Hailin Liao
email: h.liao@lboro.ac.uk

Xiaohui Liu

O11 -Macroeconomic Analyses of Economic

Keywords: TFP, exports, Bound tests, lag-augmented VAR, East Asian economies TFP, exports, Bound tests, lag-augmented VAR, East Asian economies

Despite increasing interest in the relationship between trade and macroeconomic performance in development economics, very limited studies have been conducted on the causal links between exports and productivity growth in Asian economies. This paper examines empirically the interplay between exports and productivity growth for eight East Asian economies in a multivariate framework by applying bound tests and modified Wald tests. The results indicate that causality is bi-directional in the case of Korea, Singapore and Taiwan, while unidirectional from productivity to exports for Mainland China, Hong Kong, Indonesia, Malaysia and the Philippines. These findings provide little support for the conventional export-led growth hypothesis.

Export-Total Factor Productivity Growth Nexus in East Asian Economies

Introduction

Empirical investigation of the causal link between exports and growth is an important theme in trade and development literature. In particular, the rapid growth of many East Asian economies over the last four decades has stimulated much debate about sources of economic growth. The extend to which the East Asia's success is attribute to protectionist policies on the one hand and outward oriented policies on the other has been the focus of attention. A large number of empirical studies have been carried out on causality between trade and growth for this region [START_REF] Hsiao | Tests of Causality and Exogeneity between Exports and Economic Growth: the Case of Asian NICs[END_REF][START_REF] Chen | Export Performance and Productivity Growth: The Case of Taiwan[END_REF][START_REF] Ahmad | Unit Roots and Cointegration in Estimating Causality between Exports and Economic Growth: Empirical Evidence from the ASEAN Countries[END_REF][START_REF] Islam | Export Expansion and Economic Growth: Testing for Cointegration and Causality[END_REF][START_REF] Kwan | Exports, Economic Growth and Exogeneity: Taiwan 1953-88[END_REF]1999;[START_REF] Liu | Relationships between Economic Growth, Foreign Direct Investment and Trade: Evidence from China[END_REF][START_REF] Jin | On the Relationship between Openness and Growth in China: Evidence from Provincial Time Series Data[END_REF][START_REF] Awokuse | Export-led growth and the Japanese economy: evidence from VAR and directed acyclic graphs[END_REF][START_REF] Yao | On economic growth, FDI and exports in China[END_REF]. These studies have shed some light on the issue and provided some useful insights regarding the evaluation of trade and development strategies. However, previous studies have focused on the link between exports and GDP or output growth explicitly, or the relationship between exports and labour productivity growth [START_REF] Marin | Is the Export-led Growth Hypothesis Valid for Industrialised Countries?[END_REF][START_REF] Yamada | A Note on the Causality between Export and Productivity: an Empirical Reexamination[END_REF][START_REF] Thangavelu | Is There an Export or Import-led Productivity Growth in Rapidly Developing Asian Countries? A Multivariate VAR Analysis[END_REF].

What is absent in this area is an explicit assessment of the relationship between exports and technological progress represented by growth in total factor productivity (TFP) 2 although the possibility has already been postulated in development theory [START_REF] Marin | Is the Export-led Growth Hypothesis Valid for Industrialised Countries?[END_REF][START_REF] Ben-David | Trade and the Neoclassical Growth Model[END_REF]. Partial productivity measures have long been criticized for their incomplete picture of performance, thereby causing misleading analysis. They are also subject to the input substitution effect. It would be difficult to distinguish whether labour productivity being high in a sector is because of a high degree of technological efficiency or because of a large stock of 2 The World Bank (1993) supported the view that the promotion of exports had been a significant source of rapid productivity change through greater access to best practice technologies. However, this proposition has been strongly criticized by [START_REF] Rodrik | Getting Interventions Right: How South Korea and Taiwan Grew Rich[END_REF] and others on the basis of empirical work by Young (1995). The main argument was that substantial TFP growth was not observed (Dessus, 1999). However, results of many previous studies which tried to relate GDP growth to the growth of trade may be biased due to a simultaneity problem. 3 physical capital, given that labour productivity fails to capture all of the influences on productivity. TFP growth, on the other hand, measures the ratio of output to the sum of all basic inputs, and therefore mitigates the impact of factor substitution and scale economies. In addition, TFP evaluates technological progress and constitutes a measure of the efficiency with which all the factors of production are employed. Therefore, using TFP instead of labour productivity allows us to assess the impact of exports on technological progress. This study aims to fill the gap by evaluating the causal links between exports and productivity growth in East Asian economies.

Besides using TFP instead of labour productivity, we adopt the frontier approach to calculate the TFP, which enables us to overcome some drawbacks of the non-frontier measures of productivity, and represents an improvement over the previous studies (see Appendix A). The most important difference between the frontier approach and the traditional growth accounting method lies in one assumption i.e. the existence of an unobservable and idealized production possibility frontier with production-unit specific, one-sided deviation from the frontier which explicitly allows for inefficiency [START_REF] Farrell | The Measurement of Productive Efficiency[END_REF][START_REF] Lovell | Production Frontiers and Productive Efficiency[END_REF]Kumbhakar and Lovell, 2000).

We also apply different empirical methods to capture the nature of the links between these variables, which represents one of the contributions of our study. Bounds testing and modified Wald (MWALD) tests, both of which avoid the pre-testing problem in possible nonstationary time series, will be applied to examine causal links. The paper is structured as follows.

Section 2 discusses the theoretical foundations for possible bi-directional linkage between exports and TFP growth. Section 3 introduces the methodology and data used in the study, while the following section presents and analyses empirical results. Section 5 concludes with policy implications.

F

Theoretical Links between TFP and Exports

The new endogenous growth models establish the links between long-run growth and technological progress, and provide a framework in which trade can permanently increase the rate of growth in the host country through technology transfer, diffusion, and spillover effects. [START_REF] Romer | Idea Gaps and Object Gaps in Economic Development[END_REF] points out that one benefit that trade brings is access to new ideas. [START_REF] Grossman | Innovation and Growth in the Global Economy[END_REF] have constructed a theoretical model to show formally that trade in goods serves as a conduit for knowledge flows between countries. These flows in turn serve to increase the productivity of capital and labour, and hence the growth rate of per capita output. In addition, the human capital building model as presented by [START_REF] Lucas | On the Mechanics of Economic Development[END_REF] may suggest that trade could enable inter-country technology transfers.

Although trade is important for economic growth, the causal link between them is not necessarily unidirectional as productivity growth can also influence trade. Theoretical justifications for reverse causation from growth to trade have long been discussed in development literature [START_REF] Kanamori | Economic Growth and Exports[END_REF][START_REF] Diaz-Alejandro | Trade Policy and Economic Development in[END_REF]Kravis, 1970;[START_REF] Findlay | Growth and Development in Trade Model[END_REF]. It is argued that economic growth via increased productivity or reduced unit costs is expected to act as a stimulus to exports [START_REF] Kaldor | Strategic Factors in Economic Development[END_REF]. [START_REF] Jung | Exports, Growth and Causality in Developing Countries[END_REF] suggest that internal growth mechanisms better explain export growth rather than the reverse. In new trade theory, the market structure and output expansion may trigger significant changes in exports through a process of "cumulative causation" [START_REF] Venables | Trade Policy, Cumulative Causation, and Industrial Development[END_REF].

Given the possible bi-directional relationship between exports and TFP growth postulated by the theories, the issue becomes empirical and can be verified through statistical tests.

Attempts have been made to establish a causal link between exports and TFP growth empirically.

Previous studies in this area fall roughly into two groups: cross-country/sectional investigations and individual country analysis over time. correlation coefficient studies represented by the [START_REF] Balassa | Exports and Economic Growth: Further Evidence[END_REF] work, simple OLS regression between exports and output, and using production function approaches based on either the growth equation or [START_REF] Feder | On Exports and Economic Growth[END_REF] two-sector (exports and non-exports) model. The general conclusion from cross-country studies is that high levels of economic growth are significantly associated with high levels of export growth, though there are exceptions.

The main arguments against cross-sectional data analysis are that 1) it implicitly imposes or assumes a common economic structure and identical production technology across countries which are most unlikely to be held; 2) correlation does not necessarily imply/prove the direction of causality; 3) exports, via national accounting identity, are themselves part of national production. Therefore, any export growth studies which do not consider the endogenous nature of the growth process may be subject to simultaneity and specification biases [START_REF] Greenaway | What Does Liberalization Do for Exports and Growth[END_REF].

The critiques for the cross-sectional studies have led to the use of time-series data for causality tests, but the results are inconclusive. Using a regression equation similar to the sources of growth equation, with trade being specified as the growth of exports, about 20 time-series studies reached nearly identical results as cross-sectional studies (e.g [START_REF] Ram | Exports and Economic Growth in Developing Countries: Evidence from Timeseries and Cross-section Data[END_REF][START_REF] Sengupta | Exports and Economic Growth in Asian NICs: an Econometric Analysis for Korea[END_REF][START_REF] Kwan | Exports, Economic Growth and Exogeneity: Taiwan 1953-88[END_REF][START_REF] Greenaway | What Does Liberalization Do for Exports and Growth[END_REF][START_REF] Greenaway | Trade Liberalization and Growth in Developing Countries: Some New Evidence[END_REF][START_REF] Van Den Berg | The Relationship between International Trade and Economic Growth in Mexico[END_REF][START_REF] Gutierrez De Pineres | Export Sector Dynamics and Domestic Growth: the Case of Colombia[END_REF][START_REF] Vohra | Export and Economic Growth: Further Time Series Evidence from Less-Developed Countries[END_REF]. However, the size of the effect is sharply reduced when a quarter of these estimation procedures are adjusted in accordance with unit root test results. For example, the average value of the real export variable goes down from 0.26 to 0.08 [START_REF] Lewer | How Large is International trade's Effect on Economic Growth[END_REF]. More interestingly, many researchers have found evidence that trade generates more growth in developed economies than in developing countries. The probable reasons for the differences in coefficient value between high income and low income countries are attributed to low adsorptive capacity to foreign technology in developing countries [START_REF] Coe | North-south R&D Spillovers[END_REF]. Clearly, further time-series analysis is called for due to the following reasons. First, the simple formation of Granger causality regression lacks a theoretical foundation, and the twovariable relationship between exports and growth is examined in isolation; therefore, omitted variable bias (e.g. import variable) is likely to occur [START_REF] Riezman | The Engine of Growth or its Handmaiden? A Time Series Assessment of Export-led Growth[END_REF]. Second, there are some problems in relation to estimation procedures. For example, the lag length is selected arbitrarily in some previous studies [START_REF] Jung | Exports, Growth and Causality in Developing Countries[END_REF]. Moreover, F-test statistics for causality tests have been applied. It is now well established that the F-test statistics are not valid if times series are integrated [START_REF] Toda | Statistical Inference in Vector Autoregressions with Possibly Integrated Processes[END_REF][START_REF] Zapata | Monte Carlo Evidence on Cointegration and Causation[END_REF]. Finally, there are limited studies on the causality between TFP and exports, probably due to data unavailability. Extending previous work in this area, the present study attempts to estimate the complex interaction between exports and TFP growth in a multivariate framework (including import variable) for selected Asian economies using bound tests and MWALD tests.

