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Comparative Political Economy of Wage Distribution: The Role of Partisanship and Labour Market Institutions

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Through a pooled cross-section time-series analysis of the determinants of wage inequality in sixteen OECD countries from 1973 to 1995, we explore how political-institutional variables affect the upper and lower halves of the wage distribution. Our regression results indicate that unionization, centralization of wage bargaining and public-sector employment primarily affect the distribution of wages by boosting the relative position of unskilled workers, while the egalitarian effects of Left government operate at the upper end of the wage hierarchy, holding back the wage growth of well-paid workers. Further analysis shows that the differential effects of government partisanship are contingent on wage-bargaining centralization: in decentralized bargaining systems, Left government is associated with compression of both halves of the wage distribution.
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B.J.Pol.S. 32, 281–308 Copyright 2002 Cambridge University Press
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Comparative Political Economy of Wage
Distribution: The Role of Partisanship and
Labour Market Institutions
JONAS PONTUSSON, DAVID RUEDA AND CHRISTOPHER R.
WAY*
Through a pooled cross-section time-series analysis of the determinants of wage inequality in
sixteen OECD countries from 1973 to 1995, we explore how political-institutional variables
affect the upper and lower halves of the wage distribution. Our regression results indicate that
unionization, centralization of wage bargaining and public-sector employment primarily affect
the distribution of wages by boosting the relative position of unskilled workers, while the
egalitarian effects of Left government operate at the upper end of the wage hierarchy, holding
back the wage growth of well-paid workers. Further analysis shows that the differential effects
of government partisanship are contingent on wage-bargaining centralization: in decentralized
bargaining systems, Left government is associated with compression of both halves of the wage
distribution.
It is well known that wage inequality has increased dramatically in the United
States over the last three decades. From 1973 to 1998, the hourly earnings of
a full-time worker in the ninetieth percentile of the American earnings
distribution (someone whose earnings exceeded those of 90 per cent of all
workers) relative to a worker in the tenth percentile grew by 25 per cent, and
the corresponding figure for men only was nearly 40 per cent. As we document
in this article, wage inequality has increased in most Organization for Economic
Co-operation and Development (OECD) countries, but the extent of this
phenomenon varies a great deal, and cross-national differences in levels of wage
inequality remain as great as they were in the 1970s. In the United States, the
worker in the ninetieth percentile earned 4.63 times as much as the worker in
the tenth percentile in 1996. At the other end of the cross-national spectrum, the
90–10 ratio for Sweden was only 2.27.
While political commentators in Europe and the United States alike
frequently invoke inegalitarian labour market trends to explain various
manifestations of working-class political disaffection (not only support for
right-wing populist parties, but also falling turnout among working-class
* Department of Government, Cornell University. For helpful comments and suggestions, we
thank James Alt, Rob Franzese, Richard Freeman, Larry Kahn, Michael Wallerstein and three
anonymous referees.
282 PONTUSSON,RUEDA AND WAY
voters), recent work by labour economists demonstrates that supply and demand
factors alone cannot account for cross-national variation in wage inequality.
1
Wage inequality appears to have political determinants as well as political
consequences. On both counts, it deserves to be a central concern of comparative
political economy as conceived and practised by political scientists. Within
political science, however, the paucity of research on wage inequality stands in
sharp contrast to the large number of quantitative studies that take various
measures of macroeconomic performance or government spending as their
dependent variable. Drawing on a new dataset published by the OECD,
2
which
enables us to engage in a pooled cross-section time-series analysis of the
determinants of wage inequality in sixteen OECD countries for the period
1973–95, we seek to make up for some of this neglect.
3
From the perspective of comparative political economy, the manner in which
labour economists deal with cross-national differences in government policy
and the organization of wage bargaining leaves something to be desired. Most
commonly, rather vaguely specified institutional factors are invoked to explain
whatever variance remains when the effects of supply and demand have been
taken into account. And when economists incorporate cross-national differences
into their models, they typically reduce these differences to a single,
one-dimensional variable, such as wage-bargaining centralization.
Our analysis demonstrates that it is possible to distinguish discrete effects of
several political-institutional variables. Controlling for certain supply and
demand conditions as well as country-specific fixed effects, we find that
bargaining centralization, public sector employment, union density and left
government are all negatively associated with overall wage inequality,
measured by 90–10 ratios. In other words, higher values on these independent
variables are associated with lower 90–10 ratios. While bargaining centraliza-
tion and the size of the public sector can be characterized as ‘institutional’ or
‘structural’ features of the political economy, union density and government
partisanship pertain to the distribution of power between labour and employers,
and also to the distribution of power among different categories of wage-earners.
Institutions matter for the distribution of wages, and so do politics.
To get a better handle on the causal mechanisms at work, we engage in
separate regression analyses of the determinants of the ratio of earning in the
1For example, Richard Freeman and Lawrence Katz, eds, Differences and Changes in Wage
Structures (Chicago: The University of Chicago Press, 1995), pp. 1–24; Francine Blau and Lawrence
Kahn, ‘International Differences in Male Wage Inequality’, Journal of Political Economy, 104
(1996), 791–836; and Peter Gottschalk and Timothy Smeeding, ‘Cross-national Comparisons of
Earnings and Income Inequality’, Journal of Economic Literature,35(1997), 633–87.
2OECD, ‘Earnings Inequality’, Employment Outlook (July 1993), 157–84; OECD, ‘Earnings
Inequality, Low-paid Employment and Earnings Mobility’, Employment Outlook (July 1996),
59–108.
3For similar efforts by other political scientists, see Torben Iversen and Anne Wren, ‘Equality,
Employment and Budgetary Restraint’, World Politics,46(1998), 527–55; and Michael Wallerstein,
‘Wage-Setting Institutions and Pay Inequality in Advanced Industrial Societies’, American Journal
of Political Science,43(1999), 649–80.
Comparative Political Economy of Wage Distribution 283
ninetieth percentile to median earnings (the ‘90–50 ratio’) and the ratio of
median earnings to earnings in the tenth percentile (the ‘50–10 ratio’). Basically,
we ask if these political-institutional variables promote a more egalitarian wage
structure by holding back wage growth for highly-paid wage-earners at the
upper end of the distribution, or by raising the relative wages of people at the
bottom of the wage hierarchy. Our results indicate that bargaining centralization
and public sector employment have discernible egalitarian effects in both halves
of the wage distribution, although both primarily operate in the bottom half of
the wage hierarchy. In contrast, the egalitarian effects of union density appear
to be entirely confined to the lower half of the wage distribution, and the opposite
holds for left government, which operates almost entirely in the upper half of
the wage hierarchy.
In the light of conventional wisdom in the comparative political economy
literature, the absence of egalitarian effects of left government at the lower end
of the wage distribution is puzzling. Much of this literature leads us to expect
that labour-affiliated left parties strive to raise the floor for competition among
unskilled, low-paid workers by providing generous unemployment compen-
sation and by boosting mandated minimum wages. As we document below, the
cross-national association between left government and levels of unemploy-
ment compensation is less strong than conventional wisdom implies, and the
association between generosity of unemployment compensation and com-
pression of the lower half of the wage distribution is also quite weak. More
importantly, the final iteration of our regression analysis tests the hypothesis that
the wage-distributive effects of government partisanship are contingent on the
degree of wage bargaining centralization. In comprehensive and centralized
wage bargaining systems of the Northern European variety, we should not
expect minimum wage legislation to have much impact on the distribution of
wages. The results of an interaction model support this hypothesis: under
decentralized wage bargaining, left government turns out to have egalitarian
effects at the lower end as well as the upper end of the wage distribution, but
the egalitarian effects of left government at the lower end diminish as bargaining
centralization increases.
4
Our presentation is organized as follows. We begin with a quick look at the
cross-national patterns of wage inequality that our analysis seeks to explain. We
then review the existing literature, present our hypotheses and independent
variables, briefly explicate the methodology of pooled cross-section time-series
analysis, and discuss the results of linear regressions with 90–10, 90–50 and
50–10 wage ratios as the dependent variable. Along the lines indicated above,
we end by further exploring the wage-distributive effects of government
4The interaction argument builds on David Rueda and Jonas Pontusson, ‘Wage Inequality and
Varieties of Capitalism’, World Politics,52(2000), 350–83, which argues that the determinants of
wage inequality differ across political economy types. Exploring how the determinants of wage
inequality differ across the wage hierarchy, this article was conceived as a companion to the World
Politics piece. In future work, we plan to integrate the two approaches.
