older male labour force participation in oecd countries: pension reform and “the reserve army of...

International Labour Review, Vol. 149 (2010), No. 3

Older male labour force participationin OECD countries:

Pension reform and “the reserve armyof labour”

Martin O’BRIEN*

Abstract. Many governments have treated older workers as a “reserve army oflabour”, allowing early exit from the labour force by various means, so the OECDrecommends that the fiscal strain associated with population ageing be addressedthrough pension reform to increase older workers’ labour force participation.Modelling the labour force participation of males aged 55–59 and 60–64 in12 OECD countries over the period 1967–2007, the author finds that labour marketvariables dominate pension reform variables – and country-specific causal factorsalso – in explaining older males’ participation rates. Without consideration of labourmarket conditions, the OECD’s standard prescription of pension reform would thusseem doomed to fail.

demographic shift toward an older age profile is presenting the govern-A ments of many developed countries with a number of policy challenges.Indeed, ageing societies face issues associated with changing consumption pat-terns, labour force participation, health care and pension usage. As a result,population ageing has inspired research and analysis of older workers’ labourforce participation and pension usage both in the countries concerned and ininternational institutions. The primary concern is that an increasing proportionof older people – who are traditionally reliant upon publicly funded pensionsand health care – coupled with a decreasing proportion of working-age peoplepaying income tax will put future government budgets under strain.

The World Bank, the OECD and the ILO are advocating different policiesto address the labour force participation of older workers and pension reform inageing societies. The World Bank recommends gradual replacement of publiclymanaged pay-as-you-go schemes by a multiple-pillar retirement-income system

* School of Economics, University of Wollongong, email: [email protected] for opinions expressed in signed articles rests solely with their authors and

publication does not constitute an endorsement by the ILO.

Copyright © The author 2010Journal compilation © International Labour Organization 2010

240 International Labour Review

(World Bank, 1994; Holzmann, 1997), while the OECD recommends both pen-sion reform and increasing the labour force participation of older workers(OECD, 1995 and 1998; Blöndal and Scarpetta, 1998; Casey, 1998; Duval, 2003;Burniaux, Duval and Jaumotte, 2004). The OECD’s researchers contend that thelabour force participation of older workers has fallen over recent decadesbecause of generous social security benefits and eligibility criteria, which haveencouraged older workers to exit the labour force prematurely. The policies theypropose in order to reduce fiscal exposure and increase older workers’ labourforce participation rates therefore consist in scrapping pensions that allow earlyexit from the labour force, reducing other financial incentives for early exit, andincreasing the standard age of retirement. Such policies, however, fail to takeaccount of the crucial influence of labour market conditions in determining thelabour force participation of older workers.

In contrast, the ILO advocates a full-employment strategy, action againstage discrimination, and flexible retirement-age policies (ILO, 1995 and 2003;Auer and Fortuny, 2000). Furthermore, the ILO largely rejects the other insti-tutions’ rationale for moving from public to privately funded retirement incomeschemes (Gillion, 1997). Yet most countries appear to have embraced theOECD and World Bank strategies of trying to reduce government responsibil-ity for pension financing and attempting to increase older workers’ labour forceparticipation by addressing social security availability and the financial incen-tives for early exit inherent in various social security schemes.

Government efforts to encourage or discourage labour force participationof certain groups bring up the concept of the “reserve army of labour”, a phraseattributed to Marx (Power, 1983). Indeed, while “orthodox” economists mightemphasize a microeconomic focus and financial incentives associated with non-wage forms of income (e.g. pensions, private savings) in explaining olderworkers’ labour force participation, other labour market economists argue thatthe State plays an active role in manipulating the labour force status of margin-alized groups or segments of labour, such as older workers, in response to aggre-gate labour market conditions (see, for example, Offe and Hinrichs, 1985;Laczko and Phillipson, 1991; Peck, 1996). In other words, government policymechanisms can remove older workers from the labour force in periods oflabour surplus and mobilize them back into the labour force during periodsof labour shortage. At the heart of this rationale is the notion that this “reservearmy of labour” can be mobilized to reduce constraints on increased productionand reduce upward pressures on wages in the interests of capital. The “army” isthen deemed expendable in periods of downturn or stagnation.

Recent pension reforms aimed at increasing the labour force participationof older workers were undertaken primarily in response to concerns over popu-lation ageing rather than labour market conditions. As such, they do not neatlyfit the reserve-army-of-labour concept. Furthermore, modelling results suggestthat the pension reforms advocated by the OECD would be ineffective as policytools for stimulating older workers’ labour force participation in many countries.Some OECD governments will therefore struggle to mobilize older workers

Older male labour force participation in OECD countries 241

through pension reform alone without addressing the role of aggregate labourmarket conditions. Indeed, the findings of this article suggest that many countriesmay benefit from accepting ILO policy recommendations to stimulate aggregatelabour demand rather than rely solely on policies addressing the supply side.

The remainder of the article is organized into seven sections. The firstlooks at trends in the labour force participation of older male workers in OECDcountries and analyses various theoretical explanations. It also presents thebackground to recent policy reforms that are expected to affect older workers’labour force participation. The second section presents the methodology usedfor quantitative modelling of the influences of social security and labour marketvariables upon the labour force participation rates of male workers aged 55 to64 years in selected OECD countries. In contrast to previous OECD modelling,allowance is made for both country-specific intercepts and slope terms in vari-ous model specifications. In addition, the analysis extends to stationarity andcointegration issues, which were also neglected in the previous OECD model-ling by Blöndal and Scarpetta (1998) and Duval (2003). This allows for the esti-mation of both long-run models and short-run dynamic models incorporatingerror correction mechanisms. The third section presents the data and unit roottests, while the following three sections report and discuss the estimation resultsof the various macroeconometric models. The final section concludes with asummary of findings and policy implications.