Empirical Methodology and Data

Bounds Testing

The concept of Granger causality is well known, and the test for causality is straightforward with stationary data. However, if the series are non-stationary, testing the causal link between variables in their levels leads to statistics which do not have the relevant F-distribution under the null hypothesis [START_REF] Toda | Vector Autoregression and Causality[END_REF]. One way to deal with this is to apply a cointegration approach. However, this approach is typically affected by pre-tests on unit roots and the cointegration rank. The power of such pre-tests is low, and subsequent tests of long-run relations and causality conditioned on these pre-tests could induce severe pre-test bias 3 . Pesaran et al.

3 A potential weakness of the techniques for cointegration (for example, Engle and Granger procedure and Johansen's procedure) is that they require certain pre-testing for unit roots and also that the underlying time series to be integrated are of the same order. If this is not the case, the evaluation of the long-run relationship may not be performed within the context of cointegration, at least in the two-variable context.

F

+ + + + + + = = = 0 1 1 2 1 1 1 0 , t=1, 2… (1)
In equation ( 1), y t is the dependent variable; x t is a vector of independent variables; 1 and 2 are the respective long-run multipliers for x and y; and represent the short-run dynamic coefficients 4 ; c 0 is the drift component; c 1 is the time trend and t is a zero mean stationary process (model V in [START_REF] Pesaran | Bounds Testing Approaches to the Analysis of Level Relationships[END_REF]. If there is no deterministic trend but an intercept (model III in [START_REF] Pesaran | Bounds Testing Approaches to the Analysis of Level Relationships[END_REF], the model can be defined as 5

t b j i t j i t a i i t t t x y x y c y + + + + + = = = 0 1 1 2 1 1 0 (2)

MWALD Approach to Causality

The modified Wald (MWALD) test developed by [START_REF] Toda | Statistical Inference in Vector Autoregressions with Possibly Integrated Processes[END_REF] and further developed by [START_REF] Dolado | Making Wald Tests Work for Cointegrated VAR Systems[END_REF] is applied to investigate long-run causalities as this approach does not require knowledge of cointegration properties of the system 'so long as the order of integration of the process does not exceed the true lag length of the model' [START_REF] Toda | Statistical Inference in Vector Autoregressions with Possibly Integrated Processes[END_REF].

We consider a trivariate system, namely Y (productivity growth in manufacturing), X (export growth) and Z (import growth), in the following trivariate VAR system. (3)

+ + + + + = + + + + z x y d k t d k t d k t d k k t k t k t
4 Due to the uncertainty whether x is the long-run forcing variable for y, we exclude the current value of firstdifferenced x in the regression of equation ( 1) and ( 2), according to [START_REF] Pesaran | Working with Microfit 4.0: Interactive Econometric Analysis[END_REF]. 5 For a detailed discussion of the model, see [START_REF] Pesaran | Bounds Testing Approaches to the Analysis of Level Relationships[END_REF].

= = = = k a a a H
for the test that export growth x does not Granger-cause TFP growth y, where ) ( 12 i a are the coefficients of export growth x, i=1,2,…k in the first equation of the system. Equation (3) can also be augmented with a constant and a linear trend.

There are two reasons for an import variable to be included. First, bivariate systems are often criticised as incomplete, omitting potentially important variables. Failure to account for omitted variables can lead to misleading causal ordering, and yield biased results. Second, [START_REF] Riezman | The Engine of Growth or its Handmaiden? A Time Series Assessment of Export-led Growth[END_REF] have pointed out that 'standard methods of detecting export-led growth using Granger-causality tests may give misleading results if imports are not included'. Hence, import growth in the trivariate system is treated as an auxiliary variable for 'indirect measures of technological adoption' [START_REF] Coe | North-south R&D Spillovers[END_REF]. This specification allows us to investigate whether the causality results would remain the same by switching from the bivariate system to a trivatriate one 6 . Hence, we study only direct causality between exports and TFP growth by controlling the possible indirect causality running from imports to productivity.

Robustness Check

We also employ more conventional methods to check the robustness of our results from the methods discussed above. ADF unit root tests are performed to determine the order of integration of the variables 7 . As far as the causality is concerned, the VECM is performed in order to compare its results with those from MWALD tests.

Data 6 Since trivariate tests incorporate more information than bivarite ones, the causal inferences drawn appear more reliable. 7 Results are available upon request. The sample used in this study consists of the first and second wave of newly industrialised economies (NIEs) in East Asia, notably Hong Kong, Singapore, South Korea, Taiwan, Indonesia, Malaysia and the Philippines. In addition, China, as a large developing country in the region, is also included. TFP growth rates in manufacturing sectors calculated in [START_REF] Liao | Productivity Growth of East Asia Economies' Manufacturing: A Decomposition Analysis[END_REF] using a stochastic frontier approach are weighted by their share in the manufacturing value-added in the current year in order to obtain annul aggregate data for each economy 8 . Data for exports and imports are derived from World Development Indicators (2000) and various annual reports from the Asian Development Bank. All the data are in real terms, adjusted to 1990 prices. For example, the data for exports and imports were adjusted to the 1990 prices using their respective export and import price indices. When price indices were not available, GDP deflators were used. The sample period is from 1963 to 1998 for most of the sample economies, but from 1981 to 1998 for China, due to data unavailability.

Empirical Results

Results from Bound Tests

The key step in bound tests is to determine the appropriate lag length. A trade-off must be considered between choosing p sufficiently small so as to avoid the problems of overparameterisation in small samples, and sufficiently large to avoid serial correlation in the residuals. We follow the approach of [START_REF] Pesaran | Bounds Testing Approaches to the Analysis of Level Relationships[END_REF] to estimate the conditional ARDLerror correction model by the OLS, with and without a linear time trend, and determine the appropriate lag length by a combination of Schwarz Bayesian Criterion (SBC) 9 with the maximum lag length set to T 1/3 . Various diagnostic tests are also performed such as residual 8 Detail on the frontier approach to TFP calculations applied in this paper is presented in Appendix A. 9 The SBC usually selects a lower lag order when taking the small sample into consideration. In our case, the approach we employed is similar to Hendry's general-to-specific approach, where the rationale is to re-estimated the basic model by dropping the lagged variables with insignificant parameters from the system. Insert Table 1 The results of the bounds testing presented in Table 1 suggest the existence of a long-run relationship between TFP growth and exports in all economies in question. It appears that high TFP growth leads to an increase in exports in the case of China, Indonesia, and the Philippines,

given that the calculated F-statistics are greater than the upper level of the critical value bounds at the conventional significant level. Reverse causation running from exports to productivity growth exists in the case of Korea, Singapore and Taiwan. Earlier conventional views on East Asia's success have frequently stressed export orientation as a major source of growth. However, the preliminary results from our study provide evidence that exports might have played a smaller role toward productivity growth than was previously thought, at least in some sample economies.

There are also some inconclusive results from the ARDL cointegration test based on the F-statistics. For example, the results for Hong Kong and Malaysia are quite mixed. As Bahmani-Oskooee and Miteza (2002) argue, a more efficient way of establishing cointegration is to estimate the entire equation ( 1) and ( 2), and check the significance of the error-correction term (ECT) since the long-run effects are reflected in the significant coefficient obtained for the lagged ECT. We adopt this approach to compare whether the results are consistent (Appendix table 1B). Most of the coefficients of ECT are statistically significant with the expected negative sign, confirming the results for cointegration from the F-test. We also estimate the long-run coefficients for the case where a significant long-run relationship was found (Appendix table 2B).

Insert

Causality Analysis

MWALD Test for Causality

The same approach is adopted for selecting the lag length when testing causality. That is, the optimal lag order (k) is selected by AIC and/or SBC statistics from unrestricted VAR (k) estimated by the OLS over the same sample, with the maximum lag length set to T 1/3 . One of the criteria for lag length selection is to eliminate autocorrelation. However, we cannot take the risk of over-parameterization by choosing a higher order for the VAR, given that our sample period is relatively short. Hence, the causality tests are performed by estimating VAR (k+1), where k is the optimal lag order in the VAR as the variables are, at most, I(1) processes. [START_REF] Rambaldi | Testing for Granger non-causality in Cointegrated Systems Made Easy[END_REF] proved that the MWALD method for testing Granger nocausality can be computationally simple by running a Seemingly Unrelated Regression (SUR) in several build-in econometric packages, such as Shazam and Microfit 10 , which use the F-statistic involving coefficients of lags 1 through k. The results from the MWALD test are summarised in Table 3. It can be seen that the results are sensitive to model specifications (i.e. with and/or without linear trend and constant) which will affect the inferences drawn from the causality tests. 10 The tests are performed using the SUR routine in both software packages for comparison. The results are quite similar. However, as Microfit4.0 utilizes maximum likelihood estimation (MLE) rather than generalised least square (GLS) used in Shazam, the estimated standard errors for the MLE are generally smaller and therefore test statistics are smaller. We only report test statistics and rejection p-value from Shazam for simplicity. In order to avoid contradictory results, we follow Mills' (1998) approach and use information criteria (SBC here) to select jointly the lag order of the unrestricted VAR and the order of any additional trend polynominal to determine the optimal models, and these are shaded in Table 3.

