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¹⁷ We discuss relaxing this assumption below.

¹⁸ The assumption that u is constant is made to simplify the mathematical expressions, but it is not necessary. The assumption that u > 1 is critical for our results. Both assumptions regarding u are supported by studies of the allocation of household savings. See Friend, Irwin and Blume, Marshall E., ”The Demand for Risky Assets,” American Economic Review 65 (December 1975Google Scholar).

¹⁹ We have made the modeling choice to represent differences in social-insurance policies in terms of differences in the distribution of benefits, rather than in terms of differences in the tax that finances the benefits. A more general approach would be to define post-tax and transfer income as a function of pre-tax and transfer income, as in Roemer, John, ”Does Democracy Engender Equality,” in Mansfield, Edward and Sisson, Richard, eds., Political Knowledge and Public Interest (Columbus: University of Ohio Press, 2003Google Scholar). For the purposes of this paper, however, the assumption of a flat tax is a reasonable approximation. In most of the countries we study, much of the welfare budget is financed by a payroll tax that is usually flat. (Denmark is an outlier in relying almost exclusively on income and value-added taxes.) Moreover, a recent study of the progressiveness of the personal income tax in twelve OECD countries by Wagstaff et al., found ”no link between pre-tax inequality and the degree of redistribution brought about by the personal income tax”; see Adam Wagstaff et al., ”Redistributive Effect, Progres-sivity and Differential Tax Treatment: Personal Income Taxes in Twelve OECD Countries,” Journal

²⁰ The assumption that income-replacement policies are redistributive, or that ξ > 0, is consistent with the way most social-insurance programs are designed. For example, the average replacement ratio for unemployment insurance, b_N(w, t)/w in the notation of the paper, is 18 percent higher for a worker who receives two-thirds of the median wage than for a worker who receives the median wage in the countries in our data set; Organization for Economic Cooperation and Development, Dataset on Benefit Replacement Rates (Paris: OECD, no date). Of course, it would be preferable if ξ > 0 was a conclusion rather than an assumption. However, the one-dimensional model of politics no longer applies when both ξ and b are chosen simultaneously. For a model where ξ is chosen at a ”constitutional” stage to maximize a social welfare function while b(i) is chosen by self-interested voters in a second ”electoral” stage, see Georges Casamatta, Helmuth Cremer, and Pierre Pestieau, ”Political Sustainability and the Design ofSocial Insurance,” Journal ofPublic Economics 75 (March 2000).

²¹ The U.S. is exceptional in devoting roughly 25 percent of public health expenditures to a family of means-tested programs known as Medicaid; United States Bureau of the Census, Statistical Abstract of the United States 2001 (Washington, D.C.: U.S. Government Printing Office, 2001).

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²⁴ Organization for Economic Cooperation and Development, Employment Outlook (July 1993); idem, Employment Outlook (July 1996).

²⁵ For a discussion of the properties of the lognormal distribution and its use as an approximation of the distribution of income, see Aitchison, J. and Brown, J. A. C., The Lognormal Distribution (Cambridge: Cambridge University Press, 1957Google Scholar).

²⁶ Both our model and the data we use to measure income inequality refer to wage and salary earners who are either working full time or are temporarily without employment. Of course, not all voters fit into these categories. Some work part time. Others are permanently outside the dependent labor force. To take all categories of attachment to the labor market into account would greatly complicate the analysis, both theoretically and empirically.

²⁷ The countries and years in the data set are listed in Appendix 2.

²⁸ The regression equation is

with R ² = 87.7 and n = 50, where y, is total welfare expenditures as a share of GDP in period t, and the standard error of the coefficient on the lagged dependent variable is .050.

²⁹ The possible endogeneity of unemployment is discussed below.

³⁰ Lijphart, Arend, ”Unequal Participation: Democracy's Unresolved Dilemma,” American Political Science Review 91 (May 1997Google Scholar); Franzese, Robert J., Macroeconomic Policies of Developed Democracies (Cambridge: Cambridge University Press, 2002CrossRefGoogle Scholar).

³¹ Roemer(fn. 15,2001).

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³³ The impact of these three variables on the distribution of wages and salaries is analyzed in Michael Wallerstein, “Wage-Setting Institutions and Pay Inequality in Advanced Industrial Societies,” American Journal of Political Science 43 (July 1999). For related studies, see Richard B. Freeman, “Labour Market Institutions and Economic Performance.” Economic Policy 3 (April 1988); Francine D. Blau and Lawrence M. Kahn, “International Differences in Male Wage Inequality: Institutions versus Market Forces,” Journal of Political Economy 106 (August 1996); and David Rueda and Jonas Pontusson, “Wage Inequality and Varieties of Capitalism,” World Politics 52 (April 2000).

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³⁵ Traxler, Franz, Blaschke, Sabine, and Kittel, Bernhard, National Labour Relations in Internationalized Markets (Oxford: Oxford University Press, 2001CrossRefGoogle Scholar). The measure of union participation in policymaking with respect to nonwage issues is described by Traxler, Blaschke, and Kittel as “associational (union) participation in state regulation (non-wage issues)” (p. 68). The data are available by decade. We assigned the 1980–90 figure to 1985 in our data set, and the 1991–96 figure to 1995 in our data set. For 1990, we used the average of the 1980–90 and 1991–95 figures. We rechecked our results with 1990 removed from our data. In neither case did the inclusion of the index of union participation alter our findings with respect to inequality.