Older male labour force participation,population ageing, and policyFigure 1 charts the average labour force participation rates of males aged 55–59and 60–64 across the OECD countries and by region from 1960 to 2008. Predict-ably, the labour force participation rates of those aged 60–64 years are notice-ably lower than the rates of those aged 55–59. Also, the trend decline in labourforce participation rates from the 1970s to the 1990s is more dramatic for thoseaged 60–64 than for those aged 55–59. An increase in labour force participationthen occurred as from 2000, particularly for those aged 60–64. Lastly, labourforce participation rates in Europe are noticeably lower than the OECD aver-age, with rates in North America marginally higher. However, these aggregateddata conceal a great deal of variation at the individual country level.

Theoretical explanations for older workers’ labour force participationrates typically include the effects of labour-supply choice, demand for labour,and/or state social security and labour force participation policies. The basic,orthodox model of labour force participation is derived from the general modelof consumer demand. It therefore hinges on individual choice between work andleisure emphasizing individual responses to relative price (i.e. financial) vari-ables, which in turn are conditioned by individuals’ tastes and preferences forwork and leisure (Hicks, 1946). These propositions are generally estimated usingmicro-level data. The basic static model has been extended to a dynamic frame-work with the life-cycle model of labour force participation (see, for example,Ghez and Becker, 1975).


Alternative theories argue that elements left exogenous to this labour-supply choice model are fundamental to explaining older males’ labour forceparticipation, namely: demand for labour and the institutional framework.These theories include the neoclassical labour demand theory and segmentedlabour market theory, which address the firm’s design of financial incentives, aswell as the potential role of structural changes in industry-level employment andchanges in employers’ labour use strategies within industries (see, for example,Lazear, 1979; Standing, 1986 and 1997; O’Brien, 2005). Other explanationsgrounded in the labour market include the discouraged worker hypothesis andthe notion of the reserve army of labour, both of which emphasize the state ofthe aggregate labour market, a disadvantaged or marginalized position for olderworkers, and/or the systematic manipulation of the labour force status ofolder workers by government policy-makers (for example, Bowen and Finegan,1969; Standing, 1978; Peck, 1996).

The labour force participation of older workers has recently receivedrenewed attention from governments in connection with population ageing andassociated policy reforms. The populations of many developed countries areindeed ageing as a result of both declining fertility rates and increasing lifeexpectancy. For example, the average fertility rate in OECD countries declined


from over 3.23 children per woman aged 15–49 years in 1960 to less than 1.65 in2006. And this trend is exacerbated by increasing life expectancy, with theOECD average increasing from 65.8 and 70.8 years in 1960 to 76 and 81.7 yearsin 2006 for males and females, respectively (OECD, 2009a). The main policyconcern in the countries experiencing population ageing centres on support fora growing proportion of older citizens traditionally reliant on the public pursefor social security pensions and health-care expenditure. This challenge, ofcourse, is compounded by the shrinking proportion of the population in thetraditional working-age range of 15–64.

Three international institutions have contributed policy platforms de-signed to address the underlying dilemma, namely, the World Bank, the OECDand the ILO. Each has a different political agenda and policy recommendations,with different implications for older male labour force participation rates andsocial security pension use.

The World Bank’s involvement in pension reform derives from its role asa lender to countries with dysfunctional pension systems – in Latin America andcentral and eastern Europe in particular. The World Bank recommends a grad-ual shift away from publicly managed pay-as-you-go schemes that are unfundedand the norm in these countries. Instead, it advocates a multiple-pillar retire-ment income system, consisting of a mandatory publicly managed unfundedscheme, supported by a privately managed funded scheme, and supplementedby voluntary savings schemes (World Bank, 1994; Holzmann, 1997).

The OECD’s policy research also focuses predominantly on supply-sideissues. Its approach can be described in terms of a three-stage process. Firstcame the identification of the future budget exposure likely to result from popu-lation ageing, especially on account of publicly funded pensions (Leibfritz et al.,1995; Roseveare et al., 1996). Second, conditions of eligibility and the value ofpensions available to workers aged 55–64 were examined as factors explainingthe decline in the labour force participation of those in that age group (OECD,1995; Blöndal and Scarpetta, 1998; Duval, 2003). Third, this research was thenused to justify the primary role of pension reform in reversing the trend towardsearly retirement through restrictions on eligibility and lowering of the value ofsocial security pensions – thereby also justifying a diminishing role for publicfinancing. Specifically, the OECD prescription for pension reform consists of:(i) scrapping pension provisions that allow early retirement; (ii) moving towardthe actuarial neutrality of pension systems; and (iii) securing “convergence” ofretirement ages to 67 years (Burniaux, Duval and Jaumotte, 2004). Despite therhetoric about reviewing the barriers to employment faced by older workers,these pension reforms remain at the forefront of the more recent policy recom-mendations tendered by the OECD (2006).

The ILO has a longer history of concern with older workers’ interests thanthe World Bank or the OECD. For example, it was instrumental in improvingdisability pensions for older workers in the 1930s, in promoting policies forearly retirement and the training and placement of older workers in the 1960s,and in supporting job protection for older workers in the 1970s. The ILO’s Older


Workers Recommendation, 1980 (No. 162), called for older workers to be in-cluded within a strategy of full employment, ensuring that unemployment is notshifted from one group to another. Also recommended were legislation againstage discrimination, access to employment suited to older workers’ skills, experi-ence and qualifications, measures to enable them to continue in employmentunder satisfactory conditions, a gradual transition to retirement and a flexiblepensionable age (ILO, 1995 and 2003; Auer and Fortuny, 2000). Governments,however, largely appear to have followed the OECD’s recommendations forpension reform, with aggregate demand stimulus policies implemented onlyrecently to combat the global financial crisis, not to address population ageing.