Insert Table 3, here,

The evidence shows that the export-led TFP growth exists in Korea and Singapore, while the productivity-led exports appear to be likely in China, Hong Kong, Indonesia, Philippines and Malaysia. There is bi-directional causality in Taiwan, implying that exports and productivity growth have reinforced each other. For the second-tier East Asian economies, such as Indonesia, Philippines and Malaysia, these results suggest that the productivity growth does not respond to lagged changes in exports. This is contradictory to the conventional argument that openness, especially exports, trigged economic and productivity growth in East Asia. The evidence provides little support for the export-led growth hypothesis. In contrast, productivity is found significant in explaining the future path of exports, confirming the productivity-led export hypothesis. Hence, our results support the traditional argument that improvements in productivity growth translate into greater competitiveness in export growth in the standard Ricardian model. It may be the case that the economy improves its productivity through technology spillovers from advanced economies according to the endogenous growth theory, and this enhanced productivity further stimulates exports. On the other hand, the export-led productivity scenario found in the first-tier East Asian economies, with the exception of Hong Kong, provides support for outward-oriented trade strategies under which competition in export markets could lead to greater efficiency as local firms face greater competition from foreign firms.

VECM and Standard Granger Causality

We also perform a standard Granger causality test augmented with an appropriate ECT taken from the appropriate long-run relationship to detect causality between productivity growth and exports. Concerning the choice of optimal lag length, we adopt Akaike (1969)'s final prediction error (FPE) criterion. The FPE can be calculated as (SER) 2 (N+k)/N, where SER stands for the standard error of the regression, k is the lag length used in the regression and the optimal lag length corresponds to the minimum FPE, taking into account residual serial correlation.

Intuitively, the test of whether exports Granger-cause productivity growth or vice versa is simply a test of the hypothesis that the coefficients of lagged independent variables are all equal to zero and/or the lagged ECT is significantly different from zero.

Insert Table 4 The results presented in Table 4 indicate that there is a bi-directional causality between exports and productivity growth in Korea, Singapore, Taiwan and Malaysia. The channel of influence from exports to productivity is the significant ECT for Korea and Singapore, and both short-run and long-run effects exist in Taiwan. Conversely, there is short-run reverse causation from productivity to export performance for Singapore and Taiwan, and long-run causation for Korea, based on the adjustment toward long-run equation. Causality is unidirectional, running from productivity to exports, for China, Hong Kong, Indonesia, Malaysia and the Philippines.

Both long-run and short-run causal links exist in Hong Kong, whereas only long-run causality is found in the remaining four economies.

Since there are some arguments 11 that the standard Granger-causality test is very sensitive to the choice of the lag length, we perform the tests with lag length from k=1 to k=4 to see whether the lag length selected will affect the results considerably (results not report here).

However, the results in general do not change markedly across lag lengths. Therefore, the specification, using the ECT derived from model V or model III of ARDL-cointegration test with various lag lengths, does not affect the results from causality tests in our study. Taken together, the results obtained from both MWALD and VECM tests are consistent (Table 5), thus providing robust evidence on the causal link between exports and TFP growth for the sample economies.

Insert Table 5, here

Conclusions

This paper investigates empirically the linkage between exports and TFP growth for eight East Asian economies by applying ARDL bounds tests and MWALD tests. Our main findings are that 1) the long-run relationship between productivity growth and export performance exists for all economies under investigation; and 2) causality is bi-directional in the case of Korea, Singapore and Taiwan, while unidirectional from productivity to exports for the rest, which provides little support for the export-led growth hypothesis.

The diversity of results across countries/economies undoubtedly reflects a wide variety of outward orientation strategies (especially exports) implemented across the region. It is also probably related to the fact that these economies have different patterns and are at the different stage of development.

In some economies, export expansion and TFP growth go hand in hand, suggesting that exports and TFP growth are integral elements in the growth process. However, in other economies, TFP growth has played a more important role in promoting exports, as found at firm

11 For example, Bahmani-Oskooee and Alse (1993, p540) pointed out the danger of an arbitrary choice of lag length by claiming that one must 'select a strategy for choosing the optimum number of lags on each other when there is level data, where the more productive firms are more likely to become exporters [START_REF] Bernard | Exporting and Productivity[END_REF][START_REF] Jin | On the Relationship between Openness and Growth in China: Evidence from Provincial Time Series Data[END_REF]. Our finding also indicates that the role of imports should be considered in the export-TFP growth link as imports also indirectly affect both exports and TFP growth. (2) ** the null hypothesis of on long-run relationship is rejected at 5% significance level.

(3) ? the result is inconclusive as the F-statistic falls inside the band. where, variables with a dot over them represent growth rates, and the first term on the right-hand side is the output elasticity of frontier output with respect to time, defined as TP, the second term measures the input growth weighted by output elasticities with respect to input j,

+ = = • • ) ( ) 1 ( ) , ( ln (A.6)
12 Due to the lack of data on input prices, the output elasticity with respect to input j is equal to input share in the total production cost under the assumption of perfect competition. 13 Returns to scale can be defined as RTS= e j In this study, we are going to estimate equation (A.5) allowing for it v , a symmetric component capturing random variation across production unit and random shocks that are external to its control, into the composed error term with an attempt to distinguish the effects of statistical noise from those of inefficiency so as to obtain consistent and efficient estimates.

  functional form and normality. We also employ the bond tests to detect which variable (productivity or exports) is the long-run forcing variable.

  , the conventional conceptualization of TFP growth can be defined as output growth unexplained by input growth 12 , i.e. jt is the price of j-th input and c is the total costs. Combining equation (A.2) and (A.3of perfect competition and constant returns of scale, the output elasticities with respect to input j is equal to input share in the total production cost, therefore, TP is the only source of TFP growth. In case of unavailability of input price information, we follow

  A.6), TFP growth consists of three components: technical change (innovation and shifts in the frontier technology), technical efficiency change (catching-up) and returns to scale (SEC). That is, TFP growth is useful in distinguishing innovation or adoption of new technology by 'best practice' production units from the diffusion of technology. Coexistence of a high rate of TP and a low rate of change in technical efficiency may reflect the failures in achieving technological mastery or diffusion(Kalirajan, Obwona & Zhao, 1996). However,Nishimizu & Page (1982, p926) ignored the presence of measurement error ( it v ) in estimating the parameters of the translog approximation to equation (A.5) by using a deterministic frontier.

  (2001) propose bound tests as one way of avoiding pre-test bias. The bounds arise from two polar cases, namely joint stationarity of all variables, and non-stationarity of all variables, with cointegrated cases being intermediate. The tests are formulated via the following regression:

									a				b
	y	t	c	c	t	y	t	x	t	i	y	t	i	j	x	t	i	t
									i				j
						o r
								P
									e
									e r
											R
													e
													v i e
																		w

Table 2

 2

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

  Table2summarises these two tests for the long-run relationships using F-statistics and ECT representation; the latter only shows the results that are inconclusive in the F-test as a supplement. Taken together, the results from these two tests are consistent, indicating that the long-run relationship from productivity growth to export performance exists in the cases of Mainland China, Hong Kong, Indonesia, Malaysia and the Philippines, and from exports to productivity growth in Korea, Singapore and Taiwan.

Table 1 :

 1 F-statistic for testing the existence of long-run relationship (ARDL) Critical value bounds of the F-statistic [4.87 5.85] for modelV and [3.79 4.85] for model III at 5 % significance level.

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

more than one independent variable' in a VAR model.

Table 2 :

 2 Summary of ARDL Cointegration Tests

			Using F-statistics	Using EC representation
	Dependent Variable	Model V	Model III	Model V	Model III	Conclusion
	CHN	DTFP	X	X
		DEXPT
	HK	DTFP	?		X	X
		DEXPT
	IND	DTFP	X	X
		DEXPT		X
	KOR	DTFP
		DEXPT	?	X	X
	MAL	DTFP		?		X
		DEXPT	?
	PHL	DTFP	X	X
		DEXPT
	SGP	DTFP		X
		DEXPT	X	X
	TW	DTFP
		DEXPT	X	?		X
	Note: -a long-run relationship exists; X -no long-run relationship; ? -inconclusive

Table 3 :

 3 MWALD Non-Causality Tests : tfp stands for productivity and expt represents for exports; the arrow in the second column shows the direction of non-causality; the last column indicates the decision: means there is a causal effect. ***, ** and *denote significance at the 10%, 5% and 1% levels, respectively.

