labor market structure (changes in the rate of unemployment, 8 unemp, female participation, femlab) where significant, bear the expected signs.

As income per-capita rises so does the share of contributors to the social security system. Higher rates of life-expectancy also increase the share of workers who contribute. Changes in the rate of unemployment capture cyclical effects, and show a fall in the share of contributors as the number of people who find themselves without work rises. The share of women in the labor force has no significant effect to the share of contributors in either the random or fixed effects estimation. The dummies for the t - I years in the time series, included to control for time-specific effects but omitted from Table 3, are all significant at the 1% level in both estimations.

As expected from the theoretical discussion in Section II, the estimated coefficient on the social security tax variable is negative and significant (at 10% using random effects, and 1% in fixed effects). In both the random and the fixed effects regressions the social security tax variable bears the largest coefficient in the equation. The fixed effect3 coefficient on the variable measuring the extent of structural reform to social security, [RA - the portion of the total payroll tax for social insurance accumulating in a privately - managed individual retirement account, rather than pooling in a public PAYGO system is positive and highly significant (at 5%). While the random effects coefficient on IRA is not statistically significantly different from zero, it bears the expected sign.

Tro capture a possible lag in the reaction of the labor market to reformns - either because of uncertainty due to the dramatic changes they entail, or a learning effect as agents became familiar with the set of new social security institutions - I experimented with several different polynomials on the years-since-reform variable, YR. A third-order polynomial was preferred by the data. The signs and size of the coefficients suggest a ".1 curve" effect in the share of workers contributing after a reform (in the random and the fixed effects estimations, the coefficients on YR, YR<sup>2</sup> and YR<sup>3</sup> were all significant at 1%).11 A glance at the plotted dependent variable in Figure 4.a. also suggests such an effect, as the share of contributors initially dips after the introduction of individual retirement accounts, begins to rise in the second or third year, and then gradually levels. Rteaders will note that this effect is most pronounced in the Chile data. Therefore, to test whether the pattern of contribution was different after the reform in Chile than in other reforming countries, I interacted a Chile dummy variable with each of the polynomial terms and included these in both the random and fixed effects regressions. Testing for the joint significance of the coefficients on the Chile interacted variables, I was unable to reject the null hypothesis that these are significantly different from zero in either

> **1** I experimented with several alternatives to estimate transition dynamics after reforms, including using lags of the dependent variable and the policy (sstax and ira) variables. The lagged variables where nosignificant in either random or fixed effects, both when the year since reform polynomials were left in the regression and when they where removed. With respect to sstax and ira, the insignificance of the lagged terms may reflect the fact that for each country the variables changes only once, at the time of reform. T hus, in each case applying a lagg operator generates a variable that varies from the original variable for only seven out of 287 observations. Further, the relatively small size of the panel's cross-section - 18 countries - dissuaded me from using a GMM (generalized method of moments) estimator, since this would have introduced more biases to the model.

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