MethodologyThis section specifies the macroeconometric panel models of older male labourforce participation, which are estimated below using annual data on 12 OECDcountries covering the period 1967–2007. In the light of the foregoing discussion,the labour force participation rates of older men are assumed to be a function offinancial variables that may “pull” workers away from the labour force in linewith orthodox theory, as well as aggregate labour market constraints that may“push” workers out, as suggested by explanations based on the discouragedworker hypothesis and the idea of “the reserve army of labour”. The chosenvariables also reflect the policy tools proposed by the OECD and the ILO,focusing on pension reform and the aggregate labour market, respectively.Separate models are estimated for males aged 55–59 and 60–64 years, whichtake account of the different age-eligibility rules that may be associated withearly retirement routes as well as differences in standard pensionable ages atcountry level. The basic fixed-effects models are specified as follows:

where LFPR denotes the labour force participation rate, k denotes the agegroup (55–59 years or 60–64 years), i denotes an OECD country (i = 1, …, 12),t represents time (1967 to 2007), ITAX is the implicit tax on five years’ con-tinued work in the “early retirement” route for each age group (percentage),SSRR is the social security replacement rate (percentage) based on the value ofunemployment benefits, RETAGE is the standard retirement age for receiptof the old-age pension, UNEMP denotes the male prime-aged (25–54 years) un-employment rate (percentage), and PRIME represents the percentage of prime-aged males in the total population of males aged 15–64 years.

Two alternative variable specifications are included in the models to cap-ture financial incentives for early labour force exit under countries’ pension sys-tems, namely, the implicit tax on continued work (ITAX) and the social securityreplacement rate (SSRR). Following the work of Stock and Wise (1990), Blun-dell and Johnson (1998), Börsch-Supan and Schnabel (1998), Gruber and Wise(1998), Kapteyn and de Vos (1998) and more recently Duval (2003), the implicit


tax on continued work is calculated as the change in net pension wealth fromworking an additional five years (expressed as a percentage).1 If additional con-tributions outweigh additional benefits an implicit tax is present. In theory, theeffect of this variable should be ambiguous because of opposing income andsubstitution effects. However, evidence suggests that implicit taxes have a nega-tive effect on labour force participation (Lumsdaine and Mitchell, 1999).Implicit taxes are calculated for “typical early retirement” routes, such as earlyretirement pensions, disability pensions and unemployment-related pensionsthat can be drawn prior to eligibility for an old-age pension, thus effectivelyallowing exit from the labour force. They are calculated separately for thoseaged 55 and 60 years for use in the respective age-group models. OECD reformpolicies specifically entail a reduction of this implicit tax.

The social security replacement rate (SSRR) is based on unemploymentbenefits as a proxy for the financial generosity of available social security pen-sions as a non-wage source of income. It is used to represent social security valuerelative to potential labour market earnings in order to take account of the par-ticipation/non-participation choice facing older individuals. Again, one wouldexpect a negative coefficient if the generosity of pensions entices workers out ofthe labour force. The replacement rate of unemployment benefits is expected tobe a good proxy for other social security programmes available to older malesallowing early exit from the labour force, including disability pensions and earlyretirement pensions during the period of analysis. First, replacement rates fromunemployment benefits are significantly and positively correlated to other non-employment benefit schemes, such as disability pensions, which are the pre-dominant form of non-employment benefit used by those aged 55–64 years.Second, many unemployment benefit schemes available to older males do notrequire active job search (Blöndal and Scarpetta, 1998). In such cases, recipientsare expected to have “not in the labour force” status in labour force surveys.ITAX and SSRR are used in alternative specifications as one would expect astrong positive correlation between these two variables. The replacement ratedirectly captures the income replacement effect, while the implicit tax capturesage eligibility and pension accrual.

The standard retirement age (RETAGE) is included to account for vari-ation in the age of eligibility for an old-age pension. In most countries it is65 years, though some countries such as Canada, France and Italy have set it aslow as 60 in some years. These latter countries might thus be expected to havelower labour force participation rates, particularly for those aged 60–64. Thisvariable has taken on increased relevance in recent years in the light of OECDpolicy recommendations to increase standard retirement age to 67 years.

Cyclical labour force discouragement is captured by the UNEMP variable,the unemployment rate for prime-aged males. Here, one would expect a nega-tive coefficient if labour force discouragement has a significant influence on the

1 The implicit tax data were obtained directly from Duval (for more detail on the calculationof these data, see Duval, 2003, Appendix 2).


labour force participation of older males. Prime-aged males are used for thismeasure in order to avoid endogeneity issues, with the labour force participa-tion rate of older males being determined jointly with their own unemploymentrates. Also, past research has identified significant hidden unemploymentamong the older male labour force (for example Beatty, Fothergill and MacMil-lan, 2000; Fothergill, 2001; O’Brien, 2001). The use of older male unemploymentrates would fail, by definition, to address this mismeasurement.

As first proposed in Blöndal and Scarpetta (1998), the percentage ofprime-aged males in the total working-age male population (PRIMEPR) isincluded in this specification as an additional aggregate labour market con-straint. This influence is similar to the long-run (rather than cyclical) labour-force discouragement concept proposed by Standing (1978), the BLMR (1983)and Peck (1996). The hypothesis that any increase in the proportion of prime-aged males will crowd out older workers from the labour market implies anegative coefficient. This proposition rests on two assumptions. The first is thatprime-aged workers are more attractive to employers than older workers, per-haps in terms of education and productivity. And the second is that employersare rationing available jobs to the most attractive workers within rigid labourmarket structures, with relative wages unable to adjust to clear the market.

As noted by Duval (2003), panel-data macroeconometric models of olderworkers’ labour force participation are scarce. The model specification pre-sented here shares some similarities with the previous OECD fixed-effectspanel models, while combining the relative strengths of Blöndal and Scarpetta(1998) and Duval (2003). A notable difference is that Blöndal and Scar-petta (1998) modelled the labour force participation of those aged 55–64 as onegroup rather than two subgroups aged 55–59 and 60–64. Duval (2003), by con-trast, specified the dependent variable as the difference in labour force partici-pation between two consecutive five-year age groups, not as the labour forceparticipation rates of those age groups. However, the restriction that wasimposed upon these models in order to obtain common slope coefficientsacross all countries, even with a fixed-effects specification, may have resulted inoversimplification given the unique features of each OECD country’s labourforce participation, social security systems and labour market features.Accordingly, one of the contributions of the research presented in this article isto estimate both basic fixed-effects models and other models allowing country-specific slope coefficients. This makes it possible to determine whether labourforce participation is shaped by common or unique influences across theOECD countries, which may, in turn, have implications for the most suitablepolicy to increase labour force participation in these countries.