			With time & constant	With constant only	Without time & constant
		hypothesis F-statistic p-value	F-statistic p-value	F-statistic p-value	Decision
	CHN Expt tfp	20.3348*	0.0000			0.5249	0.7692
		tfp expt	30.1042*	0.0000			10.8415* 0.0044
	HK	expt tfp	3.3576	0.1866			3.6787	0.1589
		tfp expt	9.9760*	0.0068			9.1600*	0.0102
	IND	expt tfp	0.1182	0.7310	0.2743	0.6005
		tfp expt	10.6602*	0.0011	3.5837*** 0.0584
	KOR expt tfp	2.3887	0.3029	13.3396* 0.0013
		tfp expt	13.6509*	0.0011	7.2535** 0.0266
	MAL expt tfp	4.2764	0.1179	0.8921	0.6401
		tfp expt	1.3881	0.4996	7.5254** 0.0232
	PHL expt tfp	0.3065	0.5799	0.1814	0.6702
		tfp expt	0.6881	0.9334	0.2093	0.6473
	SGP	expt tfp	1.1277	0.2883			0.4225	0.5157
		tfp expt	4.5672**	0.0326			1.1146	0.2911
	TW	expt tfp			41.0349* 0.0000	36.3419* 0.0000
		tfp expt			7.4597*	0.0063	9.1492*	0.0025

Note

Table 4 :

 4 Results of Causality between productivity & export growth (the VECM approach)

			Model V	Model III
		hypothesis	(var.)	EC t-1	(var.)	EC t-1
	CHN expt tfp 2.4393	0.8330	2.2834	0.8045
			(0.157)	(0.432)	(0.172)	(0.448)
		tfp expt 4.9102	2.8108***	0.0167	4.3014*
			(0.112)	(0.067)	(0.899)	(0.001)
	HK	expt tfp 0.0456	0.1479	0.0605	0.0776
			(0.956)	(0.884)	(0.941)	(0.939)
		tfp expt 4.2219**	3.6787*	4.4185**	3.4112*
			(0.032)	(0.002)	(0.028)	(0.003)
	IND	expt tfp 0.0689	0.2225	0.4083	0.7672
			(0.796)	(0.826)	(0.530)	(0.452)
		tfp expt 5.1150**	4.4307*	5.3631**	2.8913*
			(0.034)	(0.000)	(0.031)	(0.009)
	KOR expt tfp 0.4997	6.4286*
			(0.613)	(0.000)
		tfp expt 0.4894	2.9276*
			(0.619)	(0.007)
	MAL expt tfp 1.0828	0.8034	3.0164***	2.767*
			(0.386)	(0.434)	(0.063)	(0.014)
		tfp expt 2.8840***	1.2178	5.1859*	2.5656**
			(0.071)	(0.242)	(0.012)	(0.022)
	PHL expt tfp 1.3218	1.0542	1.0898	0.9166
			(0.261)	(0.302)	(0.306)	(0.368)
		tfp expt 0.0882	2.2902**	0.1396	2.4099**
			(0.916)	(0.032)	(0.871)	(0.025)
	SGP	expt tfp 2.3269	3.8801*
			(0.119)	(0.001)
		tfp expt 2.9894***	0.4443
			(0.069)	(0.661)
	TW	expt tfp 4.2100*	4.4015*	4.0528*	4.2747*
			(0.013)	(0.000)	(0.015)	(0.000)
		tfp expt 6.9263*	1.708	7.3827*	1.9073***
			(0.001)	(0.104)	(0.001)	(0.072)
	Note: The arrow in the second column shows the direction of non-causality under H 0 . The joint significance of the
	coefficients is ascertained by the standard F-test. The level of significance for the error-correction term (lagged by
	one period) is determined by the standard t-statistics. ***, ** and *denote significance at the 10%, 5% and 1%
	levels, respectively.

Table 5 :

 5 Comparison of Two Causality TestsTotal factor productivity in this paper is measured by using frontier approach. The frontier approach is capable of capturing both efficiency change and technological change as components of productivity change, while the non-frontier approach assumes the presence of full technical efficiency and technical progress. The latter is resulting from the advanced technology embodied in capital and represented by the outward shifts in the production frontier over time, which is synonymously considered to be the unique source of TFP growth. Defined this way, TFP growth derived from non-frontier approach is at best a measure of Hicks-neutral disembodied technological change and at worst nothing more than 'a measure of our ignorance'(Abramovitz, 1956). We define this so called 'best practice' function f(.) as,

		VECM		MWALD
		S/L	L/R	L/R
	CHN	exp tfp
		tfp exp
	HK	exp tfp
		tfp exp
	IND	exp tfp
		tfp exp
	KOR	exp tfp
		tfp exp
	MAL	exp tfp
		tfp exp
	PHL	exp tfp
		tfp exp
	SGP	exp tfp
		tfp exp
	TW	exp tfp
		tfp exp

  As the estimates of the error correction representation selected by AIC and SBC are very similar, we only report the results of the models selected by AIC. Number inside the parenthesis is the absolute value of the t-ratio.

	Table 1B	Error Correction Representation of ARDL Model
	Dependent	S/R				Model V			Model III
	variable		forcing k=0		k=1		k=2	EC t-1	k=0	k=1	k=2	EC t-1
	CHN DTFP	dexpt	0.0052		-0.0106	-0.0070	-0.8420	-0.001	0.0026	-0.1993
					(0.8758)	(0.8754)	(1.1029)	(1.3683)	(0.6375)	(0.6462)	(1.8816)
		DEXPT dtfp		52.9557		-18.0087	-73.3377	-4.3955	29.4230	-3.3100
					(0.8758)	(3.7818)	(2.3166)	(7.1234)*	(0.9865)	(3.4093)*
	HK	DTFP	dexpt	0.0004					3.0563	0.0004	2.6515
					(1.2559)				(3.7847)	(1.155)	(4.0484)
		DEXPT dtfp		134.9978				-1.7415	152.0988	-1.7518
					(3.0602)				(6.1653)*	(4.0855)	(6.2875)*
	IND	DTFP	dexpt	0.0099		-0.0103	-0.0060	-1.044	0.0050	0.1179
					(2.5552)	(1.9781)	(1.8928)	(1.8260)	(1.5558)	(2.7557)
		DEXPT dtfp		22.1974		66.1577	-111.025	-2.1533	15.1431	76.4854	-141.12	-1.6942
					(2.2738)	(2.1754)	(3.1835)	(7.5221)*	(1.3404)	(2.1131)	(3.553)	(6.0715)*
	KOR DTFP	dexpt	-0.0001					-0.3912	0.0002	-0.0008	-0.0004	0.0002
					(0.6823)				(3.1198)*	(1.0865)	(2.9232)	(1.945)	(0.0022)
		DEXPT dtfp		-142.08		251.56		124.41	-0.3661	279.41	-1.000
					(0.7933)	(2.7977)	(2.5289)	(0.4781)	(1.8130)
	MAL DTFP	dexpt	0.0024		0.0014		0.0020	-0.1995	0.0012	0.0046	0.0028	-0.0001
					(2.3042)	(0.7201)	(2.1130)	(2.446)	(1.1017)	(2.5675)	(2.780)	(0.0042)
		DEXPT dtfp		66.3934					-1.8086	4.9733	-1.6198
					(2.4143)				(7.9809)*	(1.4151)	(6.7408)*
	PHL	DTFP	dexpt	-0.0001					-0.4208	0.0001	-0.0004
					(0.2117)				(2.8132)	(0.0475)	(0.2199)
		DEXPT dtfp		-31.8002				-1.6684	-0.7139	-1.6432
					(0.3200)				(6.3104)*	(0.7284)	(6.6448)*
	SGP	DTFP	dexpt	0.0003		-0.0007		-0.6390	0.0025	-0.0043	0.0104
					(0.2450)	(1.8173)		(3.3277)*	(1.5079)	(3.2703)	(1.065)
		DEXPT dtfp		-116.8951				-1.0000	3.7794	-1.0000
					(1.7289)					(1.4401)
	TW	DTFP	dexpt	0.0083		-0.0471	-0.0133	-0.3171	0.0099	-0.0488	-0.0132	-0.2597
					(0.9835)	(3.1549)	(1.8028)	(3.5784)*	(1.2418)	(3.3479)	(1.815)	(6.1791)*
		DEXPT dtfp		6.1095		-4.6426		-1.000	8.1239	-4.73	-1.000
					(1.6171)	(1.5388)			(2.5169)	(1.5669)
	* represents 5% significant
	Table 2B	Estimated long-run coefficients
	Dependent	Model V underlying ARDL				Model III underlying ARDL
	variable		order	regressor IMPT	const	T	order	regressor IMPT	Const
	CHN Expt (3,3,3) 25.994	0.5412	-16.834	-1.3311	(3,0,3) 8.889	0.4412	-55.290
				(4.6076)	(6.0447)	(4.2169)	(4.2884)		(1.0531)	(2.6728)	(0.8585)
	HK	Expt (0,0,0) -16.0056	0.915	46.2372	-0.1274	(0,0,0) -10.126	0.9467	27.9211
				(0.6838)	(2.3385)	(0.7187)	(0.9209)		(0.4511)	(9.1187)	(0.4581)
	IND	Expt (3,3,1) 27.8176	0.7933	-16.1388	-4.2956	(3,3,1) -2.0688	1.2331	8.5152
				(3.0146)	(4.4082)	(1.8546)	(3.2724)		(1.2527)	(5.8912)	(1.6437)
	KOR TFP (3,0,3) -0.0003	-0.0038	5.9011	-0.0032
				(0.7782)	(2.1254)	(183.4289)	(3.6254)
	MAL Expt (2,1,0) 18.5859	0.3285	-49.3962	-0.8849	(2,0,0) 3.0702	0.3206	-4.4071
				(3.0771)	(4.4782)	(2.8311)	(2.5176)		(1.4795)	(3.5411)	(0.5606)
	PHL Expt (2,0,0) -19.0608	0.3567	-89.3058	5.4254	(2,0,0) -0.4345	0.3694	4.4463
				(0.3246)	(2.6098)	(0.3022)	(0.3172)		(0.7365)	(2.8206)	(2.4658)
	SGP TFP (1,2,0) 0.0006	-0.001	5.0916	-0.0686
				(0.2604)	(0.4856)	(105.7088)	(85.959)
	TW	TFP (2,3,2) 0.2897	-0.0903	1.5706	-0.0695	(2,3,2) 0.3799	-0.1255	-0.8946
				(2.6354)	(1.3321)	(0.5667)	(0.8954)		(6.6151)	(1.8362)	(1.7879)

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

Development < O1 -Economic Development < O -Economic Development, Technological Change, and Growth, F43 -Economic Growth of Open Economies < F4 -Macroeconomic Aspects of International Trade and Finance < F -International Economics

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


D58 -Computable and Other Applied General Equilibrium Models < D5 -General Equilibrium and Disequilibrium < D -Microeconomics, F15 -Economic Integration < F1 -Trade < F -International Economics, F12 -Models of Trade with Imperfect Competition and Scale Economies

Keywords: Trade Costs, European Single Market, Enlargement

This paper addresses 'deeper economic integration' 1 -the idea that regional integration, by addressing institutional, …nancial, regulatory and infrastructure issues using, for example, harmonisation and mutual recognition agreements, 2 can promote trade beyond what can be attained by tari¤ removal alone. In particular, I analyse the likely impact of the recent accession of several Central and Eastern European Countries (CEECs) to the European Union. To this end, I derive estimates of aggregate border costs, including many of which are not explicit trade barriers. 3 These are inferred by comparing actual trade with that predicted from country size and transport costs. The novel aspect is the use of a fully modelconsistent procedure for identifying such costs, by calibrating a multi-country Dixit-Stiglitz general equilibrium model directly upon observed trade ‡ows.

