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An Exclusive Country Club: The Effects of the GATT on Trade, 1950–94

Published online by Cambridge University Press:  13 June 2011

Joanne Gowa  and
Soo Yeon Kim
Joanne Gowa
Affiliation:
Princeton University
Soo Yeon Kim
Affiliation:
University of Maryland
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Abstract

Using data on bilateral trade flows from both before and after World War II, this article examines the impact of the General Agreement on Tariffs and Trade on trade between its members and on the system of interwar trade blocs. It shows that the distribution of the benefits produced by the GATT was much more highly skewed than conventional wisdom assumes. The article also shows that the gold, Commonwealth, Reichsmark, and exchange-control blocs exerted positive and significant effects on trade after 1945. The authors attribute these effects to the bargaining protocol that governed successive rounds of GATT negotiations, the signature element of the postwar trade regime.

Type
Research Article
Information
World Politics , Volume 57 , Issue 4 , July 2005 , pp. 453 - 478
Copyright
Copyright © Trustees of Princeton University 2005

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References

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26 Leddy (fn. 17), 37. A heated debate exists about the RTAA and the trade preferences of presidents relative to Congress. See, e.g., Lohmann, Susanne and O'Halloran, Sharyn, “Divided Government and U.S. Trade Policy: Theory and Evidence,” International Organization 48 (Autumn 1994)CrossRefGoogle Scholar; Hiscox, Michael, “The Magic Bullet? The RTAA, Institutional Reform, and Trade Liberalization,” International Organization 53 (Autumn 1999)CrossRefGoogle Scholar; Bailey, Michael A., Goldstein, Judith, and Weingast, Barry R., “The Institutional Roots of American Trade Policy: Politics, Coalitions, and International Trade,” World Politics 49 (April 1997)CrossRefGoogle Scholar. We do not engage this debate here, since we are concerned about only this one instance of executive-congressional interaction.

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29 Leddy (fn. 17), 21.

30 More generally, Article XXIV sanctioned preferential trading agreements (PTAS) as long as they were not “more restrictive” than they had been previously. For the GATT text, see http://pacific.commerce.ubc.ca/trade/GATT. 31Snnivasan, T. N., Developing Countries and the Multilateral Trading System: From the GATT to the Uruguay Round and the Future (Boulder, Colo.: Westview, 1998), 10.Google Scholar

32 U.S. Tariff Commission (fn. 14), 65; U.S. Tariff Commission, Operation of the Trade Agreements Program, 9th Report, July 1955-June 1956 (Washington, D.C.: Government Printing Office, 1957), 65.

33 Havana Charter, chap. 4, Art. 17.2(a). For the Havana Charter text, see www.globefield.com/havana.

34 Ibid., Article 28.2(a) (bis).

35 In the 1950s, for example, several countries sought to impose maximum tariffs on broad product groups instead of continuing with reciprocal trade concessions. About the same time, a majority of GATT contracting parties, including Belgium, Denmark, France, Germany, and the Netherlands, proposed roughly 30 percent cuts on average tariffs in ten sectors. U.S. allegiance to the “product-by-product and country-by-country” approach, however, left the status quo intact. Kock, Karen, Intemational Trade Policy and the GATT, 1947-1967 (Stockholm: Almqvist and Wiksell, 1969), 98Google Scholar; see also Dam (fn. 7), 67.

36 The Trade Expansion Act of 1962 authorized the use of a linear approach. For discussion, see Dam (fn. 7), 67.

37 Ibid., 76. The sectors were chemicals, cotton textiles, iron and steel, aluminum, and pulp and paper.

38 Curzon, Gerard and Curzon, V., “The Management of Trade Relations in the GATT,” in Shonfield, Andrew, ed., International Economic Relations of the Western World, 1959-1971, vol. 1, Politics and Trade (London: Oxford University, 1976), 185-86Google Scholar.

39 Hoekman and Kostecki (fn. 24), 130.

40 U.S. Tariff Commission (fn. 32), 140, 125. Of the 1,014 tariff reductions that the United States agreed to between the wars, almost half involved tariff reclassifications. Verdier, Daniel, Democracy and International Trade: Britain, France, and the United States, 1860-1990 (Princeton: Princeton University Press, 1994), 196.Google Scholar41Bidwell (fn. 22), 54, emphasis in original. The growth of the U.S. tariff code attests to the effort to specialize tariffs. In 1963 it included 6,421 lines. By 2000 it had grown to 10,175 lines, even though the U.S. trade-weighted average tariff rate fell from 11.9 percent to 7.4 percent ad valorem in the interim. U.S. Tariff Commission, Simplification of the Harmonized Tariff Schedule of the United States (Washington, D.C.: Government Printing Office, 2000)Google Scholar.