Other issues not considered in previous OECD modelling are stationarity,unit root testing and cointegration. This would appear to be a major shortcom-ing given the relatively long time-series used and the availability of tests that canreadily be conducted with contemporary econometrics software packages. Inparticular, Duval’s (2003) inclusion of the difference between the labour forceparticipation rates of two consecutive age groups as the dependent variable,


likely to be I(0), with explanatory variables expected to be I(1), could be a seri-ous misspecification.

A number of unit root tests are performed on the data presented here. TheLevin, Lin and Chu test (LLC) and the Breitung test (BR) assume a commonunit root process across all cross-sections (Levin, Lin and Chu, 2002; Breitung,2000). In addition, several other unit root tests are employed in this article thatallow ρi to vary across countries, namely, the Im, Pesaran and Shin test (IPS)and the Fisher-ADF and Fisher-PP tests (Im, Pesaran and Shin, 2003; Maddalaand Wu, 1999; Choi, 2001). The null hypothesis assumes a unit root for all i, withthe alternative that some cross-sections (countries) do not have a unit root. Inaccordance with standard practice in stationarity testing, all variables are firsttested in their level form. If a unit root is present, the test is then repeated on thefirst differenced data. If the first differenced data is stationary then it is deemedintegrated of order 1, I(1).

It must be noted that no attempt has been made to allow for structuralbreaks in this work. Perron (1989) noted that some unit root test results may bebiased towards not rejecting the null hypothesis of a unit root in the presence ofstructural break(s). However, panel unit root tests incorporating structuralbreaks have only been formulated relatively recently (Im, Lee and Tieslau,2005; Hadri and Rao, 2008) and have yet to be incorporated into mainstreameconometric software.

Non-stationary data should not be modelled in level format as a long-runrelationship unless a linear combination of the variables is stationary, i.e. thevariables are cointegrated (Engle and Granger, 1987). If variables are non-stationary and modelled in their level form, there is a danger of spurious regres-sion results. The presence of a cointegrating relationship can be tested with unitroot tests of residuals from the long-run equation, which should resemble whitenoise and be I(0).

Furthermore, if variables are cointegrated, an error-correction model is avalid dynamic specification (Engle and Granger, 1987). This entails estimatinga short-run relationship with variables specified in first differences and alsoincluding the lagged residual (error-correction term) from the long-run modelspecified in levels. The coefficient attached to the error-correction term shouldbe negative and thus represent the speed of adjustment to the long-run model,i.e.:

where ECM is the stationary residual from the long-run model.The proposed modelling thus makes a number of contributions. First, it

relaxes common parameter restrictions and allows both country-specific inter-cepts (fixed effects) and slope terms in order to take better account of uniquecountry-specific aspects of social security systems and labour market character-istics. One can therefore simulate changes in pensions (as suggested by theOECD) or the labour market (suggested by the ILO) that are specific to each


country. Second, account is taken of the time-series aspects of the data that havebeen neglected in previous OECD modelling, and stationarity issues areaddressed with both unit-root and cointegration tests. In addition, these testsare conducted both for the panel as a whole and for individual cross-sections(countries). Finally, both the long-run and short-run dynamic models are speci-fied and estimated, incorporating error-correction mechanisms.

Data and unit root testsMost of the data for the selected 12 countries cover a maximum time span of 1967to 2007 and were collected directly from the OECD web site (OECD, 2009b).The data for labour force participation, unemployment rates and population pro-portions were constructed from the annual data available, but the social securityreplacement rates (SSRR) were only available biennially (OECD, 2009c). The

Table 1. Descriptive statistics (mean and standard deviation) for 12 OECD countries,1967–2007*

LFPR55 LFPR60 SSRR ITAX55 ITAX60 RETAGE UNEMP PRIME

Australia 79.327(6.816)

55.718(11.796)

22.537(4.056)

–0.108(1.657)

3.345(6.525)

65.000(0.000)

4.292(2.338)

61.862(2.238)

Canada 76.862(4.812)

53.345(7.263)

16.963(2.864)

1.521(4.027)

5.822(8.340)

65.183(0.581)

7.175(1.814)

63.466(3.335)

Finland 68.773(7.338)

40.552(14.878)

28.829(9.915)

55.647(20.087)

57.317(19.819)

65.000(0.000)

5.999(3.534)

62.726(3.474)

France 72.842(8.860)

34.440(19.414)

32.927(6.385)

49.161(14.054)

69.821(17.180)

61.890(2.236)

5.210(2.792)

61.883(2.236)

Germany 80.458(4.615)

41.279(12.877)

28.110(1.656)

40.568(16.867)

44.564(15.636)

65.195(0.601)

5.155(2.812)

63.404(1.077)

Italy 62.241(6.444)

36.103(5.113)

11.366(13.500)

54.950(20.366)

83.780(22.937)

61.305(2.049)

3.969(1.919)

64.184(4.130)

Nether-lands

71.412(7.634)

37.260(17.307)

49.427(5.289)

54.298(28.626)

73.882(17.844)

65.000(0.000)

4.516(2.852)

63.335(3.124)

Norway 84.516(2.319)

68.502(6.703)

28.073(13.696)

13.003(3.728)

13.647(7.029)

67.329(0.872)

2.529(1.629)

63.364(3.484)

Portugal 75.386(4.103)

58.424(7.637)

22.220(16.830)

33.666(21.374)

50.741(27.013)

65.000(0.000)

3.598(1.451)

59.711(2.444)

Spain 79.029(5.648)

53.972(11.633)

29.049(8.547)

58.225(6.486)

66.033(9.018)

65.000(0.000)

8.955(4.387)

62.466(3.192)

Sweden 86.992(2.860)

66.884(7.716)

24.610(9.647)

28.247(17.133)

48.332(29.685)

65.463(0.000)