The conclusions are that trade patterns in 1997 are consistent with border costs between the existing EU and accession states of between 0 and 33 per cent. If these barriers were overcome by deeper economic integration, one might expect overall trade between these areas to increase by 50-100 per cent, leading to gains in income of 10-20 per cent for the accession states and small gains for the existing EU members. While the conclusions broadly agree with previous studies, the implied gains are, if anything, somewhat more optimistic.

The layout of the paper is as follows. Section 2 discusses the derivation and economic role of border costs. Section 3 outlines the calibration and general equilibrium methodology, especially the model-consistent calibration methodology, as well as the data used. In section 1 See Lawrence, 1995.

2 On mutual recognition, the most comprehensive reference is Maskus and Wilson's (2000) World Bank study.

3 e.g. the e¤ects of product standards, labelling and such like, and the associated application of testing and border checks.

Estimated gravity models

Studies such as LeJour et al (2001) or the literature surveyed in Anderson and Wincoop (2004) derive costs by inference from trading patterns, using gravity models, on the grounds that 'bottom-up' estimates may miss many of the costs involved. If a pair of countries trades much less than might normally be expected, then we might well infer that there is some cost obstructing that trade. Tari¤s, NTBs and transport costs can be deducted, leaving an estimate for the net e¤ect of regulations, currency conversion costs and even the informational costs of …nding trade partners where trade has been di¢ cult in the past (see Rauch (1999), Edwards (2006)).

Trade is usually compared to an assumed underlying gravity framework, using equations (in logs) for trade in good i between exporting country, c; and importing country, cc; of the form:

X i;c;cc = i + i d ic;cc + 1i Y c + 2i Y cc + i;c;cc ; (1)
X i;c;cc denotes exports, Y is GDP and d is distance between capitals of c and cc. Theory suggests that trade should be roughly proportional to the product of country sizes and inversely proportional to distance. Gravity models can be empirical or more theory-driven (see Bergstrand (1989), Anderson and Van Wincoop (2003)).

While gravity equations usually work reasonably well for trade between large subsets of countries, there is a clear mis…t in the case of the national borders of the importing country.

Trade within almost any country substantially exceeds than that with neighbouring countries, even after taking account of distance. This 'missing trade'can be very substantial [START_REF] Bergstrand | The Generalised Gravity Equation, Monopolistic Competition and the Factor-Proportions Theory of International Trade[END_REF] , on Turkish exporters only,. while the causes are controversial [START_REF] Brown | An economic assessment of the integration of Czechoslovakia, Hungary and Poland into the European Union[END_REF] .

Despite their intentions, the EU Single Market and introduction of the Single Currency are far from eliminating home bias across member states. Nevertheless, there are also limited regional bias e¤ects linked to the regional trade bloc border. These are potentially particularly important when we come to analyse the impact of Single Market enlargement.

A fuller gravity model could therefore be written (in logs) as:

X i;c;cc = i + 1i D ijs EU + 2i d c;cc + 1i Y c + 2i y c + 3i Y cc + 4i y cc + (2)
X d d Dd i;c;cc + 1i T M i;c;cc + v 2i T E i;c;cc + i;c;cc
where Dd is a set of dummies for border e¤ects, with D EU set to 1 if both c and cc are EU members, otherwise set to zero. T M i;c;cc is the tari¤ on imports of i from c to cc.

T E i;c;cc is the export tari¤ levied by country c on country cc. y c and y cc represent income per capita, which can also be shown to have signi…cant e¤ects. In the LeJour et al (2001) study, estimated trade between EU members in the late 1990s was between 0-250 % higher than that between EU members and the CEECs.

Cost estimation

To estimate border costs, we also need estimates of demand and supply elasticities. Ideally, these should be consistent with the estimated e¤ects of tari¤s and formal NTBs. In addition, it is necessary to make an assumption whether border costs are …xed or variable with respect to trade. The standard response is to assume they are 'iceberg'costs, which eat up a …xed [START_REF] Brown | An economic assessment of the integration of Czechoslovakia, Hungary and Poland into the European Union[END_REF] Obstfeld and Rogo¤ (2000) argue that a combination of border regulatory costs, currency conversion costs, informational costs and under-measurement of transport costs explains much of the di¤erence. Anderson and Van Wincoop (2003) cost borne in the …rst instance by the importer. This could be consistent with bureaucratic or testing costs in the limited case where a …xed proportion of goods imported have to be tested, or with 'pure horizontal' regulatory barriers (in the sense that di¤erent countries' rules specify di¤erent technical speci…cations, but which have no tangible di¤erence in terms of the quality of the good or service experienced by consumers). [START_REF] Dixit | Monopolistic Competition and Optimum Product Diversity[END_REF] Unlike a tari¤, an iceberg border cost yields no revenue: consequently its abolition gives a much greater boost to overall incomes.

The above resource cost speci…cation is open to challenge. Many trading costs (e.g. the redesign of a good, or testing a variety just once to meet di¤erent standards), may impose a lump-sum cost on importers. If producers are identical, then there will be a threshold level of lump-sum cost of market entry. The resource cost of this may be greater or less than that of an iceberg cost, depending upon elasticities and the scale of reduction in observed trade volumes, but numerical analysis suggests that, if entry is seen to have been deterred, the cost of a lump-sum barrier must be high. [START_REF] Edwards | Quality Standards under Classical Oligopoly and Trade: Regulatory Protection or Just Over-Regulation?[END_REF] The idea that regulatory di¤erences are of the 'pure horizontal'variety is open to criticism (see Edwards, 2003). Di¤erent standards may be better suited to the di¤erent natonal tastes. Higher standards may raise consumer welfare, though at the expense of producers. Despite these objections, I continue to use the iceberg cost assumption in this paper, partly for consistency with previous studies and partly due to its relative simplicity of [START_REF] Dixit | Monopolistic Competition and Optimum Product Diversity[END_REF] See, again, Maskus and Wilson (2001) or Edwards (2003) for more detail. [START_REF] Edwards | Quality Standards under Classical Oligopoly and Trade: Regulatory Protection or Just Over-Regulation?[END_REF] The author's calculations indicate that, for a demand elasticity of 4, the threshold lump-sum entry cost imposes a higher welfare loss than the iceberg trade cost consistent with a halving of trade volumes.

Speci…cation of the simulation model

The most appropriate simulation tool for major trade changes is usually held to be a multicountry, multi-sector CGE model. The neoclassical Heckscher-Ohlin model framework is not seen as realistic, since it only allows for trade in one direction. The major alternatives are the Armington formulation -a somewhat ad hoc model where producers within a country are perfectly competitive, but di¤erent countries'goods are di¤erentiated -and the Dixit-Stiglitz (D-S) formulation, which is explicitly derived from a 'love-of-variety' model of consumer demand, and usually assumes Chamberlinian monopolistic competition in the long run between producers of di¤erentiated varieties. Under many circumstances, both the Armington and the D-S models share many behavioural properties (downward-sloping demand curves, improvements in consumer utility from being able to spread demand across a variety of sources), although the latter can also incorporate price mark-ups and potential gains from capacity rationalisation (see Krugman, 1979, Baldwin and Venables, 1995).

Where inputs have a variety e¤ect and where transport costs are high, the D-S formulation can result in agglomeration economies, making the model prone to multiple equilibria. In common with the 'top-down'studies summarised above, I derive implicit trade costs from observed trade. However, unlike previous work, rather than estimating equations, I derive residual border e¤ects by direct calibration of a theoretical D-S model (see below), which are then fully consistent with the model used for simulation. I calibrate residual border e¤ects for imports and exports between each pair of countries (constructing averages for intra-EU trade by model-consistent CES aggregation). This compares to the more parsimonious set of dummies used in most gravity studies, which constrain many residual border e¤ects to be equal. I also use direct estimates of transport costs from the GTAP project, rather than arc distances. In addition, compared to gravity estimation, more speci…c account is taken of the importance of relative output prices, calibrating for revealed comparative costs on the basis of a certain set of restrictive assumptions has been made about border e¤ects. [START_REF] Edwards | An Economic Assessment of the Accession of the Central and Eastern European Countries to the EU Single Market[END_REF]

Derivation of border and comparative production costs

The theoretical relationship between the D-S model and the gravity model is well-established since Bergstrand (1989), and forms the basis for the calibration in this paper. Since the calibration technique is novel, I outline it in detail. The subscript i is dropped throughout this section. I assume goods are consumed in countries c 2 1:::C, which I also index as cc.

Imports of good g into country cc are Q g;cc . Total consumer utility in country cc is assumed 1 0 In the absence of better information, I have assumed these border costs are symmetrical in both directions between a pair of countries. Note that the assumptions made about relative border e¤ects can potentially a¤ect our picture of the underlying competitiveness of the CECs in di¤erent industries -further information on this might help guide future work.

U cc = " X c X g2c ((1 c;cc )Q g;cc ) ( 1)= # =( 1) ; (3)
where is the elasticity of substitution between goods varieties, and c;cc is an iceberg cost.

Setting the marginal utility of consumption of g equal to its relative price and rearranging:

Q g;cc = U cc h (1 c;cc ) ( 1)= ( cc =P c (1 + c;cc )(1 + t c;cc ) i ; (4)
where c;cc is the proportionate transport cost and t c;cc is the net contribution of import and export tari¤s, subsidies and the tari¤ equivalents of NTBs. P c is the selling price of goods from country c at the point of export, while cc is an aggregate consumer price index for cc.