284 PONTUSSON,RUEDA AND WAY
partisanship, and conclude by speculating about future trends and the political
consequences of rising wage inequality.
PATTERNS OF WAGE INEQUALITY
Table 1 summarizes the wage inequality observations which serve as the
dependent variable of our analysis. For each country, the table provides the mean
value for each specification of the dependent variable over the entire period
1973–95 and also the percentage change from the earliest to the most recent
observation. It should be noted at the outset that these inequality measures refer
to gross income from employment for individuals: they ignore other sources of
income (government transfers, self-employment, income from capital, etc.) and
also leave out the distributive effects of taxation and income pooling within
households. What follows must not be confused with an analysis of the
distribution of disposable household income.
We also note that the OECD dataset on which we rely is restricted to full-time
employees, except in the case of Austria. Since part-time employees invariably
earn less, on an hourly basis, than full-time employees, the figures in Table 1
understate the extent of wage inequality in the other countries. And since the
incidence of part-time employment has increased in most OECD countries since
the early 1980s, they also understate the upward trend in wage inequality.
5
Keeping these qualifications in mind, income from employment still accounts
for the lion’s share of total income in all OECD countries, and wage inequality
among full-time employees, measured by 90–10 ratios, still correlates quite
closely with broader cross-national measures of income distribution.
6
Table 1 reveals dramatic cross-national variation in wage inequality. In these
sixteen countries, the average both-gender 90–10 ratio for the 1973–95 period
was 2.89. In other words, a person in the ninetieth percentile of the wage
distribution earned, on average, nearly three times as much as a person in
the tenth percentile. Sweden, with an average 90–10 ratio of 2.07, stands out
as the OECD country with the most compressed overall wage distribution. With
the notable exceptions of France and Austria, the continental European countries
included in this dataset (Belgium, Germany, Italy, the Netherlands and
5The exclusion of part-time employees does not pose a serious problem for cross-sectional
comparison, for median hourly earnings of part-time workers as a percentage of median hourly
earnings of full-time workers correlates very closely with 90–10 ratios among full-time workers
cross-nationally. But this data restriction does pose a problem from the point of view of comparing
changes in wage inequality across countries, since the growth of part-time employment varies
considerably across countries. From 1983 to 1998, the incidence of part-time employment actually
declined in the United States and the Scandinavian countries (see OECD, Employment Outlook, July
1999, p. 24, for data on part-time wages as a percentage of full-time wages and p. 240 for data on
the incidence of part-time employment).
6Cf. OECD, Income Distribution in OECD Countries: Evidence from the Luxembourg Income
Study (1995); Gottschalk and Smeeding, ‘Cross-National Comparisons of Earnings and Income
Inequality’; and Wallerstein, ‘Wage-Setting Institutions and Pay Inequality in Advanced Industrial
Societies’.
Comparative Political Economy of Wage Distribution 285
TABLE
1Means and Percentage Changes of the Dependent Variables,
1973–95
90–10 ratios 90–50 ratios 50–10 ratios
Country
and years % % %
covered Mean change Mean change Mean change
Australia 2.81 10.6 1.69 6.6 1.66 3.1
(1976–95)
Austria 3.53 6.1 1.79 1.7 1.96 0.0
(1980–94)
Belgium 2.34 6.7 1.62 4.9 1.45 1.4
(1986–93)
Canada 4.24 12.1 1.82 2.8 2.30 9.1
(1973–94)
Denmark 2.18 0.0 1.55 3.3 1.40 2.8
(1980–94)
Finland 2.45 11.7 1.68 1.2 1.46 10.2
(1977–95)
France 3.27 10.8 1.96 2.9 1.66 5.7
(1973–95)
Germany 2.79 9.7 1.71 2.4 1.63 11.9
(1984–95)
Italy 2.32 5.9 1.63 9.9 1.42 3.4
(1986–95)
Japan 3.07 1.3 1.81 5.1 1.70 6.3
(1975–95)
Netherlands 2.54 9.7 1.63 6.2 1.56 5.8
(1977–95)
Norway 2.08 3.8 1.49 2.7 1.39 6.4
(1980–94)
Sweden 2.07 1.8 1.56 1.2 1.33 0.0
(1975–95)
Switzerland 2.72 2.2 1.69 2.4 1.61 0.0
(1990–95)
United Kingdom 3.17 13.5 1.77 11.1 1.78 1.5
(1973–95)
United States 4.60 22.3 2.04 10.7 2.00 11.0
(1973–95)
Average 2.89 2.29 1.72 3.78 1.64 1.1
Standard deviation 0.74 9.79 0.15 4.40 0.26 6.38
Notes and sources: The percentage changes measure the variation from earliest to
latest available observation in the country series. See OECD, ‘Earnings Inequality,
Low-paid Employment and Earnings Mobility’, pp. 61–2 for all countries except
the United States; for the United States, OECD, ‘Earnings Inequality’, p. 161, and
OECD, ‘Earnings Inequality, Low-paid Employment and Earnings Mobility,
p. 103.
286 PONTUSSON,RUEDA AND WAY
Switzerland) fall within a rather narrow band (2.34–2.79), well below the OECD
average. At the opposite end of the spectrum, the United States occupies an even
more distinctive position than Sweden, with an average 90–10 ratio of 4.6.
Canada, France, Japan and the United Kingdom also turn out to be considerably
more inegalitarian than the OECD average.
Turning to change over time, the cross-national variation in the data is equally
impressive. From the earliest to the most recent observation available for each
country, we observe large increases of 90–10 ratios in the United States, the
United Kingdom, Canada, Australia, the Netherlands, Austria and Italy.
However, overall wage inequality fell quite significantly in Finland, France,
(West) Germany and Belgium over this period, while the remaining countries
– Denmark, Japan, Norway, Sweden and Switzerland – are best characterized
as cases of stability. For all sixteen countries, the average 90–10 ratio increased
by 2.29 per cent. It is tempting to conclude from the 90–10 data in Table 1 that
there is no common trend for wage inequality to increase in the OECD countries,
but this conclusion may be a bit hasty. In many of these countries, wage
inequality declined in the early part of the period covered by these summary
measures and subsequently rose. If we measure change since the trough of wage
inequality in each country, we observe increases of wage inequality in eleven
out of sixteen, and the magnitude of these increases are typically greater than
those shown in Table 1. Also, it is noteworthy that the tendency for rising wage
inequality becomes broader and more pronounced when one looks at data for
men and women separately.
7
In most of these countries, compression of
between-gender differentials offset the effects of growing within-gender
differentials in the 1980s. The growth of wage inequality among men has been
especially strong, and this relates to another feature of Table 1: over the period
1973–95, the tendency for wage inequality to rise was stronger in the upper half
than the lower half of the wage distribution. (This is related to the point about
male inequality because men are over-represented in the upper half of the
distribution.)
CAUSAL HYPOTHESES
Wage inequality measures such as the 90–10 ratio are summary measures of a
complex, multidimensional wage structure, which might be decomposed into
a number of different kinds of wage differentiation: most obviously,
differentials based on education, work experience, gender, race and corporate
profitability. Comparable in degree, wage dispersion may take different
forms in different countries. In view of this complexity, it would surely be
quixotic to look for a very simple explanation of the cross-national patterns of
wage inequality described above. A number of variables are bound to matter and
7See Jonas Pontusson, David Rueda and Christopher Way, ‘The Role of Political-Institutional
Variables in the Making of Gendered Patterns of Wage Inequality’ (Working paper, Institute for
European Studies, Cornell University, 1999).
Comparative Political Economy of Wage Distribution 287
TABLE
2Theoretical Expectations
Relative magnitude of effect
Direction of across the wage hierarchy
overall
effect on 90–10 Upper half Lower half
Explanatory variables ratios (90–50 ratio) (50–10 ratio)
Political Institutional Variables
Union density Negative Weak Strong
Bargaining centralization Negative Weak Strong
Public employment Negative Weak Strong
Left government:
Wage floor variant Negative None Strong
Marginal taxation variant Negative Strong None
Market Forces Variables
Unemployment Positive Weak Strong
LDC trade Positive Weak Strong
Female labour-force
participation Positive Weak Strong
Private service employment:
Demand for ‘food and fun’
variant Uncertain None Uncertain
Innovation incentives
variant Positive Strong Weak
it seems most appropriate to inquire about their marginal effects. In other
words, this is the kind of research question that calls for multiple regression
analysis.
