rate policy (the ability to devalue in order to tackle current account deficits). Whether the loss of this policy is a serious cost of joining the Eurozone depends to a large extent on whether the current account balance of Cyprus is sustainable. If this is the case, and given that Cyprus has not been following a policy of competitive devaluations to improve its current account over the period under consideration, it implies that the loss of the exchange rate policy may not be a serious cost of adopting the Euro since the country may not require the use of this policy, ceteris paribus.

To investigate this question, this paper used the approach developed by [START_REF] Husted | The emerging US current account deficit in the 1980s: a cointegration analysis[END_REF] which tests for the existence of a long-run relationship between imports and exports. If The results presented in this paper suggest that the current account balance is stationary and that a valid long-run relationship exists between real imports and real exports of goods and services. These results point to the conclusion that Cyprus is not in violation of its intertemporal budget constraint and the current account balance is sustainable in the A policy implication of these findings is that the loss of exchange rate policy following the adoption of the Euro may not be a serious cost for Cyprus. This conclusion derives from the fact that even though Cyprus has not been using competitive devaluation policies over the period under investigation, the current account balance was sustainable.

Thus, ceteris paribus, Cyprus may not require the use of this policy and, therefore, its loss may not constitute a serious cost of adopting the Euro.

However, it may be argued that the sustainability of the current account was the result of proper macroeconomic policies that ensured that any short-run imbalances were corrected and the economy was guided towards the long-run equilibrium. As we have already mentioned, Cyprus has not been following a policy of competitive devaluations over the period under investigation. However, this does not mean that the exchange rate policy followed by Cyprus had no positive effect on the current account. The policy of pegging the Cyprus pound exchange rate to some currency anchor has helped in keeping inflation low and stable and this, in turn, had a positive impact on the international competitiveness of the economy. Furthermore, the finding of current account sustainability may have been the result of proper fiscal and monetary policies. By becoming a member of EMU, Cyprus loses its monetary policy and exchange rate policy and constraints are imposed on its fiscal policy. Thus, whether Cyprus will continue to satisfy its IBC remains to be seen.

The main point that can be made at this stage is that if the finding had been that the current account was not sustainable, then the loss of exchange rate policy would have Godfrey's (1978a[START_REF] Godfrey | Testing for higher order serial correlation in regression equations when the regressors include lagged dependent variables[END_REF] Lagrange multiplier test of (no) residual serial correlation (see also [START_REF] Harvey | The econometric analysis of time series[END_REF];

Functional Form is [START_REF] Ramsey | Tests for specification errors in classical linear least squares regression analysis[END_REF] RESET test for (correct) functional form specification; Normality is the Bera and Jarque (1981) test for normality;

Heteroscedasticity is [START_REF] White | A heteroskedasticity-consistent covariance matrix estimator and a direct test for heteroskedasticity[END_REF] test for homoscedasticity; and ARCH is [START_REF] Engle | Autoregressive conditional heteroskedasticity with estimates of the variance of UK inflation[END_REF] test for (no) autoregressive conditional heteroscedasticity (see [START_REF] Pesaran | Microfit 4.0: Interactive Econometric Analysis[END_REF] for more details). Notes (Table 4): The specification of deterministic components refers to Cases III, IV and V in [START_REF] Pesaran | Bounds testing approaches to the analysis of level relationships[END_REF]. The critical values reported by [START_REF] Pesaran | Bounds testing approaches to the analysis of level relationships[END_REF] are for large samples. [START_REF] Narayan | The saving and investment nexus for China: evidence from cointegration tests[END_REF] has computed critical values for the 'bounds' test for small samples using the same GAUSS code used by [START_REF] Pesaran | Bounds testing approaches to the analysis of level relationships[END_REF]. The reported [START_REF] Narayan | The saving and investment nexus for China: evidence from cointegration tests[END_REF] critical values are based on a sample of 30 observations. The F-statistic for Case III refers to testing that = =0 in equation ( 4). The F-statistics for Cases IV and V are based on a similar equation like (4) but with a trend included in the specification.

The F-statistic for Case IV refers to testing that = =0 and that the coefficient of the trend is also equal to zero. The F-statistic for Case V refers to testing that = =0 (see [START_REF] Pesaran | Bounds testing approaches to the analysis of level relationships[END_REF]). 1976 1978 1980 1982 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2004 1976 1978 1980 1982 1984 1986 1988 1990 1992 1994 1996 1998 2000 2002 2004 2004 Page 32 of 32 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

  was unilaterally pegged to the European Currency Unit (ECU) (and later the Euro)

  and 0<b<1; and, (iii) the deficit is unsustainable if 0 b .

9

  (IFS) CD-ROM of the International Monetary Fund. The study uses annual data from 1976-2004 since quarterly data were not available in IFS for most of the sample period. (Since Cyprus was divided in 1974 following the Turkish invasion, our sample starts in 1976.The focus is the government-controlled (Greek) part of Cyprus.) As indicated in the previous section our export series represents exports of goods and services, and our imports series measures imports of goods and services plus net transfer payments and net interest payments. These series were constructed using data from the Balance of Payments section of the IFS. All dollar values were transformed into Cyprus pounds using the bilateral exchange rate (IFS code rh). All variables (except the current account balance) were transformed into natural logarithms prior to the empirical analysis. The definition of the variables used in the empirical analysis is given below.Real exports (LRX):These are exports of goods and services deflated by the GDP deflator (base year=2000; IFS code 99bip). No data are available for an export price index. The use of the GDP deflator is guided by the fact that it represents domestically produced goods (some of which are exported).Real imports (LRM): These are imports of goods and services plus net transfer payments and net interest payments deflated by the Consumer Price Index (CPI) (base year=2000;

  sample size. (The use of the Schwarz Information Criterion (SIC) yielded similar conclusions about the time series properties of the variables.) As in the case of the ADF DF-GLS test was also done with and without a trend included in the model.3

  noted this UECM is a simple reparameterisation of a general autoregressive distributed lag (ARDL) model. The null hypothesis of no long-run level relationship (i.e. no cointegration) is tested by estimating the UECM and testing the joint significance of the lagged levels in equation (4), i.e. H 0 : = =0 . Pesaran et al (2001) have tabulated the appropriate critical values for different number of regressors and various specifications of the deterministic components (intercept and trend and whether these should be restricted or unrestricted).