Next, rewrite (4) in terms of observable variables. The nominal value of exports from c to cc, E c;cc is the number of goods varieties produced in country c, n c , times sales per good, Q g;cc (g 2 c) times the export price P c , upscaled by (1 + c;cc ) to take account of the transport cost . U cc equals total expenditure in cc, Y cc divided by the aggregate price index cc . n c equals output X c , divided by the turnover of a 'representative'…rm, T c . Hence

E c;cc = (1 c;cc ) ( 1) X c Y cc T 1 c P 1 c 1 cc (1 + c;cc ) 1 (1 + t c;cc ) (5)
For cc = c we can replace E c;cc with H cc (home use). For H cc ; cc;cc = t cc;cc = 0. This means that, rearranging (5), and dividing by the version for H cc gives us:

E c;cc =H cc = (1 c;cc )=(1 cc;cc )) ( 1) (X c =X cc )(T cc =T c )(P c =P cc ) 1 (1+ c;cc ) 1 (1+t c;cc ) (6)
We can rearrange this to put (1 c;cc ) on the left hand side, and if we assume c;cc = 0 for c = cc; we can simplify somewhat:

(1 c;cc ) = f(E c;cc =H cc )(X cc =X c )(T c =T cc )(P cc =P c ) 1 (1 + c;cc ) 1 (1 + t c;cc )g 1=( 1) : (7)
An interesting result is found if we multiply together these expressions for trade in both directions between a pair of countries, c and cc, since a lot of terms can then be eliminated:

(1 c;cc )(1 cc;c ) = h q Ẽc;cc Ẽcc;c j p H c H cc i 2=(1 ) ; (8)
where the tilde represents adjustment for tari¤s, NTBs and transport costs. If the geometric average volume of trade between a pair of countries, once tari¤s and transport costs have been corrected for, is signi…cantly smaller than that of home-based consumption in the two countries, then there must be residual border costs present. Assuming residual border trade costs are the same in each direction, then

c;cc = 1 h q Ẽc;cc Ẽcc;c j p H c H cc i 1=( 1) : (9)
Once an estimated value for the elasticity of substitution, has been chosen, all the other terms on the right hand side of ( 8) are given, so that, for given observed output, consumption and trade, the higher the value of the trade cost, c;cc , the lower will be the implicit trade cost in the other direction, cc;c . We can therefore use data on observed trade

The model for simulations

Simulations on the removal of these trade costs are carried out using a multi-country static CGE model, as outlined in Appendix 1. Goods are produced using a Cobb-Douglas aggregate of intermediate inputs and 4 primary factors: unskilled labour, skilled labour, capital and land. Land is …xed sectorally. Both types of labour are mobile between sectors, but not internationally. For capital, I investigate two variants, one where it is …xed in total within a country, and one where it is internationally mobile.

Intermediate inputs and …nal consumption goods are CES aggregates of home production and imports from various sources. The elasticity of substitution between di¤erent sources of a good is set at 4 in all sectors. [START_REF] Edwards | Import Search and the Path-Dependency of Trade[END_REF] This lies at the lower end of elasticity estimates surveyed by Anderson and Van Wincoop (2004), which vary from 4 to 5 up to around 10.

However, it is more consistent with the elasticities traditionally used in Armington-based general equilibrium models. A key factor may be time-scale (in this model I am interested in medium-run simulations, where perhaps long-run elasticities would be higher).

There are also iceberg transport costs, iceberg unspeci…ed trade costs (see above) and tari¤s, as well as taxes/subsidies on output and use of a commodity.

Firms are imperfectly competitive, and charge markups dependent on their market shares.Treatment of the number of …rms within an industry and country is an issue here. Many D-S models assume that the number of …rms is endogenous. Trade liberalisation leads to a general shakeout of capacity (see Krugman, 1979). The small loss of consumer variety from the shakeout is o¤set by a larger gain from allowing …rms to reap economies of scale (see Baldwin and Venables, 1995). However, endogenised …rm numbers are really a long-run assumption, since …xed costs may not be avoidable in the short-or medium-run.

For this reason, I retain a 'medium-run'formulation, with a …xed number of …rms in each country/industry.

The top level of the consumption function uses a Cobb-Douglas structure.

Data

I use the GTAP version 5 database, which has harmonised trade and input-output data for regions across the world in 1997. I aggregate data into 8 goods 12 and 10 regions [START_REF] Harrigan | OECD Imports and Trade barriers in 1983[END_REF] , chosen for their relevance to the issue of enlargement [START_REF] Harrison | Increased Competition and Completion of the Market in the European Union: Static and Steady State E¤ects[END_REF] . For trade and protection I use 4 principal data series from GTAP for these countries and regions:exports at market (i.e. domestic) prices (VXMD), exports at world prices (VXWD), imports at world prices (VIWS) and imports at market prices (ie sales prices in the importing country before indirect tax) (VIMS). Trade volumes, transport costs and tari¤s are derived from these series. [START_REF] Hoekman | Deep Integration, Regionalism and Non-Discrimination[END_REF] 4 Results

Border costs

In 1997, imports from the CEECs into the EU faced sizeable barriers in agriculture and food processing, but formal barriers elsewhere had been removed under the Europe Agreements.

CES weighted averages of the barriers facing CEEC exports to the EU varied between 10 and 31 per cent for agriculture, while for food processing they were 25-54 per cent. Even after taking account of these formal trade barriers, there is still a considerable shortfall compared to intra-EU trade.

Table 1 shows calibrated comparative costs and country bias. In this case, average 'excess' bias on EU-CEEC trade is assumed to be the same in both directions. The …rst column shows calibrated production costs relative to the EU. This suggests the CEECs two are not on the 2005 EU accession list, whereas the 3 Baltic States, as well as Cyprus and Malta, are.

1 5 VIWS -VXWD is the transport cost margin. VXWD -VXMD is net export tax/subsidy, plus the GTAP estimates of the tari¤ equivalent of some quantitative trade restrictions whose revenue accrues to the exporting country.

VIMS -VIWS is net import tax/subsidy and the tari¤ equivalent of remaining NTBs.

Correction is made for some data errors in GTAP V5. I have removed tari¤s on trade between the EU and CEECs other than in agriculture and food processing. The second column is average calibrated home bias between EU states, which is far from having been eliminated by the Single Market. The remaining two columns show 1 6 Comparative costs in services would, of course, be expected to be lower in poorer countries (see Balassa, 1964). However, it seems that, at least for Poland, the low relative costs apply to all sectors. Only for the Other CEEC region does there seem to be clear evidence supporting the Balassa-Samuelson relationship. average calibrated iceberg costs of trade in both directions on trade between the EU and CEECs, assuming they are symmetrical [START_REF] Lawrence | Regionalism, Multilateralism and Deeper Integration[END_REF] . These vary from slightly negative (for Polish food processing only) to around 15% for Polish manufactures, 10-13% for other CEEC manufactures and 9-14 per cent for Hungarian manufactures. For agriculture they are around 7-10%. The key assumption is that Single Market entry can remove these costs.

Enlargement simulations

Table 2 (below) shows the e¤ects on consumer welfare in each region resulting from (1) customs union (removal of the tari¤s on agriculture and foodstu¤s and harmonisation of external tari¤s with those of the EU) and ( 2) assumed abolition of iceberg unspeci…ed trading costs, i ; when countries join the Single Market. Customs union has only small welfare e¤ects, though these generally bene…t the accession states by 0-2 [START_REF] Anderson | Gravity with Gravitas: a Solution to the Border Puzzle[END_REF] 2 % while having no signi…cant e¤ect on existing EU members. The former e¤ect is not surprising given the fact that most tari¤s have already been abolished, while the latter re ‡ects the small size of the CEEC economies relative to the existing EU.

Under (2) the trade between the EU and CEEC is increased to re ‡ect the supposed removal of trade costs following Single Market accession. Since it is assumed these are real resource costs, all countries can gain, including outsiders. The biggest bene…ciaries are the CEEC countries, where welfare rises by 10-20% compared to 1997 base. Gains to the existing EU members are typically around 1 2 %. Even the poorer EU countries in the South experience gains of 0.4%, with expansion outweighing the costs of cheap-wage competition.

The Former Soviet Union and LDCs also see small gains, so that trade diversion e¤ects are 1 7 If costs were asymmetric, then equation ( 8) would imply that, if border costs were lower in one direction than in Table 1, they would have to be higher in the other direction.

Table 2

Summary of results -change on 1997 base, calculated consumer utility Table 3 shows the change in trade volumes: these are typically of the order 50-100% between the EU countries and CECs on accession. appear to be few losers.

2. Prices in the EU generally fall as a result of the saving in costs of inputs (the unskilled wage in Germany is set to 1 in this model, to act as a numeraire). However in the CEEC countries -especially Poland -output prices generally rise towards EU levels.

3. Relative skilled/unskilled wages do not change greatly in any country, though there are sizeable gains to both types of labour in Poland in particular. The lack of distributional changes between types of labour may partly be because a …xed factor (land) in two sectors absorbs much of the e¤ects of changes in output prices. [START_REF]Quantifying the Impact of Technical Barriers to Trade[END_REF]

Comparison with other studies 4.3.1 Border costs

The calibrated residual border costs vary between 7 and 15 per cent of total cost of goods traded -which suggest they are roughly twice as large as implied by 'bottom-up'estimates such as Harrison et al (1996) or Zahariadis, 2002. I suggest this represents a general discrepancy between gravity models and attempts to quantify the e¤ects of observable barriers. [START_REF] Mccallum | National Borders Matter: Canada-US Regional Trade Patterns[END_REF] Possible reasons are that di¤erent regulatory regimes involve more inconvenience and uncertainty to trading …rms than simple cost estimates imply, or that there may be …xed costs involved which have not properly been modelled, or alternatively that trade costs are magni…ed by the presence of informational barriers (see Edwards, 2006). 2 0 However, a higher trade elasticity, as suggested by, for example, Harrigan (1993), could reduce the implicit border costs consistent with a particular observed trade pattern. Direct comparison of the counterfactual simulations of a comparative static model with actual country experience is not simple. Accession to the EU Single Market is a lengthy process, requiring adaptation of trade barriers, laws, standards and macroeconomic policies over a number of years, as well as adjustment by …rms and consumers. Consequently, even though the o¢ cial EU accession date for the Poland was 2005, it is by no means easy to establish precisely when adjustment began or will end. In the meantime, global prices and technology have changed, while the CEECs are also readjusting to a market economy and a more transparent political system -a process assisted by, but not entirely dependent upon EU accession.