Table 2 identifies eight independent variables that are included in our
regression models, and also summarizes our expectations of their effects on the
distribution of wages. The variables are grouped into two clusters. The first
cluster includes union density, wage-bargaining, public sector employment and
government partisanship. These political and/or institutional variables are the
variables of primary theoretical interest. The second cluster includes the rate of
unemployment, trade with low-wage countries, female labour force partici-
pation and private service employment. We conceive these as control variables,
designed to capture, in an admittedly crude fashion, supply and demand
conditions emphasized by labour economists.
8
Leaving definitions and data
sources for the Appendix,
9
let us elaborate on our theoretical expectations for
each of the eight independent variables.
8A direct test of the supply and demand arguments advanced by labour economists would require
the use of individual-level data.
9The Appendix is available with the website version of this article.
288 PONTUSSON,RUEDA AND WAY
Union Density
Following Freeman, we can distinguish two dimensions of the relationship
between unionization and wage distribution.
10
The first dimension concerns the
distribution of wages among union members, and how it compares to the
distribution of wages among unorganized wage-earners. The second concerns
wage differentials between union members and non-members; in other words,
the wage premium associated with union membership for workers with
equivalent qualifications, experience and other relevant characteristics. Several
arguments lead us to expect that the wage distribution of unionized firms or
sectors will be more compressed than that of the non-unionized firms or sectors.
First, most unions approximate the logic of democratic decision making (one
person, one vote) more closely than markets do, and whenever the mean wage
exceeds the median wage, we would expect a majority of union members to
favour redistributive wage demands. Secondly, as organizations dependent on
membership support in conflicts with management, unions have a strong interest
in curtailing wage setting based on the subjective decisions of foremen and
personnel managers, in order to curtail their ability to discourage activism by
rewarding pliant behaviour. Unions and employers operating in the same
product markets can both be expected to favour standardization of wage rates
across firms – to take wages out of competition – but unions alone have an
unambiguous interest in standardizing wage rates within firms.
The issue of union wage premia renders the relationship between unionization
and wage distribution more complicated, for the net wage-distributive effects
of unionization also depend on the distribution of union membership across the
wage hierarchy. Assuming that there is a wage premium for unionized workers,
unionism would be a source of wage inequality if high-wage workers were better
organized than low-wage workers, whereas the opposite would hold if low-wage
workers were better organized. Helping to sort out this question, recent
Eurobarometer surveys enable us to estimate union density by income quartile
for the member states of the European Union. In 1994, average union density
for the EU as a whole was 37.5 per cent in the first (lowest) income quartile,
37.8 per cent in the second quartile, 34.2 per cent in the third quartile and 23.7
per cent in the fourth quartile.
11
While the distribution of union membership
across the wage hierarchy differs somewhat from one country to another, this
general picture is consistent with conventional wisdom and leads us to expect
union density to be negatively associated with wage inequality in our pooled
regressions.
Given that we expect union density to be associated with wage compression
and that the lower half of the wage distribution tends to be more unionized than
10 Richard Freeman, ‘Unionism and the Dispersion of Wages’, Industrial and Labor Relations
Review,34(October 1980), 3–23; and Richard Freeman, ‘Union Wage Practices and Wage
Dispersion within Establishments’, Industrial and Labor Relations Review,36(October 1982), 3–20.
Cf. also Blau and Kahn, ‘International Differences in Male Wage Inequality’.
11 Eurobarometer, June–July 1994.
Comparative Political Economy of Wage Distribution 289
the upper half, it follows that, everything else being equal, the lower half of the
wage distribution should be more compressed than the upper half. However, it
does not follow that the egalitarian effect of unionization (the effect of, say, a
5 per cent increase of union density) on the lower half of the distribution should
be greater than the effect on the upper half. The question of differential effects
across the wage hierarchy requires us to introduce additional hypotheses. In the
case of union density, we hypothesize that unions which primarily organize
workers in the lower half of the wage distribution tend to be more solidaristic
in their wage demands than unions which primarily organize workers in
the upper half of the distribution. The logic behind this hypothesis hinges on the
proposition that the highly educated workers who occupy the upper end of the
wage distribution enjoy a great deal of individual bargaining power in the labour
market and stand to gain relatively little by joining a union. Unions which aspire
to organize or retain the loyalty of these workers must moderate their pursuit
of wage compression. For unions whose best-paid members occupy the middle
of the wage hierarchy this problem is much less pronounced, if not absent. Hence
we expect union density to have greater egalitarian effects on 50–10 ratios than
on 90–50 ratios.
Bargaining Centralization
The standard argument linking centralization to wage compression asserts that
centralization facilitates the reduction of inter-firm and inter-sectoral wage
differentials since it means that more firms and sectors are included in a single
wage settlement. As Rowthorn suggests, this logic presupposes that at least one
of the parties to centralized bargaining wants to achieve a reduction of inter-firm
or inter-sectoral differentials.
12
Arguably, then, centralization is a facilitating
factor or perhaps a necessary but not sufficient condition for wage compression.
In a somewhat different vein, one might argue that centralization produces
wage compression by altering the distribution of power among actors. In the
Swedish case, low-wage unions insisted on solidaristic measures as a condition
for their participation in peak-level bargaining sought by employers in the
1950s.
13
But why should centralization systematically strengthen the relative
bargaining power of low-wage unions? The logic of the median voter model
might apply here as well: if low-wage and high-wage unions bargain jointly,
organizational politics will influence the demands that they pursue – exerting
pressure for compression – and market forces will be less influential in
determining the distribution of wage increases.
14
Moreover, we hypothesize
that centralized bargaining in the extreme–asingle settlement for all wage
12 Bob Rowthorn, ‘Corporatism and Labour Market Performance’, in Jukka Pekkarinen, Matti
Pohjola and Bob Rowthorn, eds, Social Corporatism (Oxford: Clarendon Press, 1992).
13 Peter Swenson, Fair Shares (Ithaca, NY: Cornell University Press, 1989), pp. 56–8.
14 Cf. Wallerstein, ‘Wage-Setting Institutions and Pay Inequality in Advanced Industrial
Societies’.
290 PONTUSSON,RUEDA AND WAY
earners – renders wage differentials more transparent, and thus politicizes
wage-distributive outcomes. In other words, centralization not only empowers
low-wage unions, but also makes them more likely to demand redistributive
wage settlements.
Centralization is a characteristic of the process of collective bargaining, and
its effects depend on the extent to which wages across the employment spectrum
are determined through collective bargaining. Because wages at the upper end
of the wage distribution are less likely to be regulated by collective bargaining
than wages at the lower end, we expect the egalitarian effects of centralization
to be stronger for 50–10 ratios than for 90–50 ratios.
15
Government Employment
While controlling for the strong association between public-sector employment
and union density, we nonetheless expect the size of the public sector
(government employees as a percentage of the total labour force) to be
negatively associated with wage inequality for several reasons.
16
In general,
public-sector unions appear to be more inclined to favour wage solidarity than
private-sector unions. This is certainly true if one compares unions organizing
similar services on different sides of the public–private divide in the 1970s and
1980s. At the same time, public-sector employers have been more inclined than
private-sector employers to accommodate union demands for compression or
even to initiate compression. While sheltered from competition in product
markets, public-sector employers are more directly exposed to political
pressures favouring equality and robust wage growth.
17
The egalitarian logic of
public-sector wage setting is most pronounced with regard to equal pay
provisions for women and minorities. Very well-documented in the case of the
United States,
18
this point would seem to hold more broadly across the OECD
countries.
These arguments imply that the wage structure of the public sector should be
more compressed than that of the private sector, but a straight comparison of
public and private wage distributions is problematic because the public sector
encompasses a distinct set of economic activities, often characterized by a very
wide spread of educational qualifications. Our expectation that the size of the
15 To model the inertia associated with institutional change, the bargaining centralization value
for a given year used in our analysis is the average for that year and the previous four years. Developed
by Torben Iversen, the measure of bargaining centralization used here captures not only the level
at which collective bargaining occurs (‘centralization’ in a narrow sense), but also the degree of
concentration of union membership at different bargaining levels (see Appendix to the website
version of this article for further details).