3.547(2.793)

63.237(2.324)

United States

81.226(4.086)

60.273(7.380)

12.585(1.608)

4.577(0.332)

7.382(2.001)

65.390(0.802)

4.351(1.489)

63.285(3.653)

* ITAX from 1967 to 2004.LFPR55 = Labour force participation rate of males aged 55–59 years; LFPR60 = Labour force participation rate ofmales aged 60–64 years; SSRR = Social security replacement rate; ITAX55 = Implicit tax on five years’ continuedwork for males aged 55 years; ITAX60 = Implicit tax on five years’ continued work for males aged 60 years;RETAGE = Standard retirement age; UNEMP = Unemployment rate of prime-aged (25–54) males; PRIME = Per-centage of prime-aged males in the total male labour force (aged 15–64).Sources: OECD (2009b), OECD (2009c), Duval (2003), author’s calculations.


replacement-rate data for the missing years were therefore imputed by inter-polation between available years. This would appear to be acceptable becausethere is little short-term volatility in this variable. The implicit tax rates andstandard retirement age data were obtained directly from Romain Duval of theOECD for 1967–99, 2002 and 2004 (see also Duval, 2003). The missing data for2000, 2001 and 2003 were imputed by a process of interpolation of available data.These data were not extrapolated beyond 2004. Since not all variables wereavailable for all time periods for each OECD country, the data used in themodels are an unbalanced panel. The countries selected for analysis were chosenon the basis of data availability: only those countries with at least 25 years of

Table 2. Unit root tests

Variables Unit root test

IPS ADF PP LLC BR

LFPR55it 5.798[1.000]

8.064[0.999]

8.005[0.999]

2.431[0.993]

5.131[1.000]

ΔLFPR55it –14.152***[0.000]

196.195***[0.000]

375.330***[0.000]

–15.57***[0.000]

–9.810***[0.000]

LFPR60it 6.204[1.000]

11.326[0.987]

10.827[0.990]

2.492[0.994]

7.0772[1.000]

ΔLFPR60it –12.926***[0.000]

175.800***[0.000]

270.550***[0.000]

–13.39***[0.000]

–7.758***[0.000]

SSRRit 0.331[0.630]

26.949[0.307]

10.080[0.994]

0.862[0.806]

2.187[0.986]

ΔSSRRit –13.690***[0.000]

197.242***[0.000]

153.574***[0.000]

–13.62***[0.000]

–10.89***[0.000]

ITAX55it 4.214[1.000]

18.882[0.758]

8.962[0.998]

–0.763[0.223]

3.157[0.999]

ΔITAX55it –5.998***[0.000]

87.706***[0.000]

44.616***[0.007]

–5.608***[0.000]

–2.352***[0.009]

ITAX60it 2.604[0.995]

22.274[0.563]

8.650[0.998]

0.941[0.827]

4.940[1.000]

ΔITAX60it –4.698***[0.000]

75.009***[0.000]

46.965***[0.003]

–5.007***[0.000]

–1.662**[0.048]

RETAGEit 0.576[0.718]

3.611[0.729]

0.847[0.991]

–0.916[0.180]

–0.374[0.354]

ΔRETAGEit –2.476***[0.007]

18.119***[0.006]

11.591*[0.072]

–3.001***[0.001]

–3.044***[0.001]

UNEMPit –1.051[0.147]

43.859***[0.008]

10.124[0.994]

0.132[0.553]

0.884[0.812]

ΔUNEMPit –8.143***[0.000]

106.492***[0.000]

84.864***[0.000]

–6.656***[0.000]

–6.872***[0.000]

PRIMEit –0.788[0.215]

40.784**[0.018]

2.547[1.000]

–2.227**[0.013]

0.484[0.686]

ΔPRIMEit –4.036***[0.000]

68.020***[0.000]

71.165***[0.000]

–2.053**[0.020]

–1.221[0.111]

Test statistic reported with p-value in brackets.*** Significant at 1 per cent level. ** Significant at 5 per cent level. * Significant at 10 per cent level.IPS = Im, Pesaran and Shin W-Stat; PP = Fisher PP; ADF = Fisher ADF; LLC = Levin, Lin and Chu; BR = Breitung.


observations for all variables were selected for modelling. This left the following12 countries: Australia, Canada, Finland, France, Germany, Italy, the Nether-lands, Norway, Portugal, Spain, Sweden and the United States.

Descriptive statistics for each country are presented in table 1. As ex-pected, the labour force participation rates are higher for those aged 55–59 thanfor those aged 60–64 in all countries. However, there is much greater variation inthe 60–64 participation rates over time. The generosity of social security varieswidely between countries and is generally higher in European countries than inAustralia, Canada and the United States. However, retirement age has been con-stant at 65 years for many countries (with a standard deviation equal to zero).This feature precluded the use of country-specific RETAGE slope terms. France,however, had a standard retirement age at 60 years from 1983 onward, hence thiscountry’s subsequent exclusion from the 60–64 age group model, since the pri-mary purpose of the exercise is to seek quantification of early retirement influ-ences. Prime-age unemployment rates average less than 10 per cent in allcountries, albeit with relatively wide variation over time, following cyclical fluc-tuations, as expected. Finally, at around 62–64 per cent, the average percentageof prime-aged males in the total working-age male population is very similaracross most countries, and has moved very slowly over time, as would be ex-pected with demographic data.

Unit root tests are presented in table 2. Evidence suggests that all variablesare I(1), i.e. a unit root present in the level form of the variable with stationaritysubsequently established in the first difference of each variable. These I(1) vari-ables will therefore be modelled in their level form with residuals subsequentlychecked for stationarity before error-correction modelling.