³⁶ Wallerstein (fn. 33).

³⁷ Neal Beck and Jonathan Katz, ”What To Do (and Not to Do) with Time-Series Cross-Sectional Data in Comparative Politics,” American Political Science Review 89 (September 1995); Greene, William H., Econometric Analysis, 3rd ed. (Upper Saddle River, N.J.: Prentice Hall, 1997Google Scholar).

³⁸ We generated 400 data sets, each with 45 observations (15 countries, 3 time periods) and 3 regressors (a constant plus the first 2 regressors in column 1 of Table 2). The error terms were normally distributed with randomly selected, country-specific variances and randomly selected, cross-national correlations. In the simulations the 90 percent confidence intervals contained the true values of the coefficients roughly 90 percent of the time when calculated using OLS standard errors. When calculated on the basis of panel-corrected standard errors, by contrast, the 90 percent confidence intervals contained the true value of the coefficients only 77 percent of the time. See Karl O. Moene and Michael Wallerstein, “Income Inequality and Welfare Spending: Simulations” (http://faculty-web.at.nwu.edu/polisci/wallerstein/papers.html).

³⁹ Franklin, Mark N., “Electoral Competition,” in LeDuc, Lawrence, Niemi, Richard, and Norris, Pippa, eds., Comparing Democracies: Elections and Voting in Global Perspective (Thousand Oaks, Calif.: Sage Publications, 1996Google Scholar).

⁴⁰ In the case of health expenditures, the estimated coefficient on inequality is even closer to zero if one subtracts means-tested health expenditures (Medicaid) from the U.S. figures. Excluding U.S. Medicaid expenditures (roughly 25 percent of total government expenditures on health in the U.S.), column 3 of Table 2 becomes

where standard errors of the coefficients (excluding the constant) are (.103, .526, .0025, .0088, .052) and adjusted R ² = 65.8. Only the coefficient on the lagged dependent variable and on Right government are significant at the .05 level.

⁴¹ The category of income replacement in Table 2 is a subset of the policies included in insurance against loss of income in Moene and Wallerstein (fn. 4). The difference between the two is that the measure of insurance against income loss in Moene and Wallerstein (fn. 4) includes a share of expenditures on health while all health expenditures are excluded from spending on income replacement in Table 2.

⁴² An alternative way to measure the generosity of unemployment benefits is the replacement ratio, which is available from OECD (fn. 20). Using the average replacement ratio for a worker at the median wage and at two-thirds of the median wage in the first year of unemployment as the dependent variable yields

where standard errors of the coefficients (excluding the constant) are (.048, .054, .0003, .0010) and adjusted R ² = 90.5. All coefficients are significant at the .05 level. Neither the share of elderly in the population nor the rate of unemployment are significantly different from zero when the replacement ratio is the dependent variable.

⁴³ This procedure is advocated and given a Bayesian justification in Learner, Edward E., Specification Searches: Ad Hoc Inferences with Nonexperimental Data (New York: John Wiley and Sons, 1978Google Scholar). We did not consider the unemployment rate to be ”questionable” when the dependent variable included unemployment benefits.

⁴⁴ In lines 1, 4, 5, and 6, the minimum estimate is obtained by excluding Austria and the maximum estimate is obtained by excluding Finland. In line 2, the minimum is obtained by excluding Norway and the maximum is obtained by excluding the U.S. In line 3, the minimum is obtained by excluding Finland while the maximum is obtained by excluding Austria.

⁴⁵ Lorenzo Kristov, Peter Lindert, and Robert McClelland, ”Pressure Groups and Redistribution,” Journal ofPublic Economics 48 (July 1992).

⁴⁶ Iversen, Torben and Soskice, David, ”An Asset Theory of Social Policy Preferences,” American Political Science Review 95 (December 2001Google Scholar).

⁴⁷ Iversen and Soskice (fn. 46), for example, report a correlation coefficient of .73 between their measure of the extent of vocational training and the W¹⁰/W⁹⁰ ratio (p. 889).

⁴⁸ Daron Acemoglu and Jorn-Steffen Pischke, ”The Structure of Wages and Investment in General Training,” Journalof PoliticalEconomy 107 (June 1999).

⁴⁹ Moffitt, Ribar, and Wilhelm (fn. 7).

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⁵¹ Kenworthy calculates the share of individuals in advanced industrial societies who would be classified as living in poverty in the U.S., that is, living in households with incomes less than 40 percent of the median household income in the U.S. after converting their household income to U.S. dollars according to purchasing power parity and adjusting for family size; see Lane Kenworthy, ”Do Social Welfare Policies Reduce Poverty? A Cross-National Assessment,” Social Forces 77 (March 1999). The partial correlation coefficient between share living in poverty and the log of the 90/10 wage ratio is .69, controlling for GDP per capita for the fourteen countries where Kenworthy's sample overlaps with the sample of this paper.

⁵² OECD (fn.23).

⁵³ United States Bureau of the Census, Statistical Abstract of the United States 1990 (Washington, D.C.: U.S. Government Printing Office, 1990Google Scholar); idem (fn. 21).

⁵⁴ OECD (fn. 24,1996); OECD (fn. 24,1993) in the case of the U.S.

⁵⁵ Swank (fn. 1).

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