  ' test for the existence of a long run relationship was conducted based on these two estimated models, using three different specifications for the deterministic components. These three specifications were: (a) unrestricted intercept and no trend (Case III in Pesaran et al (2001)); (b) unrestricted intercept and restricted trend (Case IV in Pesaran et al (2001)); and, (c) unrestricted intercept and unrestricted trend (Case V in Pesaran et al (2001)). The results are presented in Table

  to equilibrium, which takes about one year. In addition, this error correction model passes all the usual diagnostic tests.

  such a long-run relationship exists, the time series of imports and exports will never move too far apart from each other over time, implying that the current account balance is sustainable in the long-run. This paper has studied the sustainability of Cyprus' current account in three ways: (a) By testing for stationarity of the current account balance; (b) by testing for the existence of a long-run relationship between real imports and real exports; and (c) by investigating the long-run relationship between real imports and real exports, both expressed as a proportion of real GDP.

Figure

  Figure 1: Real Exports (RX) and Real Imports (RM)

Figure

  Figure 2: The Current Account Balance (RCA)

  is an error term. 2 If b equals one and u t is stationary then the economy satisfies its intertemporal budget constraint. In other words, if M t and X t are non-stationary, then

	where u t
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  ) except the Phillips-Perron test and the DF-GLS test which were both To investigate this further we have also tested LRM and LRX for unit root behavior using the Phillips-Perron test and the DF-GLS test. The Phillips-Perron tests (see[START_REF] Phillips | Testing for a unit root in time series regression[END_REF]) have failed to significantly clarify the situation (results not presented but available on request). The Phillips-Perron Z(t) statistic confirms that LRX is an I(1) variable. This conclusion is robust to different truncation lag orders for the Newey-West adjustment of the variances and regardless of whether a trend is included or not in the model. However, as in the case of the ADF test, the Phillips-Perron Z(t) results suggest that (at the 5% significance level) LRM is I(1) if no trend is included in the model but I(0) if a trend is included. However, at the 1% level the unit root null for LRM cannot be rejected even in the case when a trend is included in the model.

	Submitted Manuscript Submitted Manuscript
	RCA is a stationary variable. This result carries the important implication that Cyprus'
	current account deficit is sustainable.
	Turning now to variables LRXGDP and LRMGDP, the ADF test results suggest that
	these variables are I(0) processes. This conclusion is also supported by the DF-GLS test
	results which clearly suggest that LRXGDP is an I(0) variable. The results also suggest
	that LRMGDP is an I(0) variable if the model with lag 1 is used (selected by SIC)
	regardless of whether or not a trend is included in the model. However, in the No Trend
	F o F o
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	over the period 1978-2004 (i.e. one more observation) the ADF t statistic is -3.68. This
	substantial reduction in the magnitude of the statistic may suggest considerable sample
	done using STATA 10 (StataCorp (2007)). I would like to thank Mona Kanaan for her dependence. In other words, with a bigger sample, the finding might have been that LRM
	assistance with STATA. is I(1).
	11 12

Page 10 of 32 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK case the MAIC selects lag 2 and based on this the analysis just fails to reject the unit root null if the

[START_REF] Cheung | Lag order and critical values of a modified Dickey-Fuller test[END_REF]

critical values are used. The unit root null, however, is rejected if the

[START_REF] Elliott | Efficient tests for an autoregressive unit root[END_REF]

critical values are used. The conclusion that LRMGDP is an I(0) variable is also supported if inference is based on the lag length chosen by the

[START_REF] Ng | Unit root tests in ARMA models with data-dependent methods for the selection of the truncation lag[END_REF]

sequential-t method. Thus, we proceed on the conclusion that both LRXGDP and LRMGDP are I(0) variables.

Based on the ADF unit root results, LRX is clearly an I(1) variable, whereas LRM is I(1) if no trend is included in the ADF regression but I(0) if such a trend is included. The latter result is somewhat surprising given that the plots of LRM and LRX indicate similar behavior over time. In fact, the ADF t statistic of -4.42 is based on the estimation period 1979-2004 which is used by the econometric software package following our specification of a maximum lag order of two. If the same ADF regression is estimated Page 11 of 32 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Turning now to the results based on the DF-GLS test the picture becomes much clearer. It is clear from Table 2 that regardless of which critical values one uses both LRX and

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  The model (results not reported but available on request) passes all the usual diagnostic tests at the 5% level (see Notes of Table3for information on these tests). The long-run relationship

	between LRM and LRX derived from the ARDL (2,2) model is given as (t-ratios in
	brackets):
	LRM t = 0.159 + 0.988 LRX t
	(0.89) (41.97)
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  The model also marginally fails the CUSUM test for structural

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	available on request).
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  Furthermore, the results suggest that the coefficient of exports is equal to one, implying that the current account balance is 'strongly' sustainable.

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	long-run.
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  cost of adopting the euro. However, the results of this paper suggest that Cyprus is not in violation of its IBC and its current account balance is sustainable in the long-run. Thus, it may be argued that the loss of a policy instrument in such a case may not be a serious cost of joining EMU. On a methodological level, the results presented in this paper support theoretical models that use an intertemporal approach to the current account.