42 Ibid., 125.

43 Hoekman and Kostecki (fn. 24), 31.

44 Ibid. For the internalization ratios for each GATT round though 1967, see Finger, J. M., “Trade Liberalization: A Public Choice Perspective,” in Amacher, Ryan C., Haberler, Gottfried, and Willett, Thomas D., eds., Challenges to a Liberal International Economic Order (Washington, D.C.: American Enterprise Institute, 1979), 424.Google Scholar

45 Ibid., 42.

46 CPI for Urban Consumers, ail items; 1982-84=100; www.economy.com/freelunch.

47 Tomz, Goldstein, and Rivers (fn. 4, 2004a).

48 That is, a state is a system member if it is a UN member or its population exceedsfivehundred thousand and it receives diplomatic missions from at least two major powers. Correlates of War 2 Project. 2003. State System Membership List, v2002.1, http://cow2.la.psu.edu.

49 The analysis utilizes directed dyads composed of these countries, for various years depending on system membership and data availability.

50 See fn.4.

51 Tomz, Goldstein, and Rivers (fn. 4,2004a), 4.

52 To construct their roster of “nonmember participants,”Tomz, Goldstein, and Rivers (fn. 4,2004a) use GATT archival material that is not yet publicly available. They have not responded to our request to allow us to use their data. We cannot be certain, therefore, that our roster of nonmember participants exactly replicates theirs. This is so, because, for example, while they note that “the maximum allowable duration of de facto status changed over time,” their papers do not define exactly how it did so. Similarly, they state that not all members accorded provisional members MFN treatment, but, with one exception, they do not identify these countries (pp. 7-8).

53 The quotation is from Rose (fn. 4).

54 For details, see www.faculty.haas.berkeley.edu/arose.

55 Complete results for any finding we report in the paper are available from the authors.

56 To be more specific, the GATT coefficient in Rose's OLS analysis is a statistically insignificant -0.04, (fn. 4), 104. Including dyadic fixed effects produces a statistically significant estimate of 0.15 (fn. 4), 104. We used Rose's data and variable definitions to estimate GATT'S impact between 1950 and 1994, the years we analyze here. An OLS analysis with year fixed effects produces a coefficient on GATT of 0.116 (p-value = 0.004). An analysis with dyadic and year fixed effects produces a coefficient on GATT of 0.046 (p-value = 0.150). In Subramanian and Wei (fn. 4), the GATT coefficient that an analysis with country and year fixed effects produces is -0.113, which is statistically significant at less than the 0.05 level (p. 28). Their analysis is difficult to compare directly to ours, as their regressand is the log of real imports of each state and they include data through 2000. Goldstein, Rivers, and Tomz use data from 1948 through 2001 and report a statistically significant estimate on GATT, using a model with dyadic and year fixed effects, of 0.37 (fn. 4,2004b), 26.

57 Green, Donald P., Kim, Soo Yeon, and Yoon, David H., “Dirty Pool,” International Organization 55 (Spring 2001)CrossRefGoogle Scholar.

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59 The results in Table 1 are robust to the inclusion of country fixed effects, however.

60 Since the data are in the form of directed dyads, dyadic fixed effects are specified separately for each direction. Based on a likelihood ratio test, Cheng and Wall find that a symmetry restriction on the dyadicfixedeffects rejects the null (that is, that fixed effects do not differ significantly between the dyad that reflects the imports of a from b and the dyad reflecting imports of b from a). Thus, they argue for including a separate term for each direction of the dyad. See I-Hui Cheng and Howard J. Wall, “Controlling for Heterogeneity in Gravity Models of Trade and Integration,” Federal Reserve Bank of St. Louis Working Paper 1999-010E (2004).

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62 In comparing these results with those in the existing literature, however, it is important to note that our analyses do not include Soviet bloc members.

63 As reported in Subramanian and Wei (fn. 4), 9. Belgium and Luxembourg are not in our analysis, however, because the IMF reports their trade separately only in 1997.

64 These include dyads where both members are nonindustrial countries or where one is an industrial country and the other a nonindustrial country.

65 The coefficient on industrial-country trade that Rose (fn. 4) reports is a statistically significant 0.47 (p. 108). The corresponding statistic in Subramanian and Wei (fn. 4) is 0.322, also statistically significant (p. 29),

66 We report Newey-West standard errors applied to panel data, which are robust to heteroskedas-ticity and autocorrelation. Newey, W. and West, K., “A Simple Positive Semi-Definite, Hetersoscedasticity and Autocorrelation Consistent Covariance Matrix,” Econometrica 55 (May 1987)CrossRefGoogle Scholar.

67 This is also true, of course, of members of other dyads that either joined the GATT at its inception or were allies or democracies throughout the sample period.

68 The thirteen remaining industrial countries—Austria, Denmark, Finland, Germany, Greece, Iceland, Ireland, Italy, Japan, Portugal, Spain, Sweden, and Switzerland—acceded to GATT between 1951 and 1967.

69 We are grateful to Albrecht Ritschl and Nikolaus Wolf for making the data available to us. Iceland is not included in the Ritschl and Wolf data, so we exclude it from the analysis. The following states are in the Ritschl-Wolf data set but not in ours: Argentina, Belgium, Turkey, and several members of the former Eastern bloc (that is, Bulgaria, Czechoslovakia, Hungary, Poland, Romania, the Soviet Union, and Yugoslavia). Albrecht Ritschl and Nikolaus Wolf, “Endogeneity of Currency Areas and Trade Blocs: Evidence from the Inter-War Period” (Manuscript, Humboldt University 2003).