16 OECD, ‘Trends in Trade-Union Membership’, Employment Outlook (July 1991), p. 113.
17 Geoffrey Garrett and Christopher Way, ‘Corporatism, Public Sector Employment, and
Macroeconomic Performance’, Comparative Political Studies,32(1999), 411–34.
18 For example, Ronald G. Ehrenberg and Joshua L. Schwarz, ‘Public Sector Labor Markets’, in
O. Ashenfeleter and R. Layard, eds, Handbook of Labor Economics (Amsterdam: North Holland,
1986), pp. 1219–68.
Comparative Political Economy of Wage Distribution 291
public sector has an egalitarian effect rests on the proposition that the wage
structure of, for example, government-run health care is more compressed than
the wage structure of privately run health care. Also, we expect that wage
compression in the public sector spills over into the private sector as
private-sector employers compete with public-sector employers for labour.
Observing that the differences between public-sector and private-sector wage
distributions are smaller in Sweden than in most countries,
19
we should not
conclude that public-sector egalitarianism has been particularly weak in
Sweden: more likely, the spillover effects of public-sector egalitarianism have
been particularly strong in Sweden.
The spillover effects of public-sector egalitarianism are likely to be most
pronounced at the lower end of the wage distribution, for the simple reason that
private employers compete most directly with public-sector employers for
unskilled labour. As we move up the wage hierarchy, the educational
backgrounds and career patterns of public-sector and private-sector employees
become more distinct. Similarly, the importance of equal pay provisions in the
public sector should affect 50–10 ratios more than 90–50 ratios, since women
tend to be over-represented in the lower half of the wage distribution. By
contrast, unionization and collective bargaining encompass more of the upper
end of the wage hierarchy in the public sector, constraining the ability of
well-paid civil servants to capitalize on their marketplace power. On balance,
we still expect the egalitarian effects of public employment to be strongest at
the lower end of the wage distribution.
Left Government
Governments might influence wage distribution through minimum-wage and
equal-pay legislation, other forms of incomes policy, arbitration in bargaining
disputes, and a variety of measures that strengthen the competitive position of
women and other disadvantaged groups (such as immigrants) in the labour
market. Less obviously perhaps, tax policy might influence the distribution of
primary as well as disposable income. Ideally, we would like to explore the
wage-distributive effects of discrete government policies, but we do not have
the data necessary to do so within the framework of pooled cross-section
time-series analysis (requiring yearly observations for each variable). Instead,
we propose to explore the role of government by including Tom Cusack’s
measure of government partisanship in our regressions.
20
Two distinct arguments lead us to expect left government to be negatively
associated with wage inequality. The first argument – the wage floor variant in
19 Janet Gornick and Jerry Jacobs, Gender, the Welfare State and Public Employment
(Luxembourg Income Study Working Paper, no. 168, 1997).
20 Tom Cusack, ‘Partisan Politics and Public Finance’, Public Choice,91(1997), 375–95. The
construction of this measure is described in the Appendix on the website version of this article. Note
that we have inverted Cusack’s index so that higher values signify more leftist government.
292 PONTUSSON,RUEDA AND WAY
Table2–hinges on the proposition that the policy preferences of left parties
raise the floor for competition in the labour market. If there is a legislated
minimum wage, left governments are likely to set the minimum wage closer to
the median wage than right governments. They are also likely to favour higher
levels of unemployment compensation and to promote other social wage
programmes, curtailing the inegalitarian effects of unemployment and, more
generally, boosting the relative bargaining power of unskilled workers. The
second argument – the marginal taxation variant – hinges on the proposition that
left parties favour progressive income taxation. As Hibbs argues, high marginal
tax rates reduce the value of an added increment of income to highly paid people,
and might discourage wage earners in the upper reaches of the wage hierarchy
from taking full advantage of their market power or perhaps using their market
power to gain non-wage benefits from their employers instead.
21
By analysing the determinants of 90–50 and 50–10 ratios, we test these two
variants of the left government argument independently. Whereas the first
argument holds that left government compresses the wage distribution by
raising the relative wages of poorly paid workers, the second argument holds
that left government compresses the wage distribution by holding back highly
paid workers (of course, the two arguments are not contradictory: both effects
may operate simultaneously).
Unemployment
In seeking to assess the wage-distributive effects of the political-institutional
variables discussed above, we ought to control for the effects of market
conditions, which also vary over time and across countries. The rate of
unemployment is perhaps the most obvious control variable. The basic insight
of the literature on labour market segmentation is that unskilled, low-paid
workers are more readily substitutable than more skilled, high-paid workers, and
consequently that their bargaining position is more immediately and more
adversely affected by unemployment.
22
However, there is another side to the relationship between unemployment and
wage inequality, also related to labour market segmentation. As many studies
have shown, employers are more likely to lay off unskilled than skilled workers
during economic downturns. To the extent that it entails a disproportionate loss
of low-paid jobs, an increase of unemployment produces wage compression by
altering the composition of the labour force. To minimize this statistical effect,
the unemployment variable in our regressions is a five-year moving average (the
21 Douglas Hibbs, ‘Fiscal Influences on Trends in Wage Dispersion’ (paper presented at the annual
meeting of the European Public Choice Society, Reggio Calabria, 1987); and Douglas Hibbs and
Ha˚kan Locking, ‘Wage Compression, Wage Drift, and Wage Inflation in Sweden’, Labour
Economics,77(1996), 1–32.
22 Cf. James Galbraith, Created Unequal (New York: Free Press, 1998); and Katherine Bradbury,
‘Rising Tide in the Labor Market: To What Degree Do Expansions Benefit the Disadvantaged?’ New
England Economic Review,32(May–June 2000), 3–33.
Comparative Political Economy of Wage Distribution 293
observation for each year is the average for that year and the preceding four
years). With this specification, we expect sustained higher rates of unemploy-
ment to be associated with higher levels of wage inequality, and the inegalitarian
effects of unemployment to be concentrated in the lower half of the wage
distribution.
23
Trade with Less Developed Countries
Wood argues that much of the trend towards increased wage inequality in the
OECD countries in the 1980s can be attributed to increased manufacturing trade
with Less Developed Countries (LDCs).
24
The basic logic of Wood’s analysis
is that relative supplies of skilled and unskilled labour are a function not only
of domestic conditions, but also of the factor content of trade. By importing less
skill-intensive goods from low-wage countries, OECD countries are essentially
importing low skill labour, so that the effective supply of unskilled labour
relative to skilled labour has grown, putting downward pressure on the relative
wages of the unskilled. To provide at least a partial test of this thesis, which
resonates much of the popular literature on globalization, our regressions
include trade with LDC countries that are not producers and exporters of oil
(non-OPEC), expressed as a percentage of gross domestic product (GDP).
Wood’s argument implies that the inegalitarian effects of this variable should
primarily manifest themselves in the lower half of the wage distribution.
Female Labour-Force Participation
To the extent that women are on average less educated and/or have less work
experience than men, an increase of the proportion of the total labour force made
up of women represents an increase of the relative supply of unskilled or less
skilled labour.
25
Everything else being equal, we expect female labour-force
participation to be associated with more wage inequality, especially in the lower
half of the wage distribution. However, there are clearly countervailing forces
at work here. Women acquire skills through labour-force participation, and
higher levels of female labour-force participation should eventually reduce the
skill gap between men and women. As women increasingly come to occupy
full-time jobs, moreover, the union density gap between men and women should
erode. As a consequence, women become a more important constituency for
23 Using yearly observations rather than a moving average, Rueda and Pontusson do not find any
wage-distributive effects of unemployment in ‘Wage Inequality and Varieties of Capitalism’.
24 Adrian Wood, North–South Trade, Employment and Inequality (Oxford: Clarendon Press,
1994).
25 Cf. Robert Topel, ‘Wage Inequality and Regional Labor Market Performance in the United
States’, in Toshiaki Tachibanaki, ed., Labour Market and Economic Performance (New York: St
Martin’s Press, 1994), pp. 101–32; Lennart Svensson, Closing the Gender Gap (Lund:
Ekonomisk-historiska fo¨reningen, 1995).
294 PONTUSSON,RUEDA AND WAY
unions, encouraging them to pursue a reduction of wage differentials based on
gender. The data at our disposal do not allow us to explore these countervailing
effects with much precision. Still, given the dramatic increase in female
labour-force participation in recent decades, it makes sense for us to control for
women’s share of total employment.