Basic fixed-effects modelsThe results of the estimated basic fixed-effects models are presented in table 3.The models display relatively high adjusted R2 values, stationary residuals andgenerally comparable, statistically significant coefficients across specificationsfor each age group, indicating fairly robust results. Slope coefficients generallyincrease with age group suggesting that the labour force participation rates ofmales aged 60–64 are more sensitive to both pension value and labour marketconditions. The coefficients are comparable to those obtained from the modelsof Blöndal and Scarpetta (1998) and Duval (2003), considering the differences independent variable and time frame of analysis. In Blöndal and Scarpetta (1998),the SSRR coefficients varied from 0.15 to 0.25, UNEMP from 0.57 to 1.64, andPRIME from 0.47 to 0.93, depending on which specification was reported for thecombined 55–64 age group. Because of his different dependent variable specifi-cation, Duval’s (2003) coefficients cannot be directly compared, though the rela-tive magnitude of the ITAX and UNEMP coefficients are similar.

Estimation results indicate that an increase in social security replacementrates (SSRR) by one percentage point will reduce labour force participationrates by 0.162 percentage points for males aged 55–59 and by 0.277 percentage


points for those aged 60–64. For comparison, a one-percentage-point increasein implicit tax rates (ITAX) will decrease labour force participation rates by0.081 percentage points for males aged 55–59 and by 0.177 percentage pointsfor males aged 60–64. The standard retirement age coefficient was statisticallysignificant only in the models incorporating SSRR. According to these models,raising the standard retirement age by two years – as would be the case in mostcountries raising it from 65 to 67 – would increase labour force participationrates by around 1.4 percentage points for those aged 55–59 and by 2.6 percent-age points for those aged 60–64.

The labour force discouragement variable (UNEMP) coefficient indicatesthat a one-percentage-point increase in the prime-age unemployment ratereduces labour force participation rates by 0.75 to 0.88 percentage points formales aged 55–59 and by 1.82 to 2.16 percentage points for males aged 60–64.Finally, there is evidence of labour market crowding out of older males byyounger prime-aged workers. A one-percentage-point increase in the propor-tion of prime-aged males in the total male population of working age reduceslabour force participation by 0.8 to 1 percentage point for those aged 55–59 andby 1.45 to 1.64 percentage points for those aged 60–64.

Table 3. Basic fixed-effects models

55–59 60–64

Intercept 92.067***[0.000]

147.045***[0.000]

73.919***[0.000]

217.429***[0.000]

SSRRit –0.162***[0.000]

–0.277***[0.000]

ITAXit –.081***[0.000]

–0.177***[0.000]

RETAGEit 0.685***[0.000]

0.007[0.971]

1.310***[0.004]

–0.683[0.189]

UNEMPit –0.752***[0.000]

–0.879***[0.000]

–1.820***[0.000]

–2.163***[0.000]

PRIMEit –0.819***[0.000]

–1.011***[0.000]

–1.450***[0.000]

–1.636***[0.000]

Number of observations 442 406 415 382

R2 0.852 0.857 0.825 0.838

R2 0.846 0.852 0.818 0.832

F statistic 162.938***[0.000]

156.874***[0.000]

134.294***[0.000]

135.851***[0.000]

LLC –5.425***[0.000]

–5.343***[0.000]

–7.272***[0.000]

–7.359***[0.000]

ADF 56.868***[0.000]

56.834***[0.000]

88.324***[0.000]

91.723***[0.000]

PP 60.975***[0.000]

59.141***[0.000]

73.622***[0.000]

67.639***[0.000]

p-value in brackets.*** Significant at 1 per cent level. ** Significant at 5 per cent level. * Significant at 10 per cent level.


Between 1995 and 2007, the labour force participation rate of males aged55–59 increased by an average of 4.58 percentage points in the 12 countries exam-ined here. The model incorporating SSRR predicted an increase of 3.72 percent-age points, composed of −0.19 points from a small average increase in socialsecurity replacement rates, 0.46 points from an increase in standard retirementage, 2.44 points from a reduction in unemployment, and 1.01 points from a reduc-tion in the proportion of prime-aged males in the labour market. By comparison,the model incorporating ITAX predicted an increase in labour force participa-tion rates of 5.91 percentage points, composed of 1.82 points from a reduction inaverage implicit taxes, 2.86 points from the reduction in unemployment and1.23 points from a reduction in the proportion of prime-aged workers. In each ofthese models, the contribution from eased labour market constraints – par-ticularly lower unemployment rates – thus outweighed by far the estimatedeffects from pension reform over the period 1995 to 2007. Significantly, pensionreforms were quite substantial during this period, with implicit taxes decreasingby 22.7 percentage points on average.

Also between 1995 and 2007, the labour force participation rate of malesaged 60–64 increased by an average 9.29 percentage points in the 12 countries.For this age group, the model incorporating SSRR predicted an increase of8.47 percentage points, composed of −0.37 points from higher social security re-placement rates, 0.87 points from the raised standard retirement age, 6.17 pointsfrom lower unemployment and 1.80 points from the declining proportion ofprime-aged workers. The ITAX model predicted an increase in labour force par-ticipation of 12.73 percentage points. Again, the contribution from lowered im-plicit tax – which pension reforms reduced by 21.1 percentage points on average– was relatively small, with the model estimating a resulting increase in labourforce participation rates of 3.59 points. By contrast, the model estimated that re-duced unemployment increased participation rates by 7.27 points, with easedcrowding-out pressures contributing a further increase of 1.87 points.

These results clearly show that labour market variables play a greater partin explaining the labour force participation rates of older workers than do pen-sion reform variables. The ILO’s recommendation to promote full employmentor, at the very least, to stimulate demand for labour would thus appear to be aneffective means of increasing older workers’ labour force participation in orderto address the concerns of ageing societies. It should be noted, however, that thismodel does not take account of the proposed policy’s possible inflationaryeffects, nor of the natural rate of unemployment or NAIRU. Yet the proposedpolicy, being focused on the demand side of the labour market, should be facili-tated by the gradual easing of crowding-out pressures from prime-aged workers.An obvious feature of ageing populations is that the proportion of prime-agedworkers in the labour force is now on a downward trend, thereby relaxing thisaggregate constraint that crowded out older workers in the past. In other words,while workers of the baby-boomer generation may have crowded out olderworkers in the 1970s and 1980s, they are unlikely to experience the same pres-sure to exit the labour market themselves as older workers.


