Adams, C. and D. Coe, “A Systems Approach to Estimating the Natural Rate of Unemployment and Potential Output for the United States,” Staff Papers, International Monetary Fund (Washington: June 1990).
Amano, Robert A., and R. Tiff Macklem, “Menu Costs, Relative Prices and Inflation: Evidence for Canada,” Manuscript, Research Department, Bank of Canada (October 1995).
Barro, R. and M. Rush, “Unanticipated Money and Economic Activity,” in S. Fischer (ed.) Rational Expectations and Economic Policy (Chicago: University of Chicago Press 1980).
Blomberg, S. Brock, and Ethan S. Harris, “The Commodity-Consumer Price Connection: Fact or Fable? Federal Reserve Bank of New York Economic Policy Review. Vol. 1 (October 1995), pp. 21–38.
Clark, P., D. Laxton, and D. Rose, “Asymmetry in the U.S. Output-Inflation Nexus,” Staff Papers. International Monetary Fund (Washington: March 1996).
Coe, D. and R. Moghadam, “Capital and Trade as Engines of Growth in France,” Staff Papers, International Monetary Fund (Washington: September 1993).
Duguay, Pierre, “Empirical Evidence on the Strength of the Monetary Transmission Mechanism in Canada,” Journal of Monetary Economics. Vol. 33 (February 1994), pp. 39–61.
Fortin, Pierre, “The Unbearable Lightness of Zero-Inflation Optimism,” Canadian Business Economics, Vol. 1 (Spring 1993), pp. 3–18.
Hostland, Doug, “Changes in the Inflation Process in Canada: Evidence and Implications,” Bank of Canada Working Paper 95-5 (May 1995).
King, R. and M. Watson, “The Post-war U.S. Phillips Curve: A Revisionist Econometric History” WP 94-14, Federal Reserve Bank of Chicago.
Laxton, D., Meredith, G. and D. Rose, “Asymmetric Effects of Economic Activity on Inflation,” Staff Papers, International Monetary Fund (Washington: June 1995).
Layard, R., Nickell, S. and R. Jackman, Unemployment: Macroeconomic Performance and the Labour Market. (Oxford: Oxford University Press 1991).
Mackinnon, J., “Approximate Asymptotic Distribution Functions for Unit-Root and Cointegration Tests,” Journal of Business and Economic Statistics, (April 1994).
Nickell, S. and B. Bell, “Would Cutting Payroll-Taxes on the Unskilled Have a Significant Impact on Unemployment?” Centre for Economic Performance Working Paper No. 276 (February 1996).
Pantula, S., G. Gonzalez-Farias, and W.A. Fuller, “A Comparison of Unit-Root Test Criteria,” Journal of Business and Economic Statistics, (October 1994).
Prasad, E., “Inflation and the Business Cycle,” United States - Background Papers (SM/93/183, August 16, 1993), Chapter III, pp. 26–34.
Prasad, E., “Indicators of Economic Slack,” United States - Background Papers (SM/94/223, August 18, 1994), Chapter III, pp. 27–30.
Prasad, E. and A. Thomas, “Aggregate and Provincial Labor Market Adjustment in Canada,” (mimeographed, International Monetary Fund, July 1995).
Santaella, J., “The Output Gap and Inflation,” United Kingdom - Recent Economic Developments (SM/95/276, October 18, 1995), Chapter V, pp. 79–90.
Staiger, D., Stock, J. and M. Watson, “How Precise are Estimates of the Natural Rate of Unemployment?” (mimeographed, Harvard University, February 1996).
Thomas, Alun H., “Estimates of the Structural Rate of Unemployment,” Canada - Recent Developments and Policies (SM/95/181, April 20, 1995), Chapter III, pp. 11–17.
Some recent papers have argued for the existence of nonlinearities in the relationship between inflation and the output gap. For example, see Chapter VI of the Background Papers for the 1995 Article IV consultation with the United States (SM/95/181, July 26, 1995), Laxton et al. (1995), and Clark et al. (1996). Such nonlinear specifications are not addressed in this paper.
See “Inflation and the Business Cycle” in SM/93/183 and “Indicators of Economic Slack” in SM/94/233 for a more detailed discussion of these methods.
See the Economic Report of the President (1996).
The NBER defines April 1958 as a business-cycle trough and April 1960 as the corresponding business-cycle peak.
The HP filter fits a curve through the data, thereby allowing changes in the trend. However, it is estimated in a manner that seeks to minimize frequent changes in the trend.
Available data on the capital stock do not incorporate the most recent revisions to the national income and product accounts. Hence, there could be biases in the estimates of total factor productivity and potential GDP depending on the scope of the revisions to the capital-stock data. See Coe and Moghadam (1993) for the use of this approach on French data.
The 1974 break point is consistent with the analysis of the Council of Economic Advisors (1996) and corresponds to the first oil price shock. Other determinants of total factor productivity, such as the share of youth in the labor force and the ratio of exports and imports to GDP were not found to add significant explanatory power in the regression equation.
See “Estimates of the Structural Rate of Unemployment” in SM/95/81 for a similar approach.