	Submitted Manuscript
	F o r TABLES AND FIGURES Table 1: Augmented Dickey-Fuller unit root tests Variable L t µ t RCA -3.51 (1) -4.37 (1) been a serious F o r F o r F o r F o r F o r LRX -2.13 (0) -2.48 (0) LRM -1.45 (0) -4.42 (1) LRXGDP -4.07 (2) -3.85 (2) LRMGDP -4.83 (1) -4.72 (1) F o r	t µ -5.01 (2) -5 (0) -4.68 (2) -4.73 (0) -4.75 (2)	L	t -4.95 (2) -5.27 (0) -4.82 (2) -4.79 (0) -4.64 (2)
	P P P P P P P
	e e r e e r e e r e e r e e r e e r e e r
	R e v i e w R e v i e w R e v i e w R e v i e w R e v i e w R e v i e w R e v i e w
	22 27 28

Notes (Table

1

): L and L denote the level and first difference of the series, respectively. t µ is the ADF t-ratio with a constant but no trend and t is the ADF t-ratio with a constant and a trend included in the ADF regression. The order of augmentation of the ADF tests (given in brackets) was selected by the Akaike Information Criterion using a maximum order of two. The use of a maximum order of three yields similar conclusions. The 5% critical values are -2.98 for the t µ test and -3.6 for the t test.

Table 2 :

 2 [START_REF] Elliott | Efficient tests for an autoregressive unit root[END_REF] DF-GLS test statistics

	Model	RCA	LRX	LRM	LRXGDP LRMGDP CL-5% ERS-5%
	Trend	-4.22 (1)	-2.66 (1) -3.13 (1)	-4.11 (1)	-4.17 (1)	-3.37	-3.19
		-2.84 (2)					-3.29
	No Trend -2.78 (1)	-0.19 (1)	0.09 (1)	-4.13 (1)	-3.11 (1)	-2.44	-1.95
		-1.43 (2)				-2.32 (2)	-2.39
	Notes (

Table 2 )

 2 : The column entitled Model indicates whether or not a trend is included in the model. The order of augmentation of the DF-GLS tests (given in brackets) was based on two information criteria reported by STATA 10: the Modified Akaike Information Criterion (MAIC) (see[START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF]) and the Schwarz Information Criterion (SIC) using a maximum lag order of two. In those cases where the MAIC and the SIC selected different lag lengths with different conclusions about the time series properties of the variables, the table reports the results for both. In all such cases the SIC selected lag 1 and the MAIC selected lag 2. The table reports two critical values given by STATA 10: CL-5% reports the 5% critical values based on the response surface analysis of[START_REF] Cheung | Lag order and critical values of a modified Dickey-Fuller test[END_REF] and ERS-5% reports the 5% interpolated critical values based on[START_REF] Elliott | Efficient tests for an autoregressive unit root[END_REF]. The CL-5% column reports two critical values: the first number applies to models with lag 1 and the second number to models with lag 2. The ERS-5% critical values reported by STATA are the same for both models.

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Table 3 :

 3 Results of the unrestricted error correction model for LRM

	Dependent variable is LRM
	OLS estimation from 1978-2004
	Regressor	Coefficient	t-statistic
	Constant	0.165	0.87
	LRM (-1)	0.402	2.5
	LRX	1.046	7.19
	LRX (-1)	-0.485	-2.05
	LRM (-1)	-1.039	-4.92
	LRX (-1)	1.027	4.89
	DIAGNOSTIC TESTS
	Adjusted R 2	0.79
	F-statistic	F(5, 21)=[0.000]
	SER	0.052
	Durbin's h-statistic	[0.972]
	Serial correlation	F(1, 20)=[0.748]
		F(2, 19)=[0.718]
	Functional Form	F(1, 20)=[0.554]
	Normality	2 (2)=[0.282]
	Heteroscedasticity	F(1, 25)=[0.331]
	ARCH	F(1, 20)=[0.738]
		F(2, 19)=[0.921]
	Notes (

Table 3 )

 3 : Values in square brackets are p-values. In the regressor column the number in brackets next to the variable name indicates the lag length. SER is the standard error of the regression. Serial Correlation is the F version of

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Table 4 :

 4 'Bounds' test results for the existence of a long-run relationship

	Deterministic	F statistic	5% Critical Values	5% Critical Values
	components		Pesaran et al (2001)	Narayan (2005)
	Unrestricted intercept	12.11	4.94-5.73	5.395-6.35
	and no trend (Case III)
	Unrestricted intercept	7.71	4.68-5.15	5.377-5.963
	and restricted trend
	(Case IV)
	Unrestricted intercept	11.54	6.56-7.3	7.36-8.265
	and unrestricted trend
	(Case V)

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Table 5 :

 5 Results of the autoregressive distributed lag model for LRMGDP

	Dependent variable is LRMGDP
	OLS estimation from 1978-2004
	Regressor	Coefficient	t-statistic
	Constant	0.169	0.9
	LRMGDP (-1)	0.37	1.93
	LRMGDP (-2)	-0.397	-2.57
	LRXGDP	1.154	5.32
	LRXGDP (-1)	-0.558	-1.7
	LRXGDP (-2)	0.573	2.12
	DIAGNOSTIC TESTS
	Adjusted R 2	0.6
	F-statistic	F(5, 21)=[0.000]
	SER	0.052
	Durbin-Watson statistic	2.02
	Serial correlation	F(1, 20)=[0.757]
		F(2, 19)=[0.906]
	Functional Form	F(1, 20)=[0.791]
	Normality	2 (2)=[0.271]
	Heteroscedasticity	F(1, 25)=[0.169]
	ARCH	F(1, 20)=[0.598]
		F(2, 19)=[0.829]
	Notes (

Table 5 )

 5 : See Notes to Table3.

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Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

ACKNOWLEDGEMENTS

I would like to thank an anonymous referee of this journal for useful and constructive comments on an earlier draft. I am, of course, solely responsible for any remaining errors.