70 Panel data analysis can accommodate time-invariant variables using a technique that Hausman and Taylor developed; Hausman, Jerry A.. and Taylor, William E., “Panel Data and Unobservable Effects,” Econometrica 49 (November 1981)CrossRefGoogle Scholar. However, their method assumes that not all explanatory variables are correlated with the fixed effects, an assumption that is not tenable here. The time-invariant nature of these variables also precludes an estimation based on first-differencing the data to account for the possible endogeneity of the GATT regime. Baier, Scott L. and Berg-strand, Jeffrey H., “Do Free Trade Agreements Actually Increase Members’ International Trade?” (Manuscript, 2005)Google Scholar.

71 Ritschl, and Wolf, (fn. 69); Nitsch, Volker, “Honey, I Shrunk the Currency Union Effect on Trade,” World Economy 25, no. 4 (2002)Google Scholar; Persson, Torsten, “Currency Unions and Trade: How Large Is the Treatment Effect?” Economic Policy 33 (October 2001)Google Scholar.

72 Blundell, Richard and Dias, Monica Costa, “Evaluation Methods for Non-Experimental Data,” Fiscal Studies 21, no. 4 (2000)Google Scholar.

73 The privileged group and the founding-member group are mutually exclusive. The former consists of pairs of privileged-group members (Britain, Canada, France, Germany, and the United States). The latter consists of country pairs formed among founding members that were not in the privileged group, as well as founding members paired with each member of the privileged group.

74 Ritschl and Wolf (fn. 69), 9.

75 Our post-World War II data begin in 1950; hence, we use this as the first of the GATT years.

76 An F-test for all the interaction terms (for GDP and group terms) is F(10, 17356) = 20.04 (Prob > F = 0.0000), which shows that all the interaction terms are jointly statistically significant. An F-test for all the interwar bloc variables (main and interaction effects) is F(1O,17356) = 29.80 (Prob > F = 0.0000), which indicates that the effects of the control groups and their corresponding interaction terms are jointly statistically significant.

77 The MID variable is insignificant. Joint democracy exerts a statistically significant impact on trade, increasing it by about 17 percent. 78These results are robust to the inclusion of country or dyadic fixed effects for countries that are not members of any of the groups in the analysis. In our sample, these countries are Greece and Japan.

79 We did not include the European Community in this analysis, because it did not come into existence until 1958 and because its membership did not remain stable over the postwar period. We have, however, tested whether our results are robust to the inclusion of the EC6, since it formed relatively early in the postwar period. We find that the EC had a large positive (0.84) and significant (p-value = 0.000) impact on its members’ trade. Our results are robust to including the EC with these exceptions: the treatment effect for the gold bloc becomes insignificant (p-value = 0.772); the coefficient on the interwar alliance term becomes significant (p-value = 0.008), while its treatment effect becomes insignificant.

80 Eichengreen and Irwin used interwar and early postwar trade to predict trade in 1949,1954, and 1964; Eichengreen, Barry and Irwin, Douglas A., “Trade Blocs, Currency Blocs and the Reorientation of World Trade in the 1930s,” Journal of International Economics 38 (February 1995)CrossRefGoogle Scholar; and idem, , “The Role of History in Bilateral Trade Flows,” in Frankel, Jeffrey, ed., The Regionalization of the World Economy (Chicago: University of Chicago Press, 1998)Google Scholar.

81 Our sample includes, however, only two Reichsmark bloc members (Austria and Germany). 82To calculate these values we sum the coefficients of the control and treatment effect variables (that is, the group dummy and its corresponding interaction term). The corresponding significance level is based on F-tests for the joint significance of each group's control and treatment effect estimates. They are statistically significant for all groups, except for founding members: F(2,17356) = 1.85 (Prob > F = 0.152); privileged group: F(2,17356) = 121.52 (Prob > F = 0.000); exchange-control bloc: F(2,17356) = 5.78 (Prob > F = 0.003); gold bloc: F(2,17356) = 11.85 (Prob > F = 0.000); Commonwealth bloc: F(2,17356) = 62.00 (Prob > F = 0.0000); sterling bloc: F(2,17356) = 35.71 (Prob>F =0.000); and Reichsmark bloc: F(2,17356) = 107.71 (Prob > F = 0.000).

83 To ensure that these results are not due to the abnormally low levels of trade during the Great Depression, we reanalyzed the data using only 1928 trade as the reference level. The results are robust, with these exceptions: the coefficient on alliances becomes positive and statistically significant for the interwar period (p-value = 0.048); its treatment effect is insignificant (p-value = 0.960).

84 Ritschl and Wolf (fn. 69), 11.

85 The Ritschl and Wolf results (fn. 69) are sensitive to model specification. When they control for trade diversion, they find that trade within the Commonwealth bloc, for example, increases significantly after its formation (p. 24). Eichengreen and Irwin (fn. 80,1995) find that the formation of the Commonwealth and Reichsmark blocs exerted a positive and significant effect on the trade of its members, while the formation of the currency blocs did not (p. 15).