Looking at the period between 1997 and 2004 (just prior to o¢ cial accession), Poland's GDP grew by around 29% (or 3.7% per annum) [START_REF] Obstfeld | The Six Major Puzzles in International Macroeconomics: Is There a Common Cause?[END_REF] , much faster than the Eurozone as a whole. According to the Polish Ministry of Economic A¤airs and Labour, Poland's exports more than doubled, rising as from 20% to 33.7% of GDP, [START_REF] Piazolo | The integration process between Eastern und Western Europe[END_REF] with the growth being spread widely across commodities (as our model would indicate). Imports grew less rapidly in percentage terms. The EU accounted for 78% of Poland's exports and 68% of imports in A tentative conclusion would be that the 80-100% surge in Polish-EU trade predicted by the model has e¤ectively happened in anticipation of accession (2004 saw particularly rapid trade growth). [START_REF] Rauch | Networks versus markets in international trade[END_REF] The timing of this surge should not be seen as surprising, given that formal trade barriers (except in agriculture and food products) were removed in the mid-1990s, while implementation of the various EU regulations and standards has been a gradual process. The degree to which Poland's rapid GDP growth can be attributed to rising trade with the EU would require much more detailed study: if one assumes that there are up-front costs in changing a country's regulatory regime, and that developing products and market ties to new countries also involves search and learning-by-doing, [START_REF] Tre ‡er | The Case of the Missing Trade and Other Mysteries[END_REF] then, the long-run welfare gains could take longer to emerge than the increase in trade.

Summary and Conclusions

This paper addresses the potential bene…ts of deeper integration of the CEECs into the EU's Single Market, associated with the 2005 accession. I assume that di¤erent regulatory and currency regimes pose a signi…cant barrier to trade, proxied by an 'iceberg'-type border cost, whose removal is potentially relatively cheap. Such barriers are estimated by comparing the trade volumes between countries which are already members of the Single Market in 1997 with those which were not. This involves assumptions about the nature and speci…cation of the barriers imposed by di¤erent regulatory regimes, and the underlying structure of the economy. With these assumptions, and using a Dixit-Stiglitz type general equilibrium 2 3 Some of the trade growth observed would, of course, re ‡ect underlying steady-state economic growth. Based upon this approach, trade between the EU and accession states in 1997 fell well short of its potential, as judged by inter-EU trade patterns. Consequently, entry to the Single Market could be seen as potentially raising trade between the accession states and existing EU by 50-100 % compared to 1997, with trade between Poland and the EU rising particularly fast. Unlike purely gravity-based studies, the general equilibrium approach in this paper is able to derive estimates of associated GDP gains: for the EU's new members from Central and Eastern Europe, these are of the order of 13-20 per cent from entry into the Single Market and abolition of remaining tari¤ barriers. I do not look at the e¤ects of agricultural or regional subsidies. These cost estimates are somewhat larger than those of Le Jour et al 's (2001) model, which included fewer competitive and variety e¤ects. Hence, the study can be viewed as giving a bullish interpretation of the trade e¤ects of 'deep integration' of the accession states, although some other potential bene…ts from institutional reforms are omitted. [START_REF] Zahariadis | Modeling Deep Integration in the Context of the EU-Turkey Customs Union: Working Notes[END_REF] A quick comparison with Poland's actual experience since 1997 that the sizeable increase in predicted trade has largely taken place in the years prior to actual accession. However, there may be ongoing welfare bene…ts as …rms and consumers adjust to membership. 2 5 On this point, Piazolo's (2001, Chapter D) estimates suggest that an improvement in institutional quality by one third, as measured by the EBRD index, would increase the level of GDP by at least one tenth purely due to the greater e¢ ciency of resource use estimate with our CGE model. The total increase in GDP in some of the institutionally more backward states, allowing for additional investment over time, could be 20 to 30 percent. Notes on the structure of the model:

The model is based on an imperfectly competitive structure, using a Dixit-Stiglitz framework. 26 In this paper I have used the simpler version of the model, where the number of goods produced within each country, c, and industry, i, n c;i , is …xed. Each good is produced by one …rm and denoted g. However, unlike many Armington models, it does allow for monopolistic markups. I also allow capital to ‡ow between countries.

Production of goods.

Each …rm combines labour, land and capital using a Cobb-Douglas function to form a value added input: i.e.

V A g = c;i (LU

u c;i g LS s c;i g ) (1 k c;i d c;i ) K k c;i g Dg d c;i ; (
= V A c;i P V c;i u c;i (1 k c;i d ci )=W U c ; (A2) LS c;i = V A c;i P V c;i s c;i (1 k c;i d c;i )=W S c ; K c;i = V A c;i P V c;i k c;i =R c;i ; D c;i = V A c;i P V c;i d c;i =LDP c;i ;
where W U , W S; R and LDP are factor rents.

The price of value added,

P V c;i = (W U c LU c;i + W S c LS c;i + R c K c;i + LDP c;i D c;i )=V A c;i : (A3)
Higher level of production function.

Output of i, Y c;i , is a Cobb-Douglas aggregate of intermediate inputs II of other goods,

ii; and value added, V A:

Y c;i = c;i V A v c;i c;i Y ii II I c;ii;i c;ii;i : (A4)
Assuming cost-minimisation, with P U and P Y denoting input and output prices,

II c;ii;i = I c;i Y c;i P Y c;i =P U c;ii;i ; and (A5) V A c;i = v c;i Y c;i P Y c;i =P V c;i : (
c;ii = (V A c;ii P V c;ii + X i II c;i;ii P U c;i )=Y c;ii : (A7)
The marginal per unit price including tax and subsidies:

P Y c;ii = P P Y c;ii (1 + OU T T AX c;ii ) SU BSIDY c;ii =Y c;ii : (A8)
Trade and the aggregation of goods.

I use a two-stage nested utility function. Total demand for industry bundle i in c is a Dixit-Stiglitz aggregate T U c;i :

T U c;i = 0 @ X g2i g;c U g;c 1 A 1= ; (A9)
where U g;c is use of g in c and is a parameter re ‡ecting compatibility and home bias in consumption. The substitution parameter = ( 1)= , where is the elasticity of substitution between g in i.

If depends only on country of origin, cc, country of use, c, and industry, i, then we sum, …rst, the various varieties g 2 (cc \ i); produced in each country cc, and then sum across countries: Total expenditure in country cc on goods in industry i:

T U c;i = 0 @ X cc X g2(cc\i) cc;c;i U g;c;i 1 A 1= : (
V U cc;i = X c QU i;c;cc P U U i;c;cc : (A12)
The price P U of the aggregate bundle T U :

P U cc;i = V U cc;i =T U cc;i : (A13)
Consumption (top level of the nested utility function).

Utility U T is a Cobb-Douglas aggregate across i of consumption CN i :

U T c = Y i CN c i;c i;c : (A14) CN i;c = c i;c CE c =P U c;i : (A15)
Competition and pricing.

Within export markets, it is assumed that a …rm has a very small market share and so its own-price elasticity is . By contrast, in the home market, the …rm's market share S g;c is assumed to be signi…cant, so the …rm's price elasticity The top level utility function is Cobb-Douglas in functional form (so the elasticity of substitution between consumption of the produce of each industry, i, is unity). Share parameters for each product class are calibrated from value shares in total expenditure.

h c;i = + (1=n c;i )(1 SM c;i )(1 ); (
The lower level utility function has an elasticity of substitution between goods g in industry i of : This is assumed to equal 4 in all industries.

Supply side: technology is assumed to be Cobb-Douglas, so elasticities of substitution between inputs are unity, and share parameters can be directly calibrated from total costs (once monopoly pro…t has been subtracted).

  g;c;cc of the value of exports from c to cc. For example, Baldwin et al (1997) and LeJour et al (2001) assume that non-membership of the Single Market imposes an iceberg

  Of previous simulations of Eastward EU enlargement, LeJour et al (2001) used …tted coe¢ cients based upon a gravity model of trade and a multi-country Armington CGE model. There was no guarantee of theoretical consistency between the …tted gravity equations and the simulation model. Baldwin et al (1997) utilised a D-S general equilibrium model, but simply assumed that trade between EU members and non-members carried a (fairly arbitrary) 10% iceberg cost. Zahariadis'(2001) single-country on possible Turkish accession utilised more conservative bottom-up estimates of border costs.

  a) between existing EU member states and b) between existing and future member states (as well as that between pairs of future member states) to infer the cost-equivalence of assumed regulatory barriers to trade.

1 1

 1 The love of variety characteristic of the Dixit-Stiglitz model requires the substitution elasticity > 1: Note that the originalDixit-Stiglitz (1977) paper was written in terms of a substitution parameter, = ( 1)= ; which correspondingly has to lie in the range between zero and unity.

1 4 Page

 4 Due to data limitations, I am unable to carry out simulations on the precise accession list of 2004. The other CEEC region comprises the Czech Republic, Slovakia, Slovenia, Romania and Bulgaria. The latter 11

  low-cost producers compared to the EU in almost all industries, especially services[START_REF] Krugman | Increasing Returns, Monopolistic Competition and International Trade[END_REF] , agriculture, and light and heavy manufactures. Hungary is low-cost in textiles, while the OCEC region is high-cost in iron and steel.

  by the e¤ects of the overall expansion of the EU and CEEC economies.