26
Private Service Employment
Relative to manufacturing and public services, private services became an
increasingly important source of employment in all the OECD countries over
the period covered by this analysis (1973–95). Based on the American
experience, it is commonplace to argue that wage inequality and private service
employment are associated. Treating wage inequality as a precondition for
employment growth, the standard version of this argument asserts that the scope
for productivity growth in personal services is inherently limited, that pricing
closely reflects labour costs, and that demand for these services is highly price
sensitive.
27
In countries with a high wage floor, the expansion of ‘food and fun’
sectors, employing largely unskilled labour, will be sluggish at best. Though the
causal arrows are reversed, the implication of this argument is that the
coefficient for private service employment in our regressions should be positive
and largest in the regression that uses 50–10 ratios as the dependent variable.
If we relax the assumption that the production of personal services with a high
content of unskilled labour is tightly constrained by labour costs, we might
expect the opposite association between wage inequality and private service
employment. After all, the expansion of such services entails a relative increase
in demand for unskilled labour, boosting the bargaining position of workers at
the bottom of the wage hierarchy. In a somewhat different vein, it seems
plausible to argue that productivity growth in high-end services involves radical
innovation and depends on wage incentives to a much greater extent than
productivity growth in manufacturing, which tends to be piecemeal and closely
linked to capital investment. By this reasoning, the inegalitarian effects of
private service employment should manifest themselves primarily in the upper
half of the wage hierarchy.
As with female labour-force participation, our goal here is not to sort out the
complex relationship between wage inequality and private service employment.
Again, we are first and foremost interested in the wage-distributive effects of
our four political-institutional variables; the other four variables are included as
controls.
26 On similar grounds, we would like to be able to control for immigration, but the data that we
would need to do that is simply not available. Also, we have been unable to locate comparable
cross-national data on the distribution of educational qualifications (data on average years of
education are available, but not theoretically relevant to our research question).
27 Cf. Iversen and Wren, ‘Equality, Employment and Budgetary Restraint’.
Comparative Political Economy of Wage Distribution 295
METHODOLOGY
Pooled cross-section time-series analysis has become common practice in
quantitative comparative political economy in recent years.
28
In this type of
analysis, country-years are the units of observation of dependent and
independent variables; in other words, regressions are run on multiple
observations for each country, allowing us to take advantage of variation
between, say, Sweden 1990 and Sweden 1991 as well as the variation between
Sweden 1990 and Germany 1990. Pooling is particularly attractive when time
series are short and/or the number of cross sections small, as is often the situation
for comparative political economists because our theories and data pertain to
a small number of countries (typically, ten to twenty OECD countries). By
incorporating over-time variations, pooling dramatically increases the total
number of observations, and this in turn enables us to determine the statistical
significance of results with greater precision and to test more complex causal
models by including more variables in regression.
However, pooled data analysis must not be seen simply as a technical solution
to the small-Nproblem of comparative political economy. This methodology
is inextricably linked to the idea that cross-national variations and changes over
time have common determinants. As Shalev points out, it is common for
cross-sectional and time-series regressions with the same variables to produce
different results.
29
By engaging in pooled data analysis, we ask: What explains
the observed variance across both space and time?
Ordinary least squares (OLS) regression rests on a number of assumptions
about the data-generating process producing the observed values on the
independent variables and the error term. The pooling of data is likely to violate
some of these assumptions for reasons articulated by, among many others, Beck
and Katz.
30
Beck and Katz show that one common solution to some of these
problems, the Parks–Kmenta method, consistently underestimates parameter
variability. Following their recommendations, now accepted by many compar-
ativists, we report OLS estimates of the coefficients and panel-corrected
estimates of their standard errors.
In our regression models, we deal with dynamics and auto-correlation by
including lagged values of the dependent variable on the right-hand side of the
equation.
31
On the assumption that the effects of a one-unit change in a particular
28 See Alexander Hicks, ‘Introduction to Pooling’, in Thomas Janoski and Alexander Hicks, eds,
The Comparative Political Economy of the Welfare State (New York: Cambridge University Press,
1994), pp. 169–88; and Nathaniel Beck and Jonathan Katz, ‘What to Do (and Not to Do) with
Time-series Cross-section Data’, American Political Science Review,89(1995), 634–47, for citations
and general discussion.
29 Michael Shalev, ‘Limits of and Alternatives to Multiple Regression in Macro-comparative
Research’ (paper presented at conference on ‘The Welfare State at the Crossroads’, Stockholm,
1998).
30 Beck and Katz, ‘What to Do (and Not to Do) with Time-series Cross-section Data’.
31 See Nathaniel Beck and Jonathan Katz, ‘Nuisance vs. Substance: Specifying and Estimating
Time-series Cross-section Models’, in John Freeman, ed., Political Analysis (Ann Arbor: University
296 PONTUSSON,RUEDA AND WAY
variable persist indefinitely, the total effect of a change – that is, its effects over
an infinite period of time – can be computed by dividing the value of the
coefficient for the variable of interest by one minus the coefficient for the lagged
dependent variable.
32
In what follows, we report long-run as well as short-run
effects for each variable.
Our models also include dummy variables for each of the countries in our
dataset, though we do not report the coefficient estimates for these variables.
33
Put somewhat crudely, the country dummies control for the values that all
observations for a given country share by representing the variance unique to
that country. Controlling for omitted variable bias, the inclusion of country
dummies in the regression facilitates the estimation and interpretation of the
coefficients by clearing out the influences of country-specific factors.
Our regression models, then, take the following general form:
Y
it
0
Y
i(t1)
1
X
1it
2
X
2it
n
X
nit
1
C
1t
2
C
2t
16
C
16t
⫹␧
it
where i1,2…,irefers to the cross-sectional units, t1, 2, …, trefers to the
time units; Y
it
is the dependent variable; Y
i(t-1)
is the lagged dependent variable;
X
1
,…,X
n
are the other explanatory variables;
0
,
1
,…,
n
are the slopes of the
explanatory variables; C
1
,…,C
16
are the country dummy variables;
1
,…,
16
are the intercepts for each country; and
it
is an independent random error term
normally distributed around a mean of 0 and with a variance of
2
.
It should be noted, finally, that we applied logarithmic transformations to all
our variables (except for the country dummies) before running the regressions
reported below.
34
The slope estimates yielded by these regressions should
therefore be understood as percentage changes or elasticity measures represent-
ing the relationship between the variables. In other words, the regression
coefficients should be interpreted as the percentage change in the dependent
variable associated witha1per cent change in the independent variable in
question.
(F’note continued)
of Michigan Press, 1996), vol. 6, pp. 1–36. As with all time-series data, the possibility of
non-stationarity must be considered. Dickey–Fuller tests for the pooled data revealed no evidence
suggesting that the inequality series was non-stationary: the null hypothesis of non-stationarity was
rejected at better than the 95 per cent level in tests with and without a time trend. The results of
Breusch–Godfrey tests indicate that there is no significant auto-correlation in the reported regressions
after the inclusion of the lagged dependent variable. See William Greene, Econometric Analysis
(Englewood Cliffs, NJ: Prentice Hall, 1990), pp. 426–8.
32 George Box and Gwilyn Jenkins, Time Series Analysis (Oakland, Calif.: Holden-Day, 1976),
chap. 1.
33 Technically, this means that we estimate Least Squares Dummy Variable (LSDV) models. We
are able to avoid perfect colinearity with a full battery of country dummies because we estimate our
models without a general regression intercept. In this set-up, the coefficients for the country dummies
represent the (unique) intercept of the country in question. The results of F-tests and Wald tests
confirm that country dummies belong in the specification of our regression models. See Greene,
Econometric Analysis.
34 In addition, we note that the analysis includes linearly interpolated data for a handful of missing
observations in the wage inequality series. We did not interpolate across gaps of more than three
years, and interpolated observations account for only nineteen out of 211 used in the analysis.