Other studies have also included a measure of the tax wedge as a determinant of the natural rate of unemployment (see, for example Layard et al. (1991) and Adams and Coe (1990)). However, in the long run, wages are usually thought to adjust fully to payroll taxes and the effect on employment would be minimal. The tax wedge also could affect employment if wages at the bottom end of the wage distribution are not flexible because of minimum wage laws and interactions with the benefit system. This qualification may be relevant for European countries, which have generous benefit systems and relatively high minimum wages but should not be an issue in the United States. Moreover, Nickell and Ball (1996) have found that the tax wedge is not significant in a long-run cointegrating relationship between various structural variables and unemployment in the United Kingdom.
In a recent review of the relationship between real wages and the business cycle, Abraham and Haltiwanger (1995) argue that labor demand shocks play the major role in accounting for the positive cyclical comovement in real wages and employment since the early 1970s. The identification of employment shocks as labor demand shocks also underlies recent analyses of the regional and aggregate labor markets in the United States and Canada by Blanchard and Katz (1992) and Prasad and Thomas (1996).
The weighted symmetric r test-statistics in Table 1 indicate that the null hypothesis is accepted and that all series are I(1) except for the output and wage gaps and the relative minimum wage. However, the relative minimum wage is I(1) using the Dickey-Fuller T test-statistic, and the output and wage gaps are borderline stationary in levels.
The replacement rate—defined as the ratio of weekly unemployment benefit to the average weekly wage—also was included in the analysis, but this variable had the wrong sign in all cointegrating vectors.
Estimating the cointegrating relationship by OLS is appropriate if it is assumed that the error in the cointegrating equation is uncorrelated with the errors in regressions of the explanatory variables on their own lags. See Hamilton (1994) pp. 602-603 for more details.
The long-run employment-population and participation rates are obtained by setting the output and wage gaps equal to zero in the estimated equations.
The Delta method assumes that the explanatory variables are fixed so that the coefficient estimates provide the only source of uncertainty in the estimate. The Gaussian method is an alternative method of generating standard errors in which errors are randomly sampled with replacement and used to generate artificial draws. Staiger et al. (1996) find that in Monte Carlo simulations of both methods, the confidence intervals based on the approximating distribution are close to the confidence intervals for the true distribution. Moreover, for alternatives near the null, both methods have comparable size-adjusted power. However, for more distant alternatives, the Gaussian method has substantially greater power.
In their U.S. Economic Outlook. Laurence H. Meyer and Associates estimate that the loss in employment resulting from the rise in the minimum wage would be 200,000, rising to 330,000, if account is taken of a possible effect on workers currently earning wages slightly above the minimum. Under this latter scenario, workers currently earning between $4.25 and $6.00 receive wage increases ranging from 21 percent (for those currently at the minimum) through progressively smaller increases. The wages of workers earning more than $6.00 are assumed to be unaffected.
The inflation rate for the CPI excluding food and energy is analyzed in this paper. The starting date for the analysis is 1971 because data on unionization is only available from this date onward.
Empirical work on the determinants of inflation is vast. For the United States, a number of authors have focussed on the unemployment gap as the measure of the business cycle. For example, Sargent (1976) and Barro and Rush (1980) consider the relationship between the unemployment rate and unexpected movements in prices and money respectively, whereas King and Watson (1994) analyze structural vector autoregressions explaining the change in the unemployment rate and inflation. For Canada, Duguay (1994) and Amano and Macklem (1995) use the output gap as their measure of cyclical conditions and add real oil prices and an asymmetric price variable to their respective specifications. Ericson and Brouwer (1995) explain inflation in Australia on the basis of changes in unit labor costs, import prices, and the price of oil, adding the output gap to proxy cyclical movements in the markup.
The basis for this derivation is that the economy is driven by shocks to aggregate demand which trace out an aggregate supply function i.e., firms increase output and require more labor when the price rises.
Quarterly data are used to estimate the equation and four-quarter differences are taken because some of the data used later in the analysis are only available on a seasonally unadjusted basis. Dickey-Fuller tests indicate that both the inflation rate and the unemployment gap are stationary (see Table 5).
The precision of the out-of-sample forecasts using the standard Phillips curve specification is sensitive to the initial starting point. When out-of-sample forecasts begin in the first quarter of 1993, the standard Phillips curve specification underpredicts inflation. This is because in 1993 the economy was operating below potential, putting downward pressure on inflation.
The additional cost variables were subjected to Weighted Symmetric and Dickey-Fuller τ tests to determine whether they were stationary. Table 5 presents these test-statistics using the log level and the rate of change over 4 quarters for each variable. The table indicates that all price variables are borderline stationary in rates of change and that productivity is also stationary in log levels. All variables were analyzed in rates of change in order to maintain consistency.
Petroleum and farm prices are included because they are assumed to be used as inputs for other products.
Both the Schwarz and Akaike criteria minimize the value of the determinant of the covariance matrix of the residuals but impose different penalties on increasing the number of estimated parameters (the Schwarz criterion is considerably more stringent).
In the long run, the inflation rate is constant and, given that the sum of the inflation coefficients is close to zero, the other variables must be expressed in differences so that they also sum to zero (assuming that the levels are non-zero). This can only be achieved if two lags of each variable are included. This requirement does not apply to the unemployment gap because its level is zero in the long run. However, two lags were adopted to maintain consistency with the other variables.