Models allowing country-specific slopesThe restriction of common slope terms across all countries was relaxed one vari-able at a time for ITAX, UNEMP and PRIME. At each stage, testing determinedsignificant differences in slope coefficients across the 12 countries. Ultimately,the most general specification was chosen, expanding the basic fixed-effectsmodel into a model allowing country-specific slope terms for all three of the

Table 4. Fixed-effects models allowing country-specific slopes1

55–59 60–64

ITAXit UNEMPit PRIMEit ITAXit UMEMPit PRIMEit

Australia 0.766***[0.000]

–1.86***[0.000]

–0.98***[0.000]

–0.321***[0.013]

–4.62***[0.000]

0.407[0.446]

Canada –0.306[0.338]

–.062[0.829]

–1.15***[0.000]

0.351[0.551]

0.060[0.926]

–2.54***[0.001]

Finland –0.173**[0.032]

0.221*[0.099]

–1.195*[0.014]

0.208**[0.023]

–0.226[0.401]

–5.09***[0.000]

France2 –0.097*[0.092]

–1.014**[0.012]

–0.046[0.872]

Germany –0.063***[0.010]

–1.25***[0.000]

–1.73***[0.000]

–0.408***[0.000]

–4.35***[0.000]

–1.348*[0.062]

Italy 0.097*[0.075]

–2.59***[0.000]

–1.44***[0.006]

0.087*[0.082]

–3.27***[0.000]

–0.234[0.375]

Netherlands –0.113***[0.000]

–1.11***[0.000]

–1.66***[0.000]

–0.120**[0.042]

–2.80***[0.000]

–4.71***[0.000]

Norway –0.170[0.532]

–0.926*[0.083]

–0.054[0.759]

–0.169[0.619]

–1.242[0.162]

–1.067[0.175]

Portugal –0.201***[0.000]

–1.14***[0.001]

–0.400*[0.077]

–0.231***[0.000]

–1.763**[0.014]

–2.70***[0.000]

Spain 0.190**[0.043]

–1.26***[0.000]

–0.98***[0.000]

–0.052[0.733]

–2.25***[0.000]

–2.42***[0.000]

Sweden 0.002[0.950]

–0.245[0.440]

–0.558[0.137]

–0.024[0.551]

–0.311[0.637]

–2.59***[0.000]

United States 2.110[0.309]

–1.09***[0.000]

–1.02***[0.000]

–0.169[0.693]

–2.51***[0.000]

–1.40***[0.000]

Number of observations 406 382R2 0.942 0.930R2 0.934 0.921F statistic 119.910***

[0.000]101.609***

[0.000]LLC –8.984***

[0.000]–11.068***

[0.000]ADF 118.789***

[0.000]155.763***

[0.000]PP 117.124***

[0.000]117.181***

[0.000]

p-value in brackets.1 Constant and RETAGE coefficients estimated but not displayed. 2 France is excluded from the 60–64 estimatesbecause its standard retirement age is 60.*** Significant at 1 per cent level. ** Significant at 5 per cent level. * Significant at 10 per cent level.


above variables. The restriction imposing common slope terms for all countries –as was done in all previous OECD modelling – is therefore not supported bythe data. The results of the models allowing for country-specific slope terms arepresented in table 4: they offer increased adjusted R2 values compared to thebasic fixed-effects models, indicating a higher level of explanatory power. Fur-thermore, the large amount of variation in these slopes across different coun-tries also implies that the “one-size-fits-all” policy advocated by the OECD isnot appropriate.

For men aged 55–59, the ITAX coefficient is not significant at the 5 per centlevel, or not of the expected sign, in eight of the 12 countries, namely: Australia,Canada, France, Italy, Norway, Spain, Sweden and the United States. The sameholds true for those aged 60–64 in Canada, Italy, Norway, Spain, Sweden and theUnited States. While Australia, Canada, Norway and the United States have hadfairly low implicit taxes over time, Italy, Spain and Sweden have reduced theirsdramatically from high levels, apparently to no avail. These results therefore

Table 5. Predicted change in men’s labour force participation rates, 1995–2007

ITAXit UNEMPit PRIMEit Predicted Actual Difference

55–59

Australia –4.310 7.435 2.286 5.411 3.079 2.332

Canada 3.716 3.716 5.488 –1.772

Finland 3.659 7.428 11.087 8.673 2.414

France 2.481 2.481 –4.201 6.682

Germany 2.200 –1.716 0.214 0.698 10.644 –9.946

Italy 6.462 –6.241 0.221 0.072 0.149

Netherlands 7.636 3.121 6.709 17.466 18.982 –1.516

Norway 0.000 2.319 –2.319

Portugal 7.143 –0.757 6.386 1.855 4.531

Spain –1.214 12.497 –7.370 3.913 5.215 –1.302

Sweden 0.000 2.438 –2.438

United States 0.740 4.087 4.827 0.393 4.434

60–64

Australia –7.228 18.568 11.340 9.702 1.638

Canada 8.206 8.206 10.648 –2.441

Finland –8.347 31.568 23.221 19.918 3.303

Germany 13.535 –6.021 7.514 16.900 –9.386

Italy –7.329 8.158 0.830 –1.519 2.349

Netherlands 7.274 7.943 19.036 34.254 18.167 16.087

Norway 0.000 2.405 –2.405

Portugal 13.755 –1.169 –16.298 –3.711 2.098 –5.809

Spain 22.317 –18.199 4.117 7.333 –3.216

Sweden 14.121 14.121 10.455 3.665

United States 1.697 5.610 7.307 6.064 1.243


indicate that the major pension reforms advocated by the OECD to reduceimplicit taxes are unlikely to have the expected effect on the labour force partici-pation of older workers in most countries. By contrast, the UNEMP and PRIMEcoefficients are significant at the 5 per cent level in the majority of countries forboth age groups.

As was done with the basic fixed-effects models in the previous section,changes in each country’s implicit taxes, unemployment and labour force com-position from 1995 to 2007 were simulated using the country-specific-slopemodels (table 5). Variables with coefficients not significant at the 5 per centlevel were excluded from this analysis. Countries’ labour force participation ratepredictions are generally quite accurate and within five percentage points of theobserved change, with a few exceptions such as France and Germany for thoseaged 55–59, and Germany, the Netherlands and Portugal for those aged 60–64.

For most countries, prediction results thus give further support to the pre-dominant role of labour market variables over pension reform in explaining theincrease in older workers’ labour force participation rates during this period.Only in Germany and Portugal does the effect of a decrease in the implicit taxoutweigh the influence of eased labour market conditions. The prime-ageunemployment rate increased marginally in these two countries from 1995 to2007, which had the estimated effect of reducing labour force participation inthe older age-groups. In virtually all of the other countries, however, the easingof labour market constraints contributed more to increasing the labour forceparticipation of older men than pension reforms (neither effect was dominant inNorway and Sweden). Unemployment rates declined in all countries exceptGermany and Portugal, reversing the discouraged worker effects that had beenprevalent in previous decades. Similarly, the crowding out of older workers byprime-aged males eased in all countries except Italy, Portugal and Spain,thereby allowing increased labour force participation by older workers.

Short-run error correction modelsAll of the above models specified in levels represent long-run relationships. Thestationary residuals from those models indicated quite stable and well-specifiedrelationships. Also of interest, however, are the short-run or dynamic relation-ships between the selected variables in OECD countries, which have so far notbeen explored. The results for these short-run models are presented in table 6.The residuals from these models are once again stationary. The adjusted R2

values are noticeably lower than those obtained from the long-run models,though they are not directly comparable because of the different dependentvariable. Besides, lower explanatory power is generally expected from a short-run dynamic model specified in first differences.

All coefficients are of the expected sign, but those for the pension-relatedfinancial variables (SSRR and ITAX) are low in value and not significant at the1 per cent level. This suggests that most pension reform policies to reduce thefinancial value of pensions will not have an immediate impact. Surprisingly, a


change in the standard retirement age is estimated to have an impact in the shortrun. However, there is only limited scope for altering this variable, and any pol-icy to increase the standard age of retirement would typically be phased in overa relatively long period. By contrast, the coefficients attached to both aggregatelabour market constraint variables (UNEMP and PRIME) are significant andhigher than those for the pension-related financial variables. Finally, the errorcorrection terms indicate that any deviation from the long-run model is only cor-rected or adjusted by around 10 per cent in the following year, suggesting fairlyslow adjustment to the long-run relationship. In summary, the largely insignifi-cant influence of financial variables for pension reform, combined with the sig-nificant influence of labour market constraints and relatively slow adjustmenttoward the long-run model, further strengthen the case for a policy focus on thelabour market as recommended by the ILO.

Conclusions and policy implicationsThe findings from both long-run and short-run dynamic econometric modelsemphasize the role of labour market conditions over social security variables,

Table 6. Short-run error correction models

ΔLFPR55it ΔLFPR60it

Intercept –0.172**[0.021]

–0.228***[0.004]

–0.522***[0.000]

–0.672***[0.000]

ΔSSRRit –0.031[0.258]

–0.076**[0.031]

ΔITAXit –0.0329**[0.0478]

–0.018[0.315]

ΔRETAGEit 0.841***[0.001]

0.7853***[0.003]

1.484***[0.000]

1.369***[0.000]

ΔUNEMPit –0.305***[0.000]

–0.308***[0.000]

–0.550***[0.000]

–0.529***[0.000]

ΔPRIMEit –0.541***[0.000]

–0.536***[0.000]

–0.377***[0.004]

–0.355***[0.000]

ECMit–1 –0.111***[0.000]

–0.117***[0.000]

–0.101***[0.000]

–0.110***[0.000]

R2 0.140 0.156 0.241 0.262R2 0.106 0.120 0.212 0.231F statistic 4.187***

[0.000]4.356***[0.000]

8.445***[0.000]

8.636***[0.000]

LLC –18.248***[0.000]

–18.006***[0.000]

–13.223***[0.000]

–13.944***[0.000]

ADF 325.387***[0.000]

309.837***[0.000]

240.655***[0.000]

236.062***[0.000]

PP 325.361***[0.000]

310.449***[0.000]

275.841***[0.000]

254.254***[0.000]

p-value in brackets.*** Significant at 1 per cent level. ** Significant at 5 per cent level. * Significant at 10 per cent level.


and also country-specific causal factors, in explaining older workers’ labourforce participation in OECD countries. In particular, estimation results showthat the increasing labour force participation of older males experienced sincethe mid-1990s was overwhelmingly the result of declining prime-age unemploy-ment rates and consequent easing of crowding-out pressures in the labour mar-ket, rather than the result of pension reforms and older workers’ response tofinancial incentives.

However, countries continue to focus on pension reforms without regardto labour market conditions, and the important role of the labour market doesnot feature in OECD policy prescriptions to address the concerns of ageing soci-eties. While some OECD research appears to recognize the labour market dis-advantage faced by older workers, including in comparable estimation results oflabour market variables, it is unwilling to advocate a strong active policy stanceon labour demand and employment. The resulting preoccupation with pensionreform alone means that a supply-side policy concerned with the relative priceand availability of non-wage income remains the predominant policy prescrip-tion for ageing societies. While this may ease concerns over future budget expos-ure, the main finding of this article is that pension reforms are doomed to failureunless they are complemented by policies to address the labour market.

Ongoing pension reforms inspired by OECD policy are likely to result ininequitable outcomes, merely shifting the financial responsibility for unemploy-ment or retirement to the individual, while doing little to address deficientaggregate demand and older workers’ labour market disadvantage – which isprecisely what triggered the so-called early retirement phenomenon in the firstplace. This particular concern will be exacerbated in the event of a recession oreconomic downturn. It is therefore suggested that many countries would benefitfrom adhering to the ILO policy focus on stimulating a healthy labour marketrather than relying solely on the OECD’s pension reform policy.

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