Comparative Political Economy of Wage Distribution 297
LINEAR REGRESSION RESULTS
Table 3 presents the results of the regression with 90–10 ratios as the dependent
variable, and Table 4 presents the results of separate regressions with 90–50 and
50–10 ratios as the dependent variable. Looking at the latter table, we are
interested not only in the sign, size and significance of individual coefficients,
but also in the differences between coefficients for the same variable in the two
regressions. Accordingly, Table 5 reports the results of tests for the equality of
coefficients across the wage hierarchy. Joint significance tests of the differences
in coefficients – for the entire set of variables and the political/institutional
subset – strongly reject the hypothesis that the variables have similar effects
across the wage hierarchy. Moreover, the differences in magnitude for all four
political/institutional variables not only accord with our expectations, but are
significant at better than the 5 per cent level for bargaining centralization, public
sector employment and government partisanship. While the difference for union
density is not significant by standard cut-points, it is in the right direction (with
a greater effect in the lower half of the wage hierarchy), and a joint test of the
two labour market institution variables is significant at better than the 5 per cent
level.
Let us begin by considering the effects of the four control variables. The rate
of unemployment (averaged over five years) turns out to be the only one of these
variables that has any statistically significant effects on the overall distribution
of wages. As expected, the effects of persistent unemployment are indeed
strongly inegalitarian. Contrary to our expectations, however, there is no
significant difference in the effects of unemployment across the wage hierarchy.
The results reported in Table 4 suggest that a given increase of unemployment
weakens the bargaining power of median-income workers relative to the
bargaining power of workers in the ninetieth percentile just about as much as
it weakens the bargaining power of workers in the tenth percentile relative to
median-income workers.
In the 90–10 regression, the signs of the coefficients for low-wage trade and
private service employment are contrary to our expectations. However, the
coefficients for these variables, and for female labour-force participation as
well, do not come close to statistical significance. Even when analysing the
upper and lower halves of the wage distribution separately, we still do not
observe any wage-distributive effects of trade with low-wage countries, nor can
we discern differential effects at different ends of the wage spectrum. A more
disaggregated analysis of the evolution of wages within and across economic
sectors might well show that low-wage trade has distributive effects, but the
results obtained here suggest that Wood exaggerates the significance of this
factor in the rise of wage inequality over the last two decades.
35
35 Similar findings are reported by Vincent Mahler, David Jesuit and Douglas Roscoe, ‘Exploring
the Impact of Trade and Investment on Income Inequality’, Comparative Political Studies,32(1999),
363–95. Leamer challenges the ‘factor-content-of-trade’ approach adopted by Wood and others (see
298 PONTUSSON,RUEDA AND WAY
TABLE
3The Determinants of Wage Inequality (90–10 Ratio), 1973–95
Coefficients
and standard Long-run
Variable errors p-values effects
Lagged dependent variable 0.684 0.001
(0.048)
Unemployment 0.023 0.001 0.073
(0.007)
LDC trade 0.008 0.217 0.025
(0.007)
Female labour-force participation 0.025 0.683 0.079
(0.037)
Private sector services 0.026 0.447 0.082
(0.034)
Union density 0.028 0.027 0.089
(0.012)
Bargaining centralization 0.028 0.001 0.089
(0.008)
Public sector employment 0.094 0.001 0.297
(0.024)
Left government 0.019 0.002 0.060
(0.006)
Notes and sources: All entries are least squares dummy variable estimates with
panel-corrected standard errors in parentheses. Approximate p-values are two-
sided. See Appendix in website version of this article for data sources and
description. N211.
In contrast, unpacking the wage hierarchy does clarify the effects of female
labour-force participation and private sector services. The coefficients for both
are closer to statistical significance in the regression that uses 50–10 ratios as
the dependent variable. Contrary to our expectations for female labour-force
participation but consistent with the ‘demand for food and fun’ variant of the
private services argument, the signs of both coefficients are negative. In other
words, higher levels of female labour-force participation and private sector
employment are associated with less rather more wage inequality in the bottom
half of the distribution. In the 90–50 regression, the coefficients for these
variables have the opposite sign, but only female labour-force participation
approaches significance. Moreover and most importantly, the differences in the
coefficients across the wage hierarchy are significant at around the 5 per cent
(F’note continued)
Edward Leamer, ‘Wage Inequality from International Competition and Technological Change’,
American Economic Review,86(1996), 309–14). Appealing to the Stolper–Samuelson theorem,
Leamer argues that the liberalization of North–South trade alters relative output prices, and hence
relative wages, regardless of what happens to the volume of trade. The results reported here do not
tell us anything about the wage-distributive impact of trade as conceived by Leamer.
Comparative Political Economy of Wage Distribution 299
TABLE
4The Determinants of Wage Inequality in the Upper and Lower Halves of the Wage Distribution
(90–50 and 50–10 Ratios), 1973–95
90–50 ratio 50–10 ratio
Coefficients Long-run Coefficients Long-run
Variable (s.e.) p-values effects (s.e.) p-values effects
Lagged dependent 0.629 0.001 0.539 0.001
variable (0.064) (0.059)
Unemployment 0.007 0.091 0.019 0.009 0.047 0.020
(0.004) (0.005)
LDC trade 0.002 0.676 0.005 0.004 0.456 0.009
(0.005) (0.005)
Female labour-force 0.039 0.114 0.105 0.032 0.224 0.069
participation (0.025) (0.026)
Private sector 0.022 0.991 0.059 0.037 0.073 0.080
services (0.022) (0.020)
Union density 0.010 0.273 0.027 0.026 0.005 0.056
(0.008) (0.009)
Bargaining 0.010 0.048 0.027 0.027 0.001 0.059
centralization (0.005) (0.006)
Public sector 0.031 0.033 0.084 0.086 0.001 0.187
employment (0.015) (0.015)
Left government 0.017 0.001 0.046 0.003 0.518 0.007
(0.005) (0.004)
Notes and sources: All entries are least squares dummy variable estimates with panel-corrected standard errors in
parentheses. Approximate p-values are two-sided. See Appendix in the website version of this article for data sources
and description. N211.
300 PONTUSSON,RUEDA AND WAY
TABLE
5Tests for Equality of Coefficients Across the Upper and Lower
Halves of the Wage Distribution (90–50 and 50–10 Ratios)
Variable p-value
Joint test for equality of effects of all explanatory variables 0.001
Joint test for equality of effects of political/institutional variables 0.008
Unemployment 0.756
LDC trade 0.406
Female labour-force participation 0.049
Private sector services 0.053
Joint test for labour market institution
variables (union density and centralization) 0.043
Union density 0.203
Bargaining centralization 0.024
Joint test for government variables (partisanship and public
employment) 0.006
Public sector employment 0.008
Government partisanship 0.042
Note: p-values are for Wald tests for the equality of coefficients for 90–50 and 50–10
ratios.
level for both variables, indicating notable differential effects across the wage
hierarchy.
The results reported in Tables 3 and 4 contradict the notion that the expansion
of private services entails downward adjustment of the wages of unskilled
workers, and suggest that the production of personal services with low skill
content is not as tightly constrained by labour costs as commonly supposed.
With respect to female labour-force participation, our findings call into question
the proposition that men are, on average, better trained and/or more experienced
than women and might be invoked to support the argument that the expansion
of personal services, both public and private, entails a shift in the kinds of skills
that employers value. In any case, it seems questionable to construe female
labour-force participation itself as a causal factor compressing the lower half
of the wage distribution. More likely, female labour-force participation is
correlated with compression of gender differentials and, because women are
over-represented in the lower half of the wage distribution, this is where
compression of gender differentials manifests itself most strongly. The
determinants of female labour-force participation, its relationship to public and
private service employment and its impact on the politics of wage distribution
constitute a topic for further research.
Turning to the variables of primary theoretical interest for our present
purposes, our results indicate that unionization has a very significant egalitarian
impact on the overall distribution of wages, and that this impact is primarily
concentrated in the lower half of the distribution. The coefficient for union
density in the 90–50 regression also has a negative sign, but it is much smaller
Comparative Political Economy of Wage Distribution 301
(less than half) the size of the coefficient in the 50–10 regression, and does not
satisfy conventional criteria of statistical significance. Consistent with our
expectations, unions compress the wage distribution primarily by boosting the
relative wages of poorly paid workers. The long-run egalitarian effect of a given
increase in union density is more than twice as large in the lower half of the wage
hierarchy.
The results for bargaining centralization and public employment are similar
to the results for union density, and also support the hypotheses developed
above. As Table 3 indicates, both variables have an egalitarian effect on the
overall distribution of wages, and are statistically significant at better than the
1 per cent level. When we analyse the upper and lower halves of the wage
distribution, the egalitarian effects of bargaining centralization and public
employment appear in both regressions, but they are considerably larger in the
regression with 50–10 ratios as the dependent variable. For both variables, the
egalitarian effects are roughly twice as large at the bottom of the wage hierarchy,
and the difference in the 90–50 and 50–10 coefficients is significant at better
than the 5 per cent level.