1 9

 9 See Edwards and Whalley (2006).

2

 2 Enlargement simulation resultsBaldwin et al 's work (1997) used a relatively sophisticated D-S formulation, and also incorporated estimates of the impact of the Common Agricultural Policy and of structural and regional assistance, as well as capital mobility. However, for trade barriers they simply assumed a 10% across-the-board iceberg cost (which exceeds bottom-up estimates, but is smaller than the implicit costs for many sectors in this study), as well as a capital 'risk premium'. The LeJour et al study (2001) applies a detailed (but not strictly model-consistent) gravity model estimation to a more conventional Armington multi-country CGE model. The two previous studies both found signi…cant welfare and trade gains from EU enlargement. Baldwin et al 's (1997) simulation results, based on an assumed 10% iceberg cost on trade between the EU and CEECs, are shown in Table 4 (below). Le Jour et al (2001) also …nd substantial bene…ts for the accession countries, particularly Poland, though not as sizeable as in Baldwin et al (1997). This is not surprising since the former use an Armington model, which does not model all the bene…ts which a D-S model captures. Both studies agreed that enlargement involves few costs for existing EU members, though LeJour et al (2001) imply France may have lost slightly. The study in this paper …nds GDP gains to Poland of around 20% of GDP, to Hungary of nearly 18% and just over 13% for the other CEEC countries, with small gains for the existing EU countries. These estimates exceed those in LeJour et al (2001), but are roughly in line with Baldwin et al's (1997) …gures. The di¤erence may largely re ‡ect the greater gains from trade in a Dixit-Stiglitz compared to an Armington framework, with signi…cant pro-competitive and variety gains to utility.

2 4

 4 Rauch's (1999) work on networks and trade, in particular, would indicate this kind of learning process is involved. calibrate directly for implicit trade costs -so avoiding the questions of model consistency which apply to previous work.

Appendix 1 :

 1 General equilibrium model (GEMEE).

  A10)If all …rms within an industry/country are identical in size, then we can write T U cc;i = X c n ci i;c;cc (QU i;c;cc =n c;i )

( 1 +

 1 A16) where, if HU denotes consumption from domestic suppliers and P T i;c;cc is the selling price, SM c;i = 1 HU c;i P T i;c;c =V U c;i :The overall own price elasticity is a weighted average from home and export markets:o c;i = h c;i (HU c;i =Y c;i ) + (Y c;i HU c;i =Y c;i ): (A18)Monopolistic competition markups: the …rm marks up production costs by a proportionM M c;i = 1=(1 (1= o c;i )) 1: (A19)The price of good g including monopoly markups is therefore:P M c;i = P Y c;i (1 + M M c;i ): (A20)Transport costsTransport costs are proportional to value, so the price including transport,P T R i;c;cc = P M c;i (1 + T marginc i;c;cc ): (A21)This is an iceberg cost, so, if X i;c;cc is the quantity of i leaving country c for country cc, the amount which arrives in country cc is:M i;c;cc = X i;c;cc =(1 + T margin i;c;cc ): (A22)Tari¤sTari¤s are expressed as a percentage rate, so the price including tari¤s isP T i;c;cc = P T R i;c;cc (1 + tarif f i;c;cc =100):Home use of goods, HU c;i ,equals total production less exports:HU c;i = Y c;i X cc X i;c;cc : (A35)Total use of good i in country c produced in country cc equals home use, where c = cc.Otherwise it equals imports.Use tax is an ad valorem tax on all use of i in country c, giving revenueT U Y c = X i HU c;i P T i;cc + X cc P T i;cc;c EX i;cc;c =(1 + tmargin i;cc;c ) !! u setax c;i ):(A36)Output tax, OT, is a tax per unit value of output of an industry above.Total consumer expenditure in country c;CE c = X i V A c;i P V c;i + X i Y c;i P Y c;i M M c;i + X cc X i (EX i;cc;c P M cc;i T R i;cc;c )=100) + X i OT c;i V A c;i P W c;i + X ii II c;ii;i P U c;ii ! +T U Y c T SU BY c BOT c KM c RB c : (A37)The Balance of Trade, BOT c , (including long-term net capital payments) is …xed:BOT c = X i Xcc EX i;c;cc P M c;i X cc EX i;cc;c P M cc;i KM c RB c : Key assumed parameter values: Demand side:

  Firm sizes are equal within each industry across countries. Iron and Steel and Heavy Manufacture are seen as the least competitive industries (1 and 3 …rms respectively in our smallest region, Hungary), followed by Other Primary, Light Manufacturing, Textiles and Food Processing (4-6 …rms per industry in Hungary). Services and agriculture have much smaller …rm sizes.

  argue that omitted variables and model-inconsistencies mean that McCallum's estimated border e¤ect is vastly overstated.

Table 1

 1 Calibrated relative production prices and iceberg cost of home/country bias

	POLAND	Relative Intra-EU	EU v Poland	Poland v EU
	Industry	Price	Home Bias
	AG	-0.41	0.68	0.08	0.08
	OP	-0.21	0.50	0.20	0.20
	FP	-0.35	0.68	-0.05	-0.05
	TEX	-0.30	0.55	0.09	0.09
	IS	-0.01	0.56	0.16	0.16
	MH	-0.40	0.59	0.14	0.14
	ML	-0.41	0.53	0.17	0.17
	SV	-0.38	0.82	0.06	0.06
	HUNGARY	Relative	Intra-EU	EU v Hungary Hungary v EU
	Industry	Price	Home Bias
	AG	-0.35	0.68	0.10	0.10
	OP	-0.50	0.50	0.33	0.33
	FP	-0.41	0.68	0.05	0.05
	TEX	-0.35	0.55	0.06	0.06
	IS	-0.14	0.56	0.19	0.19
	MH	-0.45	0.59	0.14	0.14
	ML	-0.39	0.53	0.09	0.09
	SV	-0.45	0.82	0.06	0.06
	OTHER CEECs Relative	Intra-EU	EU v OCEC	OCEC v EU
	Industry	price	Home bias
	AG	-0.36	0.68	0.09	0.09
	OP	-0.16	0.50	0.30	0.30
	FP	-0.41	0.68	0.06	0.06
	TEX	-0.20	0.55	0.08	0.08
	IS	0.24	0.56	0.13	0.13
	MH	-0.31	0.59	0.11	0.11
	ML	-0.34	0.53	0.13	0.13
	SV	-0.36	0.82	0.04	0.04

Table 3

 3 Changes in trade volumes with trade share shifts and mobile capital assumed

	Total trade volumes Before After % change
	Poland to EU	4.98	9.12 + 83%
	Hungary to EU	2.62	4.34 + 65%
	Other CEC to EU 6.05	9.81 + 62%
	EU to Poland	1.88	3.77 +100%
	EU to Hungary	1.45	2.20 + 51%
	EU to Other CEC 3.56	5.50 + 55%
	To summarise some further results 18 :
	1. Output gains are spread widely across all industries in the CEEC region, though

the biggest gains are to agriculture, food products and manufactures. Within the EU there

1 8

Full details and tables are available in

Edwards (2004)

. 15 Page 15 of 32 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

Table 4 :

 4 Comparison with other studies. Real income per cent change

		Baldwin et al		LeJour et al
		1997		2001
		Conservative Less conservative Custom union Single market entry
	CEEC7	2.5	18.8	1.3	5.3
	EU15	0.2	0.2	0.0	0.1
	EFTA 3	0.1	0.1	N/A	N/A
	Other OECD	N/A	N/A	0.0	0.0
	Former Sov Union 0.3	0.6	0.0	0.0
	4.4 Comparison with actual experience: the case of Poland.

  Polish Ministry of Economic A¤airs and Labour: Poland 2005. Report-Foreign Trade.Table 3 and Charts 3-4.

	2 1 Source:	Polish	National	Statistical	O¢ ce,	http://www.stat.gov.pl./dane_spol-
	gosp/…nanse_pod_gosp/Rachunki_kwartalne_pkb_latach_1995-2004/tabela38.xls.
	2 2

The articles inMaskus and Wilson (2000) are a good survey of the limited research carried out so far on the economics of technical barriers to trade.Baldwin's (2000) article in that book summarises the view that these barriers comprise 'regulatory protection'.[START_REF] Edwards | Quality Standards under Classical Oligopoly and Trade: Regulatory Protection or Just Over-Regulation?[END_REF] gives a more cautionary view on this.

[START_REF] Zahariadis | Modeling Deep Integration in the Context of the EU-Turkey Customs Union: Working Notes[END_REF] looked at EU-Turkish integration, combining data on various costs from[START_REF] Harrison | Increased Competition and Completion of the Market in the European Union: Static and Steady State E¤ects[END_REF], Hoekman and Eby Konan (1998). Total estimated costs on Turkish imports into the EU were below 5 % for most sectors, though there is also a standardisation cost[START_REF] Harrison | Increased Competition and Completion of the Market in the European Union: Static and Steady State E¤ects[END_REF]) of 1-2.8 %

Tre ‡er (1995), McCallum (1995).

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

[START_REF] Bergstrand | The Generalised Gravity Equation, Monopolistic Competition and the Factor-Proportions Theory of International Trade[END_REF] Dixit and Stiglitz (1977).

The consumer price, P U U i;c;cc , also includes a proportional tax on use:

Exports

Consumption (…nal and intermediate) in country cc of good i produced in country c is

Sales shares

Aggregate consumer price

The total value of expenditure on good i in country c is

Factor markets

Both types of labour are immobile between countries, but mobile between industries.

The wage clears each labour market, so that Capital is fully mobile between industries, and may or not be mobile between countries:

Where KM c is non-zero (so that there are international transfers of capital) the global total of KM is set to zero.

The rate of return on capital in each industry is equated to the national rate of return,

Where capital is allowed to move internationally,

Land is sectorally immobile, so

Variety of goods

For sensitivity analysis, the …xed …rm numbers version of the model assumes the total number of …rms in each country is …xed: