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David T. Coe is a Senior Economist in the Current Studies Division of the Research Department. He is a graduate of the University of Maryland and the London School of Economics and holds a doctorate from the University of Michigan. This paper was written when he was in the North American Division of the Western Hemisphere Department while on leave of absence from the Organization for Economic Cooperation and Development.
See, for example. Bean. Layard, and Nickell (1986) and Layard and Nickell (1987). For an international comparison of developments in nonwage labor costs as a percentage of wages and salaries since the mid-1960s, see Table 5 on page 63 of Chan-Lee, Coe, and Prywes (1987).
The following discussion, which is based on Green and Cousineau (1976) and Ashenfelter and Card (1986), only concerns changes in the unemployment insurance (UI) system between 1971 and 1988. In the spring of 1989 the Canadian Government announced changes to the UI system, including some tightening of eligibility requirements and a shortening of benefit periods.
The unemployment insurance system is administered in 48 regions, rather than the 10 provinces used in the calculations presented here.
If either the difference between, or the ratio of, MAXB and MINQ is used as the adjustment factor, the estimation results are very similar to those reported below.
Alternative approaches to estimating the natural rate of unemployment are discussed in Adams and Coe (1989).
When alternative lag specifications gave conflicting signals as to the significance of specific variables—for example, because of multicollinearity between the relatively large number of explanatory variables—preference was given to keeping structural variables in the estimated equations.
Variables expressed as the logarithms of ratios have been multiplied by 100, so that all variables are entered in what are effectively percentage terms, UIRR and TAXSIP are entered as two-period moving averages. Data sources are as follows: The unemployment insurance replacement rate adjusted for coverage, minimum wages, and commercial wages is from the Bank of Canada. The average rate for employers’ contributions (for social security and pension funds), expressed as a percent of total wages and salaries, is from the Organization for Economic Cooperation and Development’s Standardized National Accounts and is interpolated from annual data. The percentage of the labor force which is unionized is from Labour Canada’s annual Directory of Labour Organisations in Canada (Ottawa) and is interpolated from annual data; the 1979 observation is not available and was set equal to the average of 1978 and 1980. All other data are from Statistics Canada’s CANSIM data base.
That is, the lag distribution initially resembles a third-order polynomial, with the peak in the third quarter, before asymptotically approaching zero.
When the squares of the quadratic-gap variable were added to the equation, in order to test for nonlinearities, its estimated coefficient was marginally significant, but there was little difference in the overall estimation results. Using a linear trend with a break in 1973, rather than a quadratic trend, also had little overall effect, although the significance of the estimated coefficient on the unemployment replacement rate was reduced somewhat.
Entering separate tax rates for employers’ contributions to social security and to pension funds resulted in significant estimated coefficients of almost identical magnitude on each variable, but had little impact on the overall regression results. The significance, but not the sign, of the estimated coefficients on relative minimum wages and the replacement rate was sensitive to the inclusion of the employers’-contribution variable.
The smaller estimated coefficient on the UIRR variable in equation (1), compared with those in equations (2)-(4), reflects the fact that the adjustment factor for qualifying and benefit periods increases the magnitude and variance of the UIRR variable. (Compare Figure 2.)
Increased dispersion across industries may be a proxy for occupational mobility tending to decrease the natural rate, whereas increased dispersion across provinces may be a proxy for structural changes tending to increase the natural rate.
See Rose (1988) and the references cited therein. The higher estimate in Figure 3 is based on sample-period averages for the cyclical and supply-shock variables. If the average relative price of energy during 1979-88 were used instead of the sample-period average, the estimate of the natural rate would be correspondingly higher.
This estimate is similar to the 0.8 percentage point estimate of Grube 1, Maki. and Sax (1975, p. 187).
The estimation results suggest, for example, that each of the following policies might contribute to lowering the natural rate by about ½ of 1 percentage point during 1989—93: continued gradual declines in relative minimum wages at the same rate as occurred, on average, from the mid-1970s to 1988; the elimination of regional extended benefits for unemployment insurance benefits; or the gradual reduction in payroll taxes to about their 1982 levels.
The proposed reforms to the UI system, announced in early 1989, reduce extended original benefits somewhat, but only in regions with low unemployment rates.