Perhaps the most interesting results reported in Tables 3–4 concern the
wage-distributive effects of government partisanship. As expected, the effect of
left government on the overall distribution of wages is egalitarian. In contrast
to the other three political-institutional variables, however, the egalitarian
effects of left government are most pronounced in the upper half of the wage
distribution. In the 50–10 regression, the coefficient for left government is still
negative, but very small (about one-sixth of the coefficient for left government
in the 90–50 regression) and far from statistical significance. These results are
consistent with the hypothesis that left parties in government discourage wage
growth in the upper half of the wage distribution by raising marginal income
tax rates, but they offer little support for the hypothesis that left parties promote
relative wage gains for poorly paid workers by setting a floor for competition
in the labour market.
It is particularly intriguing to note that the differential effects of left
government across the wage hierarchy are the opposite of those we observe for
union density. In all the countries covered by our analysis, left parties have
historically been allied with trade unions and rely on the votes of union members
to a greater extent than any other parties. To the extent that unions dictate the
policies of left parties in this arena, it would appear that unions pursue one
distributive objective through collective bargaining and another distributive
objective through politics.
LEFT GOVERNMENT AND WAGE DISTRIBUTION
The preceding discussion raises two questions. First, does our hypothesis about
marginal tax rates correctly specify the mechanism whereby left government is
associated with compression of 90–50 ratios? Secondly, why do we not
302 PONTUSSON,RUEDA AND WAY
Fig. 1. The relationship between left government and top marginal income tax rates
observe the negative association between left government and 50–10 ratios
predicted by the wage floor hypothesis?
Starting with the first question, we have data on top marginal income tax rates
for early 1980s and late 1980s for eleven of the sixteen countries included in
our analysis.
36
Figure 1 plots the average of these two observations against each
country’s average score on Cusack’s (inverted) index of government partisan-
ship for 1970–90, and Figure 2 plots each country’s 90–50 ratio in 1991 against
its average top income tax rate in the 1980s. Figure 1 reveals that left government
is indeed associated with higher tax rates at the top of the income distribution,
and Figure 2 indicates that high marginal tax rates are in turn associated with
lower 90–50 ratios. Taken together, the two figures provide rather thought-
provoking, though impressionistic, evidence in favour of the argument that left
government compresses the upper half of the wage distribution via high
marginal taxes.
In Figures 3 and 4, we repeat this exercise with data relevant to the wage floor
hypothesis. First, we plot average income replacement rates provided by public
schemes during the first year of unemployment in 1985–91 against each
36 Provided by Geoffrey Garrett (Department of Political Science, Yale University), these data
are based on Arnold Heidenheimer, Hugh Heclo and Carolyn Teich Adams, Comparative Public
Policy, 3rd edn (New York: St Martin’s Press, 1990), Table 6.7 and the OECD data series on the
tax/benefit positions of production workers.
Comparative Political Economy of Wage Distribution 303
Fig. 2. The relationship between top income tax rates and 90–50 ratios
country’s average partisanship score for 1970–90.
37
We then plot 50–10 ratios
in 1991 against average unemployment replacement rates. Figure 3 suggests an
association between left government and the generosity of unemployment
compensation, but this association is considerably weaker than the association
between left government and high marginal tax rates in Figure 1. The
relationship between unemployment compensation and 50–10 compression is
weaker still, partly but not only due to the role of the United States and Canada
as outliers. Excluding the United States and Canada produces a better fit in
Figure 4, but the regression line becomes nearly flat: at best, there is a discernible
but very weak relationship between replacement ratios and 50–10 ratios. In other
words, both steps in the argument linking left government to egalitarianism via
a wage-floor effect seem to falter.
Comparing Figures 1 and 3, one might conclude that Social Democratic
parties are distinguished from their competitors on the centre-right of the
political spectrum, particularly Christian Democratic parties, by a greater
commitment to income redistribution, and not so much by a greater commitment
to high levels of basic income security. However, it must be noted that our data
37 Based on OECD, ‘Unemployment Benefit Entitlements and Replacement Rates’ (electronic
database), these unemployment replacement rates refer to someone who earned the equivalent of the
average production worker at the time that he or she became unemployed and includes various
income maintenance programmes (in addition to unemployment insurance in the narrow sense). See
OECD, ‘Unemployment and Related Benefits’, The OECD Jobs Study: Evidence and Explanations
(1994), part II, chap. 8, for more details.
304 PONTUSSON,RUEDA AND WAY
Fig. 3. The relationship between left government and unemployment benefit income replacement
rates
on unemployment compensation do not take into account the percentage of
unemployed workers who are covered by income replacement programmes in
question, and that income replacement for the unemployed constitutes but one
component of the ‘reservation wage’. Conceivably, the wage-floor hypothesis
would fare better with a broader measure of basic income support.
A somewhat different but complementary solution to the puzzle posed by the
absence of any strong association between left government and 50–10
compression holds that the wage-distributive effects of government partisanship
are contingent on labour market institutions. This line of argument is developed
by Rueda and Pontusson, who show that left government is associated with
overall wage compression (measured by 90–10 ratios) in liberal market
economies, such as the United States and the United Kingdom, but not in the
social market economies of Northern Europe.
38
In the latter countries,
collectively negotiated wage agreements typically apply to all workers in a
company or sector, whether or not they are union members, and wage
developments in different companies and sectors are closely linked to each
other. Arguably, these arrangements constrain the ability of governments to
influence the distribution of wages. Put more positively, they enable unions to
negotiate effective wage floors, and therefore reduce their reliance on minimum
wage legislation and other forms of government intervention for this purpose.
38 Rueda and Pontusson, ‘Wage Inequality and Varieties of Capitalism’.
Comparative Political Economy of Wage Distribution 305
Fig. 4. The relationship between unemployment benefit income replacement rates and 50–10 ratios
Within the framework of the preceding analysis, we test this argument by
rerunning the 90–50 and 50–10 regressions reported above with an interaction
term for left government and bargaining centralization. While Rueda and
Pontusson argue that social market conditions are conceptually distinct from
bargaining centralization, it is nevertheless true that social market economies
invariably score higher on Iversen’s centralization index than other countries.
To simplify the test, therefore, we use Iversen’s index as a proxy for social
market conditions. This argument implies that left governments would only
produce egalitarian effects at the lower end of the wage hierarchy where
bargaining is decentralized, whereas centralization would attenuate com-
pression at the upper end of the hierarchy to a much lesser degree. In accord with
these expectations, the interaction term of left government and bargaining
centralization has a positive coefficient in both regressions – indicating the two
are to some extent substitutes – but the coefficient is much larger (0.013
compared to 0.003) and more significant (p0.06 compared to p0.14) in the
regression that uses the 50–10 ratio as the dependent variable.
39
However, as
always with interaction models, the conditional coefficients and standard errors
are of primary interest, since these allow us to assess the effects of left
39 Note that we did not obtain significant interaction effects for left government with any of the
other political-institutional variables in our analysis (union density, bargaining centralization and
public-sector employment).
306 PONTUSSON,RUEDA AND WAY
TABLE
6Egalitarian Effects of Left Government
Conditional on Bargaining Centralization
Level of bargaining
centralization index 90–50 ratio 50–10 ratio
Decentralized 0.019 0.022
(United States, France) (0.011) (0.009)
[0.035] [0.011]
Moderately decentralized 0.017 0.013
(United Kingdom, Italy) (0.007) (0.005)
[0.005] [0.008]
Moderately centralized 0.016 0.005
(Belgium, Germany) (0.006) (0.005)
[0.001] [0.182]
Centralized 0.015 0.004
(Sweden, Austria) (0.008) (0.008)
[0.035] [0.318]
Notes: Panel-corrected standard errors in parentheses. Approximate
p-value from one-sided t-tests in square brackets.
government conditional on bargaining centralisation. Accordingly, Table 6
summarizes the conditional coefficients of left government and their standard
errors across the range of scores on the centralization index.
Table 6 conveys a very clear picture revealing statistically significant
conditional relationships of the type hypothesized. The effects of government
partisanship at the lower end of the wage distribution are strongly contingent
on the degree of bargaining centralization and only statistically significant at
moderate to low levels of centralization, but this is only marginally true for the
effects of government partisanship at the upper end of the wage distribution,
which decline very gradually with centralization but remain significant across
the entire range of centralization scores. In countries where wage formation is
highly decentralized, such as the United States and France, the egalitarian
effects of left government are actually greater in the lower than the upper half
of the distribution. As bargaining centralization increases, the wage-floor effects
of left government disappear while the marginal-tax effects of left government
remain.
40
These results point to the salience of minimum wage legislation and other
forms of government intervention for wage-distributive outcomes under
decentralized labour-market conditions, a result that was obscured in both the
40 In countries with highly centralized wage bargaining, such as Austria and Sweden, the 50–10
regression yields a positive conditional effect for left government, although the coefficient is
statistically indistinguishable from zero.
Comparative Political Economy of Wage Distribution 307
overall 90–10 and separate 90–50 and 50–10 regressions. Furthermore, they
suggest that wage growth at the upper of the distribution remains beyond the
reach of solidaristic unions even under centralized bargaining. While attempting
to influence their political allies’ pursuit of redistributive tax policies,
solidaristic unions operating under conditions of centralized bargaining are able
to boost the relative wages of low-paid workers on their own and also recognize
that the floor for competition in the labour market cannot be set too high.
CONCLUSION
There can be little doubt that market forces have tended to produce more wage
inequality in advanced capitalist societies over the last two or three decades. In
our regression models, persistent mass unemployment emerges as the most
important factor generating inequality, but there are probably other forces at
work here as well, not captured by our crude controls for supply and demand
conditions.
While market forces have tended to generate more inequality, there is
nonetheless no uniform or universal trend towards more overall wage inequality
among full-time employees across the OECD. Strong unions, centralized wage
bargaining, a large public sector, and left government have muted and
sometimes overcome inegalitarian tendencies. In this constellation of counter-
vailing forces, unions primarily promote the relative wages of poorly paid
workers. Under conditions of decentralized wage bargaining, the policies
associated with left government also boost relative wages at the bottom of the
wage hierarchy, but the universal effect of left government is to hold back wage
growth among highly paid workers. Bargaining centralization and public sector
employment have egalitarian effects across the wage hierarchy, though they are
strongest in the lower half.
Can the countries that have bucked the inegalitarian trend continue to do so
in the future? Our analysis offers little comfort on this score, for political-
institutional developments since the mid-1980s have not been favourable to
wage equality. In most countries, union membership has declined under the
pressures of globalization and/or as a result of the growing share of the labour
force employed in private services.
41
In many countries, centralized wage
bargaining has also fallen into disfavour with powerful employer groups and
even some unions, leading to more decentralized forms of bargaining or, at least,
more ‘flexible’ wage settlements at the national level, leaving more room for
differentiation of wage increases at lower levels of bargaining. Finally, the
near-universal expansion of public employment in the 1970s and 1980s came
41 See Torben Iversen and Jonas Pontusson, ‘Comparative Political Economy: A Northern
European Perspective’, in Torben Iversen, Jonas Pontusson and David Soskice, eds, Unions,
Employers and Central Banks (New York: Cambridge University Press, 2000), pp. 1–37.
308 PONTUSSON,RUEDA AND WAY
to a halt or was reversed in the 1990s.
42
While some public services have been
privatized, governments have more generally sought to introduce market-
oriented management principles into the public sector. More market-oriented
pay-setting practices figure prominently in such efforts to reform public
services. Not only has the public sector’s share of total employment declined
in many countries; quite likely, the egalitarian effects of public employment
have also diminished.
From the perspective of the preceding analysis, the most obvious countervail-
ing trend is the recent string of electoral victories for left parties in Western
Europe. It is too early to tell if this development will last and if the new
generation of left politicians retain the policy commitments necessary to foster
equality in the labour market. Arguably, the minimum wage introduced by the
Blair government will offset some of the inegalitarian effects of the decline of
British unions over the last two decades. However, progressive income taxation
does not appear to be high on the agenda of the Blair government or its
continental counterparts.
Our analysis suggests that if recent political-institutional developments
persist, the OECD-wide increase in inequality widely but erroneously perceived
to have characterized the 1980s and early 1990s may become reality. In turn,
this rising inequality can be expected to have important political consequences.
In particular, there may be a symbiotic relationship between moderate levels of
wage inequality and support for the universalistic welfare state. With widening
income inequality, support for means-testing may well increase as the
perception that many flat-rate benefits go increasingly to those who do not really
need them grows. Thus, under wider income inequality, governments may be
tempted to embark upon a programme of greater redistribution through more
generous but means-tested benefits for the neediest matched by a continued and
steady erosion of insurance-based benefits. In this light, the experience of the
United Kingdom is instructive, as increasing inequality over the past twenty
years has fuelled a steady drift away from universalism and insurance-based
programmes and towards greater means-testing, not only under successive
Conservative governments, but indeed under the current Labour government as
well.
43
The stakes in rising wage inequality are potentially high for anyone
committed to broad, inclusive welfare state programmes. Whether or not rising
inequality undermines the foundations of welfare-state universalism and
whether or not it promotes right-wing populism, the political economy of wage
distribution is a topic that deserves greater attention from political scientists than
it has received thus far.
42 See Richard Clayton and Jonas Pontusson, ‘Welfare-state Retrenchment Revisited: Entitlement
Cuts, Public Sector Restructuring and Inegalitarian Trends in OECD Countries’, World Politics,51
(1999), 67–98.
43 Cf., e.g., Nicholas Timmins, ‘Death to Universalism’, Financial Times,25January 1999.
Comparative Political Economy of Wage Distribution 308a
APPENDIX: DEFINITIONS OF INDEPENDENT VARIABLES AND DATA SOURCES
Union density: Employed union members as a percentage of employed labour force (‘net
density’) for all countries but Canada. The Canadian figures include unemployed and retired
people who retain their membership in the numerator and the unemployed in the denominator
(‘gross density’). Source: Jelle Visser, ‘Unionization Trends Revisited’ (Centre for Research
of European Societies and Industrial Relations, Amsterdam, 1996).
Centralisation: Wage-bargaining centralization as measured by Torben Iversen (Department
of Government, Harvard University). Higher figures signify more centralization. Iversen
classifies country-years according to the relative weight of three levels of bargaining (local,
industry and national), and multiplies these weights by a measure of the concentration of union
membership at each level. Thus there are two distinct sources of variation in Iversen’s index:
(1) index scores increase as the relative significance of higher levels of bargaining increases;
and (2) scores increase as union membership becomes more concentrated at any of these
bargaining levels (especially at those that are more significant). See Torben Iversen, Contested
Economic Institutions (New York: Cambridge University Press, 1999), for a complete
specification. These figures have been lagged, so that the value for a given year is the average
of the actual value for that year and the previous four years.
Public sector employment: Government employees (not including employees of state-owned
enterprises) as percentage of total employed labour force. Source: OECD, ‘Historical Statistics’
(electronic database).
Left government: Partisan cabinet composition as measured by Tom Cusack (Wissenschafts-
zentrum, Berlin). Cusack groups parties into five families, multiplies each family’s share of
cabinet portfolios by its weight, and sums the products. In his weighting scheme, 1 radical
Left, 2 moderate Left, 3 Centre, 4 moderate Right, and 5 radical Right. See Cusack,
‘Partisan Politics and Public Finance’ for further details. Year-by-year figures up to 1990 were
provided directly by Cusack; post-1990 figures were calculated based on cabinet data in Europa
Yearbook, and Cusack’s party classifications. The index has been inverted so that higher scores
signify more leftist government.
Unemployment: Average rate of unemployment (unemployed as percentage of total labour
force) for the year in question and the preceding four years. Source: OECD, ‘Historical
Statistics’.
LDC trade: Trade with less developed countries (LDCs) as percentage of GDP, not including
trade with OPEC countries. For all countries but Belgium, figures up to 1990 were provided
by Geoffrey Garrett (Department of Political Science, Yale University). Belgian and post-1990
figures were calculated on the basis of OECD, Monthly Trade Statistics.
Female labour-force participation: Female labour force as percentage of total labour force.
Source: OECD, ‘Historical Statistics’.
Private service employment: Service employment as percentage of total employment minus
government employment as a percentage of total employment. Source: OECD, ‘Historical
Statistics’.
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