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript


Faizul ; Haque

Thankom Gopinath

Colin Kirkpatrick

Heriot-Watt

Arun

Corporate governance and capital structure in developing countries: a case study of Bangladesh

Keywords: Corporate governance index, agency theory, debt finance, banks, Bangladesh

published or not. The documents may come

Introduction

The concept of corporate governance appears to be closely related with the financing pattern1 of a firm. While many studies investigate the influence of corporate governance on financial performance, the empirical relationship between corporate governance and the firm's capital structure has largely been unexplored. Several studies (for example, [START_REF] Du | Ultimate corporate ownership structures and capital structures: evidence from East Asian economies[END_REF][START_REF] Kumar | Corporate governance mechanisms and firm financing in India[END_REF]) have analysed the relationship between corporate governance and debt finance, but most studies consider individual governance issues, such as ownership structure, rather than overall firm-level governance practices. More importantly, there has not been any study in the context of Bangladesh2 . Even though debt financing is considered as an important corporate governance mechanism in mitigating the agency problems between shareholders and managers [START_REF] Harris | The theory of capital structure[END_REF], it is important to analyse the pattern of relationship between debt finance and the agency costs incurred between controlling and minority shareholders. In the context of this connection, the paper examines whether firm-level corporate governance has an influence on the firm's capital structure pattern in general, and debt financing in particular3 .

The Ordinary Least Square (OLS) regression framework uses a questionnaire-based Corporate Governance Index (CGI) to investigate the effect of corporate governance quality on capital structure of 98 non-financial listed firms in Bangladesh. The paper is organised as follows: section 2 reviews available literature and section 3 presents the research question and empirical model. Section 4 provides the empirical analysis (including the data), the univariate analysis and the regression results. Section 5 analyses and interprets the study results and section 6 concludes the paper.

Literature Review

The seminal works [START_REF] Fama | The theory of finance[END_REF] and [START_REF] Jensen | Theory of the firm: managerial behaviour, agency costs and ownership structure[END_REF] are widely credited with put forwarding the idea of agency theory based explanation of capital structure. Agency theory suggests that conflict of interests causes agency costs, which in turn determine the firm's capital structure decisions [START_REF] Harris | The theory of capital structure[END_REF]. While [START_REF] Jensen | Theory of the firm: managerial behaviour, agency costs and ownership structure[END_REF] refer to two types of agency conflict in a firm (between the shareholders and managers and the shareholders and creditors), [START_REF] Shleifer | A survey of corporate governance[END_REF] suggest that a conflict of interests can occur between large controlling block holders and minority shareholders. It is argued that large investors can cause enormous agency problems through direct or indirect expropriation of minority shareholders as well as employee rights.

Corporate debt policy is commonly regarded as an important corporate governance mechanism in mitigating the agency conflicts between shareholders and managers. Debt finance can resolve agency problems through reducing free cash flow and increasing the probability of bankruptcy risks and job losses4 . Large shareholders can also mitigate agency problems, since they have the incentive to collect information and monitor management [START_REF] Jensen | Theory of the firm: managerial behaviour, agency costs and ownership structure[END_REF][START_REF] Shleifer | A survey of corporate governance[END_REF].

According to another line of explanation within agency theory, improved corporate governance and associated strong shareholder rights will reduce agency costs and improve the confidence of investors in a firm's future cash flow [START_REF] Gompers | Corporate governance and equity prices[END_REF], and this in turn reduces the cost of equity capital to the firm [START_REF] Drobetz | Corporate governance and expected stock returns: evidence from Germany[END_REF]). This 3 eventually enhances the firm's ability to gain access to equity finance5 , leading to a decrease in the firm's dependence on (or preference for) debt finance. Alternatively, controlling owners of poorly governed firms are likely to prefer debt in order to meet financing needs, whilst retaining absolute ownership and control over the firm.

Apart from corporate governance, several other firm-specific factors tend to be important determinants of the firm's financing pattern. Whilst Fama andJensen (1983, cited in Suto 2003) argue that a large the ability of a firm to disclose greater amounts of information will cause lower agency costs-of-debt (e.g. positive effect on debt finance), lower information asymmetry between the insiders and outside shareholders of a large firm might have a negative effect on leverage due to the lower agency cost of equity financing [START_REF] Rajan | What Do we know about capital structure? some evidence from international data[END_REF]. [START_REF] Suto | Capital structure and investment behaviour of Malaysian firms in the 1990s: a study of corporate governance before the crisis[END_REF] argues that firm size is more likely to exert a positive influence on debt finance in developing countries as well as in bank-based financial systems, since default information is less readily available in these economies. [START_REF] Kumar | Corporate governance mechanisms and firm financing in India[END_REF] and [START_REF] Du | Ultimate corporate ownership structures and capital structures: evidence from East Asian economies[END_REF] also suggest that the relationship between firm size and leverage tends to be positive, mainly due to the advantage economies of scale have in their ability to issue long-term debt, their stronger negotiating power with the lenders and the opportunity for diversification and associated lower possibility of default risk.

According to the pecking order theory of corporate finance6 , profitability tends to be inversely related with debt finance. This is because perceived conflict of interest between the insiders and outside providers of funds drives a firm to favour retained earnings over external finance [START_REF] Suto | Capital structure and investment behaviour of Malaysian firms in the 1990s: a study of corporate governance before the crisis[END_REF]. Profitability can, however, be positively associated with the debt finance, since creditors are likely to lend more to the firms with higher cash flows [START_REF] Rajan | What Do we know about capital structure? some evidence from international data[END_REF]. [START_REF] Myers | Determinants of corporate borrowing[END_REF] and [START_REF] Rozeff | Growth, beta, and agency costs as determinants of dividend payout ratios[END_REF] [START_REF] Du | Ultimate corporate ownership structures and capital structures: evidence from East Asian economies[END_REF]. Nonetheless, the reluctance of controlling shareholders to forgo future corporate earnings, together with an insufficient cash flow to support the increased cost of raising equity funds, might drive growing firms to rely on debt rather than equity. Moreover, [START_REF] Suto | Capital structure and investment behaviour of Malaysian firms in the 1990s: a study of corporate governance before the crisis[END_REF] and [START_REF] Du | Ultimate corporate ownership structures and capital structures: evidence from East Asian economies[END_REF] suggest that nondebt tax deductions (e.g. depreciation and investment tax credits) are important substitutes for tax shield benefits of debt finance. They also suggest that asset tangibility (e.g. fixed assets) serves as important collateral in the debt contract and reduces the default risks of the lender, leading to higher debt finance.

Hypothesis and Model

This paper uses the agency theory based explanation of the capital structure of a firm. It investigates whether firm-level corporate governance has an influence on corporate debt finance in a developing economy like Bangladesh. More explicitly, the paper addresses the following hypothesis: H 0 : Corporate governance quality is inversely associated with the debt finance

The study follows, among others, [START_REF] Jiraporn | Capital structure, shareholder rights, and corporate governance[END_REF], [START_REF] Singh | The impact of corporate debt on long term investment and firm performance[END_REF] and [START_REF] Kumar | Corporate governance mechanisms and firm financing in India[END_REF] in using two alternative measures of capital structure, such as the ratio of total debt to total assets (denoted by debt-to-assets) and long term debt ratio (e.g. the ratio of long term debt to long term debt plus shareholder equity). Taking these debt financing measures as dependent variables, the following OSL regression model is developed:

Capital Structure ( ) =
+ 1 (CGI) + 2 (Concentration) + 3 (Firm size) + 4 (Profitability) + 5 (Growth) + 6 (Tangibility) + 7 (Non-debt tax shield) + 8 (Industry dummies) + ………………… (i) This cross-sectional model incorporates a firm-level Corporate Governance Index (denoted as CGI) as the main test variable, with higher CGI specifying better governance Based on the review of literature on the determinants of capital structure, several firm characteristics are used as control variables. These include firm size (measured as the natural logarithm of assets), profitability or return on assets (the ratio of net income after taxes to total assets), growth (a three-or five-year average asset growth rate), asset tangibility (fixed assets-to-total assets), non-debt tax shield (the ratio of depreciation and amortisation to total assets) and the industry dummies based on 4-digit SIC codes.

One important caveat of the study is that it is unable to address the issues of endogeneity and reverse causality 7 , primarily because of the absence of time variation in governance and financial data, along with the problem of finding appropriate instrumental variable for the simultaneous regression approach. This remains a shortcoming in the empirical study. Nevertheless, This problem can be mitigated by incorporating several firm specific characteristics as control variables, including the industry dummies 8 .

Empirical Analysis

This section explains the data including the CGI, followed by a summary statistics and univariate analysis, and the regression results. 7 Black et al. (2005) explain two different perspectives of firm level endogeneity: reverse causality and signaling effect. Reverse causality refers to the notion that firms with better financial performance prefer to adopt best practices of governance rather than vice-versa, whereas signaling effect explains that firms adopt better governance practices to signal high quality. 8 Among others, [START_REF] Klapper | Corporate governance, investor protection, and performance in emerging markets[END_REF] and [START_REF] Drobetz | Corporate governance and expected stock returns: evidence from Germany[END_REF] argue that adding appropriate control variables can be one way to mitigate the problem of omitted variables, endogeneity or reverse causality.

The Data

Corporate Governance Index

In order to quantify the firm-specific governance quality, a Corporate Governance Index (CGI) is constructed, consisting of five individual governance components 10 ; namely, the ownership pattern (sub-index 1), shareholders' rights (sub-index 2), independence and responsibilities of the board and management (sub-index 3), financial reporting and disclosures (sub-index 4) and responsibility towards the stakeholders (sub-index 5). The method follows several studies (such as Black et al. 2003;Klapper and Love 2002) to construct a CGI, although many of the governance elements have been modified in order to make the index compatible with the legal and regulatory issues in Bangladesh 11 . ***Insert Table 1 about here*** The distribution of CGI of 101 non-financial firms in Bangladesh (presented in table 1) reveals that the mean (median) value of the CGI is 40.84 (41.67), and the standard 9 The financial sector appears to dominate the market behavior of the DSE, with nearly 52 per cent (including banking firms with 47 per cent) of the total market capitalisation, and 53 per cent of the total turnover (DSE Review Dec. 2004). 10 A sub-index is constructed by summing up the values of all variables (between 1 and 0, with 1 being the compliance with better governance, and 0 otherwise) within that index, which is divided by the number of 'non-missing' variables. The ratio is then multiplied by 20 to have the sub-index value between 0 and 20. The overall CGI is finally constructed by adding up the values of all five sub-indices, and it carries a value between 0 and 100, with higher index scores denoting the better governed firms. 11 Note that the firm-specific scoring of the corporate governance practices in Bangladesh might not be comparable to international governance ratings. Given the persistent inefficiency in the legal and enforcement structures, the study is intended to measure the relative voluntary activism and/or legal compliance of the firm in corporate governance matters. 2 shows that all correlation coefficients amongst the CGI and its five sub-indices are positive, and all are statistically significant. Even though many individual elements of the sub-indices have been eliminated because of their perceived substitution or overlapping responses, the correlation coefficients among some sub-indices still remain reasonably high. This is likely to be because firms with better governance quality become visible by demonstrating their superiority in almost all categories that constitute the sub-indices.

Moreover, the adoption of better practices in all governance areas appears to be considered as complementary by some foreign and local reputed firms.

Summary Statistics and Univariate Analysis

This sub-section describes the capital structure pattern of the non-financial firms in Bangladesh, alongside the univariate relationships between corporate governance variables and several alternative measures of debt finance. financial firms in Bangladesh tend to have high degree of reliance on short-term bank debt rather than equity finance or long-term debt.

***Insert

Following [START_REF] Gompers | Corporate governance and equity prices[END_REF], governance scores are used to construct two extreme portfolios such as 'repressive portfolio' (i.e. firms with poor governance quality, with CGI < 35) and 'moderate portfolio' (i.e. better governed firms, with CGI > 48). Both portfolios represent roughly the upper (33 firms) and lower (34 firms) third of the sample.

Irrespective of debt financing measures, poorly governed (e.g. repressive portfolio) firms are found to have reasonably higher financial leverage than the firms in moderate portfolio, and the differences in Lvg.-1 (e.g. debt-to-assets) and Lvg.-3 (ratio long-term debt to long-term debt plus shareholders equity) are statistically significant.

Table 3 also reveals the degree of univariate relationships among several measures of debt finance and corporate governance indices. Columns 5 through 10 depict that all four debt financing measures are negatively correlated with the CGI and each of the five governance sub-indices. Most of the governance indices are also found to have statistically significant relationships with debt-to-assets and debt-to-equity.

The Regression Results

This sub-section presents the empirical results of the OLS regression model of the relationship between corporate governance and debt finance. Column 1 of table 4 depicts the regression result of debt-to-assets (e.g. the ratio of total debt to total assets) with CGI as the main explanatory variable, coupled with several control variables such as firm size, profitability, historical growth, asset tangibility and non-debt tax shield. It is shown that the regression coefficient of the CGI is significant and negative as expected. Both firm size and growth proxies are found to have statistically significant positive coefficients.

Moreover, the regression sign of firm profitability is negative and significant. In column 2, the new ownership concentration variable enters with the expected positive sign, which is statistically significant. The regression signs of the other explanatory variables remain identical, with only CGI, firm size, profitability and growth having significant results. The industry-adjusted results in column 3 reveal that the regression signs and the levels of significance of all explanatory variables remain unchanged12 .

***Insert

Replacing debt-to-assets with long term debt ratio (e.g. the ratio of long term debt to long term debt plus shareholders' equity) as the dependent variable, the similar specification is estimated. Column 4 of table 4 shows that the regression of long term debt ratio has not brought any change in the regression signs. However, only the CGI, firm size and profitability remain statistically significant.

The adjusted R 2 value of the regression of long-term debt ratio is around 0.720, which is increased to around 0.850 for the specification of debt-to-assets. The R 2 values suggest that the model's explanatory power in estimating the variability in debt finance is reasonably high, which provides a better overall fit for the population. In addition, higher level of significance of the F-statistics suggests an improved explanatory power of the regressor variables.

Analysis and Interpretation of the Results

The study results appear to corroborate the prediction of the agency theory that poor corporate governance and associated weak shareholder rights are linked with higher debt finance. The study thus supports the findings of [START_REF] Jiraporn | Capital structure, shareholder rights, and corporate governance[END_REF], whereby firms with poor shareholder rights favour higher debt ratio. However, this inverse relationship is unlikely to be a consequence of the role of debt in mitigating agency problems [START_REF] Jensen | Theory of the firm: managerial behaviour, agency costs and ownership structure[END_REF], rather it is an outcome of controlling shareholders' intention to retain ownership and control over the firm 13 . Although conflicts of interest between owners and managers persist, the latter do not seem to exert reasonable power and influence to cause higher agency costs, since the former have the incentive and authority to collect information and monitor management (see, [START_REF] Jensen | Theory of the firm: managerial behaviour, agency costs and ownership structure[END_REF].

The capital structure pattern in many developing economies like Bangladesh seems to suffer most from the agency problems created by the founding family or controlling shareholders, as suggested in several studies (for example, [START_REF] Shleifer | A survey of corporate governance[END_REF][START_REF] Chen | The controlling shareholder's personal leverage and firm performance[END_REF]. The controlling shareholders of poorly governed family-controlled firms tend to exert direct or indirect influence in the firm's financing decisions, as this is in their own interest, and this results in reduced rights for minority shareholder. These controlling shareholders want to preserve authority and informational advantage by choosing readily available bank debt toward meeting the firm's financing needs, whilst retaining or increasing ownership or control. This observation is supported by the evidence of positive association between ownership concentration and financial leverage.

Amongst the other determinants of capital structure, the regression results of profitability and firm size proxies are found to be significant and consistent. Evidence of a negative effect of profitability on debt ratio confirms the prediction of the pecking order theory, that firms will favour retained earnings over external finance. The positive influence of firm size on leverage also supports the theoretical prediction that large firms have an advantage over the smaller firms in obtaining long-term as well as short-term bank loans as they have the economies of scale, such as the opportunity for diversification and the ability to disclose more information [START_REF] Du | Ultimate corporate ownership structures and capital structures: evidence from East Asian economies[END_REF][START_REF] Fama | Agency problem and residual claims[END_REF].

Nonetheless, the corporate governance index, being the main test variable, is proved to be robust in all alternative specifications, even after controlling for firm-specific variables.

This corroborates the hypothesised negative influence of firm-level corporate governance 13 Alternatively, better corporate governance reduces agency cost of equity, which in turn enhances the firm's ability to raise equity finance and reduces the firm's reliance on debt. is also supported by the regression results whereby each of the five sub-indices has an inverse effect on financial leverage. The results appear to have important contribution to the existing literature on corporate governance and corporate finance in the context of bank-based financial systems in developing economies. On the one hand, the study confirms a significant negative influence of better governance quality on the firm's reliance on bank debt; [START_REF] Iturriaga | Debt ownership structure and legal system: an international analysis[END_REF] also finds that disclosure requirements have a negative influence on the bank debt of smaller firms. On the other hand, the study suggests that controlling shareholders of poorly governed firms use increased bank debt in meeting a firm's financing needs without sacrificing their ownership and control.

In the absence of a noticeable bond market in the country 14 , firms raise their debt finance mainly from the nationalised and private commercial banks. The greater negotiating power of controlling shareholders, along with their strong political ties and business relationships, appears to be the main reason the higher level of bank debt for poorly governed firms. Whilst this observation supports the findings of [START_REF] Semenov | Financial systems, financing constraints and investment: empirical analysis of OECD countries[END_REF], that close bank-firm relationships improve a firm's access to external (bank) finance, it contradicts the proposition that such relationships reduce capital market imperfections, especially in relation to a developing country like Bangladesh. This is because the crony relationship can help opportunistic businesspeople to get bank loans without improving the firm's governance practices. This eventually undermines the rights of both shareholders and depositors, leading to an imperfect capital market.

The banks (or debt finance) are therefore unable to mitigate the agency problems. Instead of serving the interests of depositors through proper vigilance and monitoring, the debt providers seem to become a part of the corrupted default system that serves the mutual interests of several parties, such as controlling shareholders of poorly governed firms, controlling owners of banks, ill-motivated bank managers of nationalised and private 14 Also, the country's stock market, with its small size (the ratios of market capitalisation to GDP in 2003 and 2004 were around 2.42 percent and 4.11 percent) does not seem to have any significant contribution to the firm's financing. In these circumstances, agency problems may arise, not only between the debt providers and minority shareholders of the borrowing firms, but also between the depositors and controlling owners of the lending firms (e.g. banks). This evidence supports the observation of [START_REF] Caprio | Corporate governance and finance: concepts and international observations[END_REF] that a large creditor is unlikely to be independent if he is directly or indirectly associated with the controlling family. It is important to mention that the creditors' rights in Bangladesh remain very weak because of incomplete and inconsistent Bankruptcy Acts, along with the ineffectiveness of the bankruptcy courts and money loan courts 15 , which in turn is the result of strong political lobbying of a group of business elites. Whilst [START_REF] Iturriaga | Debt ownership structure and legal system: an international analysis[END_REF] finds better protection of creditor rights has a positive influence on bank debt in developed economies, this proposition may not hold true for a developing economy, where poor creditor rights can cause a higher level of bank debt.

Bank-based debt finance may not only help poorly governed firms meet their financing needs without diluting the control of majority owners, but may also allow the latter to expropriate the interests of the minority shareholders and other stakeholders. The study also corroborates other empirical studies based on developing economies. [START_REF] Kumar | Corporate governance mechanisms and firm financing in India[END_REF], for instance, observes that debt can facilitate expropriation in economies like India, where institutions are generally ineffective. [START_REF] Du | Ultimate corporate ownership structures and capital structures: evidence from East Asian economies[END_REF] also argue that weak corporate governance and crony capitalism in East Asian economies have caused risky capital structures, leading to financial distress risks.

Conclusions

The paper empirically examines the relationship between firm-level corporate governance and the capital structure pattern of non-financial listed firms in Bangladesh. 15 See also Sobhan and Werner (2003). With a survey-based corporate governance index (CGI), the study investigates the effect of corporate governance on the total as well as long-term debt ratios. The cross-sectional regressions appear to confirm a significant influence of firm-level governance quality on a firm's capital structure, with the poorly governed firms having higher level of debt finance. The paper suggests that an inverse relationship between corporate governance and debt ratio is less likely to be a consequence of the role of debt in mitigating agency problems, and more likely to be an outcome of the controlling shareholders' reluctance to forgo absolute control rights. This is probably because controlling shareholders of a poorly governed firm intend to preserve their authority and informational advantage by choosing readily available bank debt, whist retaining their ownership or control of the firm. This notion is also supported by the evidence that ownership concentration is positively associated with the debt ratio. In this situation, debt providers appear to become a part of the crony relationships of powerful stakeholders, who capitalise on inefficient institutional framework to persistently cause poor firm-level governance. Notes: The OSL regressions are based on non-financial listed firms after dropping the observations that are identified as extreme outliers. ***, ** and * indicate statistical significance at the 1, 5 and 10 per cent levels, respectively. The figures in parentheses are the heteroscedasticity-adjusted standard errors. Few more observations are found to be outliers in the cross-section regression models. The problem of outliers has been treated by adding a dummy variable for the outliers in the regression model. The regression coefficients of the outlier dummy are turned out to be statistically significant at one percent level in all specifications.

F

  firm. Ownership concentration (e.g. measured as percentage of ownership by the top 10 shareholders) is also used as an additional governance variable, not been incorporated in the index. As mentioned above, the CGI and concentration variables are predicted to have negative and positive signs, respectively.

  corporate governance data based on a questionnaire survey carried out in 2004-2005. Amongst the 186 non-financial listed firms of the prime exchange of Bangladesh, the Dhaka Stock Exchange (DSE), 101 firms responded to the survey, with the response rate being around 55 percent. The sample firms capture nearly 96 per cent of the total market capitalisation (MC) of all non-financial firms, and 45 percent MC of the DSE 9 . The financial data on debt finance and other firm characteristics were collected from the annual reports of the sample firms, together with the monthly reviews of the DSE.

  .23. The standard deviation of the CGI is relatively higher, implying that the governance scores of many firms do not seem to be closer to the average governance index. This distribution is likely to be resulted from a widespread difference in governance qualities among the sample firms in various categories (such as foreign vs. local). Even so, nearly 50 percent of the sample firms appear to have CGI between 33 and 53.***Insert Table2about here***The correlation matrix of the sample firm presented in table

  . The robustness of the empirical effect of governance quality (e.g. CGI)

  leaders or bureaucrats with close relationships with the controlling shareholders of the borrowing and lending firms.

  Table 3 about here*** Column 1 of table3shows that the mean values of debt-asset ratio (Lvg.-1) and debtequity ratio (Lvg.-2) are 0.70 and 1.93, respectively, implying that the sample firms have higher proportion of debt in relation to equity. Altogether, it is evidenced that the non-

	F o
	r
	P
	e e r
	R e v i e w
	8

Table 4

 4

	about here***

Table 1 :

 1 Mean values of corporate governance index across industries

	o
	r
	P
	e e r
	R e v i e w

Source: Prepared by the authors based on questionnaire survey conducted in 2004-05.

Table 2 :

 2 Correlation matrix for corporate governance index and sub-indices

	Categories	CGI	(Sub-index-1)	(Sub-index-2)	(Sub-index-3)	(Sub-index-4)	(Sub-index-5)
	Ownership
	(Sub-index-1)	0.471*** 1
	Shareholders Rights
	(Sub-index-2)	0.864*** 0.204**	1
	Board
	(Sub-index-3)	0.749*** 0.343*** 0.628*** 1
	Disclosure
	(Sub-index-4)	0.835*** 0.194**	0.742*** 0.565*** 1
	Stakeholders Rights
	(Sub-index-5)	0.826*** 0.074	0.740*** 0.505*** 0.816*** 1
	Note: The correlation matrix is based on 101 non-financial listed firms. *, **, and *** denote significance at 10%, 5%,
	and 1% levels, respectively.

Page 17 of 17 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

Table 3 :

 3 Corporate governance and various measures of debt financeNotes: The table is based on primary data on non-financial listed firms in Bangladesh. *, **, and *** denote significance at 10, 5, and 1 per cent levels, respectively. Firms with the CGI of less than 35 are placed in the repressive portfolio (denoted as REP.), whilst the moderate portfolio (MOD.) consists of the firms with the CGI of greater than 48. Column 4 shows the difference (t-statistics) in the means of alternative leverage measures between the two portfolios. Lvg.1 = Total debt -to-Total assets, Lvg.2 = Total debt -to-Shareholders' equity, Lvg.3 = Long term debt -to-Long term debt plus Shareholders' equity, Lvg.4 = Long term debt -to-Shareholders' equity

			Mean Ratios			Correlation with CGI and Sub-indices
	Leve-	All REP. MOD.	Difference (t-stat.)	CGI	Sub-1	Sub-2	Sub-3	Sub-4	Sub-5
	rage	1	2	3	4	5	6	7	8	9	10
	Lvg1 0.70 1.02	0.50	0.52***	-0.33***	-0.17**	-0.34***	-0.04	-0.40*** -0.31***
	Lvg2 1.93 2.56	1.28	1.28	-0.20** -0.20** -0.17** -0.11	-0.15*	-0.12
	Lvg3 0.32 0.43	0.18	0.25**	-0.12	-0.01	-0.12	-0.14* -0.13*	-0.10
	Lvg4 0.44 0.58	0.30	0.27	-0.08	-0.07	-0.02	-0.09	-0.07	-0.05
	n	101	34	33		101	101	101	101	101	101

Table 4 :

 4 The OLS regression results of debt ratios against corporate governance index (CGI) and ownership concentration

	Dep.Var.		Debt-to-Assets		Long Term Debt Ratio
	Expl. Var.	1	2	3	4	5	6
	Intercept	0.014 (0.161)	-0.156 (0.157)	-0.224 (0.210)	-0.402** (0.175)	-0.487** (0.191)	-0.412* (0.249)
	CGI	-0.004*** (0.001)	-0.004*** (0.001)	-0.004*** (0.002)	-0.003** (0.002)	-0.003** (0.002)	-0.002* (0.002)
	Concentration		0.003** (0.001)	0.004*** (0.001)		0.002 (0.002)	0.003* (0.002)
	Firm Size	0.060*** (0.012)	0.060*** (0.011)	0.048*** (0.010)	0.059*** (0.016)	0.057*** (0.016)	0.047** (0.019)
	Profitability	-0.165*** (0.040)	-0.188*** (0.040)	-0.197*** (0.048)	-0.230*** (0.040)	-0.224*** (0.041)	-0.305*** (0.044)
	Growth	0.188*** (0.058)	0.169*** (0.063)	0.143** (0.063)	0.033 (0.080)	0.027 (0.074)	0.038 (0.078)
	Tangibility	-0.111 (0.086)	-0.099 (0.085)	-0.059 (0.101)	0.123 (0.093)	0.125 (0.094)	0.135 (0.110)
	Non-debt Tax Shield	-0.118 (0.100)	-0.104 (0.099)	-0.208* (0.109)	-0.180 (0.146)	-0.169 (0.143)	-0.263* (0.146)
	Industry Dummies	-	-	Yes	-	-	Yes
	F-statistics	71.63***	64.85***	30.59***	35.07***	30.93***	14.481***
	Adjusted R 2	0.836	0.840	0.853	0.719	0.720	0.734
	N. of Observations	98	98	98	94	94	94

Shleifer and Vishny (1997:737) define corporate governance as 'the ways in which suppliers of finance to corporations assure themselves in getting a return on their investment'.

[START_REF] Chowdhury | Capital structure determinants: evidence from Japan and Bangladesh[END_REF] investigates the determinants of debt finance, but does not take into account the governance issues.

Note that capital structure, leverage and debt finance are used interchangeably throughout the paper.

[START_REF] Jensen | Agency costs of free cash flow, corporate finance and takeovers[END_REF] argues that the obligation of paying debt along with its interest reduces free cash flow and thus managers refrain from using the free cash for non-optimal activities.[START_REF] Grossman | Corporate financial structure and managerial incentives[END_REF] Hart (1982, cited in Harris and[START_REF] Harris | The theory of capital structure[END_REF]) also observe that debt finance increases the probability of costly bankruptcy and subsequent job loss, and thus encourages managers to work harder, consume fewer perquisites and make better investment decisions.

Several studies (e.g. LLSV 1997;[START_REF] Shleifer | A survey of corporate governance[END_REF] suggest that corporate governance and associated better shareholder and creditor rights enhance the firm's ability to gain access to external finance.

According to the pecking order theory of corporate finance, retained earnings tend to be the most favoured forms of a firm's financing, followed by debt and then equity finance[START_REF] Singh | Corporate governance, corporate finance and stock markets in emerging countries[END_REF].

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

Note that the regression specification of column 3 is estimated by substituting the CGI with each of the five governance sub-indices, which is not shown in the paper. The regression results suggest that all five sub-indices are negatively associated with the debt ratio, and all but the ownership sub-index coefficient is statistically significant.

G28 -Government Policy and Regulation < G2 -Financial Institutions and Services < G -Financial Economics

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript