Abstract The stubbornly high unemployment experienced by European countriessince the mid-1970s have led to a major reconsideration of the natural rate para-digm. Traditional theories which describe movements of unemployment as fluctua-tions around a moving natural rate have been challenged by hysteresis theories. Thequestion arises how one can discriminate between these competing theories. To thisend, we estimate a time-varying parameter model of the unemployment rate for theUS, UK, Germany, and France. The parameters of the model were estimated jointlyby maximum likelihood estimation using the Kalman filter algorithm. When the mov-ing natural rate model is tested against the alternative of a unit root process, the unitroot hypothesis is resoundingly rejected. Among the determinants of the natural rateinstitutions that alter labour market incentives for workers appear to have been moreimportant than institutions that affect labour demand.
Keywords Hysteresis · Natural rate of unemployment · Kalman filter
JEL classifications E24 · J51 · J65
1 Introduction
Steadily increasing European unemployment rates since the mid-1970s have chal-lenged the existence of a unique (natural) equilibrium rate of unemployment and
N. Srinivasan (B) · P. MitraIndira Gandhi Institute of Development Research, Gen. A. K. Vaidya Marg, Santosh Nagar,Goregaon (East), Mumbai 400065, Indiae-mail: [email protected]
P. MitraReserve Bank of India, Mumbai, India
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nurtured the idea that the equilibrium rate of unemployment, at least to some extent,tracks the actual rate.1 This idea has come to be known as unemployment hysteresis.2
It was argued that understanding unemployment in Europe would require economiststo dispense with the natural rate hypothesis that underlies much of both Keynesianand Classical macroeconomics (Blanchard and Summers 1988, p. 186).
If so, the implications for macroeconomic theory and policy are somewhat unpalat-able. In so far as the monetary policy has short-term employment effects, hysteresischallenges the classical dichotomy between monetary and the real economy even inthe long run. This has led many economists to view hysteresis as a potential tool forreviving Keynesian ideas. For example, Ball (2009) argues that central banks facinghigh unemployment should expand demand, accepting a rise in inflation to reduce thenatural rate.
A number of studies have investigated hysteresis in unemployment.3 Blanchard andSummers (1986) estimated an ARMA(1,1) model for the unemployment rate allowingfor a time trend. They reported non-stationarity in the case of France, Germany, andthe UK and stationarity for the US.4 On the other hand, Phelps and Zoega (1998) and
1 Traditional theories describe the movements of unemployment as fluctuations around the natural rate.Most shocks cause temporary movements of unemployment around the natural rate. However, occasionalshocks cause permanent changes in the natural rate itself. Thus, unemployment could be stationary arounda process that is subject to structural breaks.2 This discussion of hysteretic versus non-hysteretic systems is related to the general discussion aboutthe nature of macroeconomic time series that has emerged since the work of Nelson and Plosser (1982).The natural rate hypothesis corresponds to trend stationary processes, while the hysteresis hypothesiscorresponds to difference stationary processes (Mitchell 1993). Formally, hysteresis means that temporaryshocks have permanent effects on the level of unemployment. Blanchard and Summers (1986) definehysteresis as “a very high dependence of current unemployment on past unemployment,” and state that“we should say that unemployment exhibits hysteresis when current unemployment depends on past valueswith coefficients summing to 1” (p. 17). More loosely, Blanchard and Summers (1986) define hysteresis asthe case in which temporary shocks have highly persistent, but not permanent, effects on unemployment,“where the sum of coefficients is close but not necessarily equal to 1.”3 Röed (1997) has reviewed the literature on hysteresis whereby a variety of theories have been proposed forthe existence of such hysteresis. What are these channels? When Blanchard and Summers (1986) introducedthe idea of hysteresis, they emphasized the insider-outsider theory of wage bargaining. When workersbecome unemployed, the remaining employed workers (insiders) increase their wage targets, preventingthe unemployed (outsiders) from getting their jobs back (see Lindbeck and Snower 1986, 2001). Anotherpopular explanation emphasizes long-lasting damage suffered by workers who experience unemployment.These workers loose human capital, become less attractive to employers and reduce their job search as theybecome accustomed to being unemployed (Layard et al. 1991).4 Similarly, results reported by Barro (1988) indicate substantial persistence during the post-World War IIperiod for these four countries. The estimated autoregressive (AR1) coefficient for US, UK, Germany, andFrance are, in turn, 0.74, 1.08, 0.90, and 1.02. Mitchell (1993) employs several alternative test proceduresto quarterly unemployment rates for 15 industrialised countries, covering the period from the mid-1960sto 1991. For some countries he also uses annual observations covering more than 100 years. His testsconsistently fail to reject unit roots even when structural break dummies are included in the regressions.Röed (1996) reports similar results in a study encompassing quarterly unemployment rates for 16 countriesbetween 1970–1994. Only in the US is the unit root rejected. Leslie et al. (1995) employ various unit roottests (including deterministic trend terms) for yearly unemployment rates for 23 countries. In most of thecountries the estimation period is from 1948 to 1992. Almost all tests fail to reject the presence of unitroots, with some exceptions for US, Israel, and New Zealand. Another avenue of research has been toexplore hysteresis within a non-linear framework within which multiple stable equilibria or natural ratesare possible (see Matthews et al. 2008 and Minford and Naraidoo 2010). Nevertheless, the implication of
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Papell et al. (2000) estimate a reduced-form model for the unemployment rate to testhow much persistence is left after allowing for structural break in mean unemployment(the natural rate).5 These authors find that the unemployment persistence parameterdrops considerably if structural break in the natural rate is accounted for. These resultslend credence to the view that changes in the natural rate might have been a moreimportant reason for the differential unemployment experience of these countries.6
This has led researchers to explore the determinants of the natural rate.7
This paper addresses two broad issues. (i) Is there clear evidence of hysteresiseffects or high persistence in unemployment? (ii) Why is the European experienceof high and persistent unemployment in the 1980s and 1990s so strikingly differentfrom that of the US and the UK? To this end, we estimate a univariate time seriesmodel of the unemployment rate for the US, UK, Germany, and France, allowingfor time-variation in both the intercept and the slope coefficient.8 The reduced-formmodel we estimate is derived from a theoretical model set out below. The parametersof the model were estimated jointly by maximum likelihood estimation (MLE) usingthe Kalman filter algorithm.
To anticipate our findings, the unemployment rate is a trend stationary process in allthese countries. In fact, the general picture that emerges is one of low persistence. Thedifferential unemployment experience of these countries can be attributed to changesin the natural rate. Among the determinants of the natural rate institutional factors thatalter labour market incentives for workers appear to have been more important thanfactors that affect labour demand.
Footnote 4 continuedthese models are very different from the standard hysteresis i.e., potentially an infinite number of equilibriaor natural rates.5 Similarly, Bianchi and Zoega (1998) estimate Markov switching-regression model to identify the datingof the infrequent shifts in the mean rate of unemployment for fifteen OECD countries.6 If one looks at the evolution of unemployment in the US and UK, on the one side, and the major continentalEuropean countries, Germany and France, on the other, the differences stand out immediately. In the USand UK, unemployment starts off at a higher level in the 1960s. It then grows steadily until the early 1980s,when it peaks. From then on, it declines continuously throughout the 1980s. After the recession of theearly 1990s once again unemployment continued to decline until the recent financial crisis. In Germanyand France, too, unemployment rose sharply in the 1970s. However, rather than peaking and declining, itcontinued to rise in the 1980s and 1990s. There is a moderate decline in the late 1990s in both countries.7 For example, Alogoskoufis and Manning 1988a,b attribute the difference to institutional factors thataffect the demand for labour. That is, higher persistence in wage aspirations and sluggish adjustment ofemployment by firms (labour turnover costs) in Europe compared to the US explains the difference inunemployment persistence. Others have emphasized the role of government policies that affect the shape ofthe supply curve. Specifically, these models attribute persistence in the natural rate to institutions that alterlabour market incentives for workers, such as, minimum wage, the level and duration of unemploymentbenefits, taxes, social security payments etc. (see Minford 1983; Layard and Nickell 1986; Davis andMinford 1986; Burda and Sachs 1987 and Burda 1988).8 This differs from the typical approach to measuring persistence, which is to estimate autoregressive (AR)models with constant parameters, potentially allowing for a break in the intercept. The reason for doing so istwofold. First, institutional characteristics of the labour market have undergone major changes in Germany,France and the UK over the last 50 years (see Siebert 1997 and Saint-Paul 2004 for a comprehensive survey).As a result the reduced form parameters are unlikely to remain fixed in the face of regime change. Second,it is well known that standard unit root test is biased toward non-rejection of the unit root hypothesis if thetrue data generating process includes breaks in its deterministic components (Perron 1990).
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The rest of the paper is organized as follows. In Sect. 2, we start by outlining astandard model of structural unemployment. We show that equilibrium unemploymentin this model depends on the nature of labor market institutions. We then combine thismodel with a government loss function to derive the reduced-form solution for theunemployment rate. In Sect. 3 we discuss the data, methodology used for testing thereduced form model, and present our results. Section 4 concludes.
2 A model of the natural rate of unemployment
We begin with an outline of the standard model of structural unemployment.9 It isassumed that industry is competitive and distributed into two sectors, unionised andnon-unionised (or “competitive”). Each firm enjoys constant returns to scale but islimited by a fixed factor (“entrepreneurship”), so that marginal product declines as theindustry expands. In this framework, a trade union typically determines an optimalunion wage which is some way above the non-union wage.10 Workers who lose theirjobs as a result will then seek jobs in the non-union sector. In principle, these additionalsupplies of labor should force wages down in the non-union sector, until there was fullemployment.
What prevents this from happening? In the standard model the nature of labourmarket institutions prevent this from happening. The social security system guaranteesa minimum income regardless of work and that taxes apply to workers with verylow incomes. As wages in the non-union sector fall, they become progressively lessattractive (after tax) to workers forced out of the union sector. They will go on thedole. The point of the analysis is that the supply of labor is partly determined by thereservation wage of potential workers, which in turn is shaped by institutions likethe minimum wage, the level and duration of unemployment benefits, taxes, socialsecurity payments, etc.
A neat way to capture these ideas is to use the framework employed by Lockwoodet al. (1998). The order of events within each time-period is as follows. First, a tradeunion sets the nominal wage, wt , (in logs), given rationale expectations about the pricelevel.11 Then, an i.i.d. labor demand shock εt occurs, with Et−1εt = 0, Et−1ε
2t = σ 2
ε .Having observed the shock, the central bank/government chooses the price level, pt or,equivalently, the inflation rate, πt . Finally, employment is determined by labor demand,
9 The model has its origins in Minford (1983) and Minford et al. (1983). Later papers (e.g., Layard andNickell 1986 and Layard et al. 1991) adopted imperfect competition instead. But the essential mechanismswere unaltered.10 There is a mass of evidence that unions force firms to pay higher wages than would be paid to similarworkers in non-union firms (see Blanchflower 1996, for a survey). The insider–outsider theory providesan explanation of what gives unions their clout. Specifically, costs associated with hiring and firing andproviding firm-specific training discourage firms from replacing their high-wage unionized employees withlow-wage non-unionized ones. Since, these costs are borne to some extent by the employers they give theinsiders market power (see Lindbeck and Snower 1986, 2001.11 Theories of trade unions (surveyed by Oswald 1985) often describe unions as being able to set wagesunilaterally. Once wages are set, firms in turn determine employment. Because firm’s demand for labornormally declines when real wages increase, labor unions face a trade-off: they implicitly choose betweenhigher real wages and more jobs.
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lt = pt − wt − εt . (2.1)
The trade union has both a real wage target, Ω , and a dynamic employment target,lut , a convex combination of (the log of) the labor force and (the log of) last period’s
employed (the “insiders”). Thus, the trade union is assumed to have per-period pref-erences over real wages and employment captured by a quadratic loss function:
Łut = Et−1
[(1 − θ)(wt − pt − Ω)2 + θ(lu
t − lt )2], (2.2)
where lt denotes employment. Then trade union’s employment target is assumed tobe dynamic:
lut = αlt−1 + (1 − α)n, (2.3)
i.e., a convex combination of the last period’s employment (lt−1) and the log of thelabor force (n). So α indexes the influence of ‘insiders’ in wage-setting; in the extremecase where θ = 1 and α = 1, the wage is set simply to insure the employment of theseinsiders, with “outsiders”, i.e., the unemployed, having no impact on wage-setting (seeBlanchard and Summers 1986). The wage is chosen to minimize Łu
t in (2.2), subjectto (2.3) and labor demand (2.1). This yields an optimal expected real wage
wt − Et−1 pt = (1 − θ)Ω − θlut = − ˜lu
t , (2.4)
where ˜lut is the union’s effective employment target and the expectation Et−1 pt is
conditional on all the variables in the union’s information set, including lagged unem-ployment.
Combining (2.4) with labor demand (2.1), we obtain
lt − lut = πt − Et−1πt − εt (2.5)
i.e., the deviation of the employment from the target is equal to the inflation surpriseplus the shock. Using (2.4) and (2.5), the unemployment rate ut , can then be writtenas:
ut = n − lt (2.6)
= (n − ˜lut ) + ( ˜lu
t − lt ) (2.7)
= (1 − θ)(n + Ω) + θαut−1 − (π − πet ) + εt (2.8)
= (1 − ρ)un + ρut−1 − (πt − πet ) + εt , (2.9)
where ρ = θα and the natural rate of unemployment, un = (1 − θ )(n +Ω)/(1-ρ).The natural rate of unemployment (un) in this model depends on demographic
factors (such as a rise in working age population), and indirectly on labor marketinstitutions. For example, a more generous benefit system will exert upward pressureon the trade union’s real wage target, Ω , because it reduces the fear of job loss on the
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part of existing employees. Institutions that affect labor demand also affect equilibriumunemployment by affecting the degree of unemployment persistence, 0 ≤ ρ < 1.12
2.1 Reduced-form model of the unemployment rate
In order to derive our reduced-form model for the unemployment rate we combine(2.9) with the government’s loss function. As Svensson (1997) and Lockwood et al.(1998) have argued persistence in unemployment means that the interaction betweenthe government and the private sector must be modelled as a dynamic, rather than arepeated game.13 Thus, the government’s intertemporal loss function is given by
L = Et
∞∑t=0
β t[λ(ut − uT )2 + (πt − π∗)2
], (2.10)
where 0 < β < 1 is the discount factor, λ > 0 is the relative weight on unemployment-gap stabilization, uT is the socially desirable unemployment rate assumed to be lowerthan the natural rate of unemployment (un) and π∗ is the socially desirable inflationrate. The government is, for simplicity, assumed to have perfect control over theinflation rate. It sets the inflation rate in each period after having observed the shock,εt . The reduced-form solution for unemployment (see appendix for derivation) in thismodel is
ut = δ + ρut−1 + dεt , (2.11)
where δ = (1−θ)(n+Ω) and d = (1−βρ2/1−βρ2+λ). The solution for unemploy-ment has a autoregressive moving average ARMA(p, q) representation. As pointed outearlier there are two broad explanations for the stubbornly high unemployment experi-enced by European countries in the 1980s. The first explanation is that unemploymentis non-stationary (ρ = 1). If non-stationary it could have potentially an infinite numberof equilibria or natural rates, each shock creating a new one. A much more palatableview from the point of view of macroeconomics and monetary policy is that labormarkets adjust very slowly toward equilibrium so that even temporary shocks willpersist.
12 Note that in this model the degree of persistence of unemployment is equal to the product of the weighton insider employment (α) in the wage-setter’s employment target, and the weight on the employment targetrelative to the real wage target (θ ) in the wage-setter’s loss function. Nevertheless, as Alogoskoufis andManning (1988b) have argued sluggishness in labor demand, because of hiring and firing costs, can also bean important determinant of unemployment persistence. Thus, for example labor turnover costs discouragefirms from hiring when labor demand rises and from firing when labor demand falls leading to involuntaryunemployment. These models predict that equilibrium unemployment depends positively on the degree ofunemployment persistence.13 With persistence the cumulative adverse effects of a given unemployment shock are greater, so the gainsto stabilization are greater. On the other hand, with persistence there is greater temptation to inflate, as anincrease in inflation today has perceived future, as well as current benefits in terms of lower unemployment.So the central bank’s optimal choice of inflation at t , is one that equates the marginal (present value) benefitof inflation surprise with the marginal (present value) cost.
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Referring to Eq. (2.11), when 0 ≤ ρ < 1, the long-run natural rate is unique and pathindependent. However, the speed of adjustment to this long-run is inversely relatedto the size of, ρ. This explanation asserts that what really changed is the equilibrium(i.e., long-run) unemployment rate. This may have been due to exogenous disturbances(demographics) or labor market institutions. According to this view persistence in thedeterminants of the natural rate might have been a more important reason for thedifferential unemployment experience of these countries than pure state dependence.
In what follows we estimate (2.11) using data for the US, UK, Germany, and France,allowing for time-variation in both the intercept and the slope coefficient. The approachwe employ is sufficiently flexible to permit the properties of unemployment to changebetween integrated and mean-reverting process over the sample. This in turn allowsus to discriminate between these competing theories.
3 Estimation
3.1 Data
We use annual data for the US, UK, Germany, and France spanning the period1955–2010. Our data for the unemployment rate is the survey-based measure asreported in the OECD’s Main Economic Indicators. In the case of UK, Germany,and France this data is not available for earlier years. For earlier years we use the datareported in Table A3 in Layard et al. (1991).
3.2 Unit root in unemployment
Following standard practice we begin by testing all these series for stationarity. Themain objective of this exercise is to provide a benchmark for our later results. Resultsfrom augmented Dickey Fuller (ADF), Phillips and Perron (1988) and test havingcomparatively more power like Elliot et al. (1996) are reported in Table 1. With theexception of the US, the ADF test fails to reject the null of a unit root at the 5 %significance level for these countries. The Philips–Perron test fails to reject the null ofa unit root for all four countries.14
Next, for each country we estimate an ARMA(1,1) model for the unemploymentrate allowing for a time trend. The reason for doing so is that a number of studies haveused this approach to estimate the degree of persistence in unemployment (see Blan-chard and Summers 1986 and Alogoskoufis and Manning (1988a). Table 2 presentsour estimates of unemployment persistence coefficient.15 While the persistence coef-ficients are of the order 0.8–0.95 for UK, Germany, and France, it is closer to 0.4 forthe US. In sum, the general picture that emerges is one of high persistence of unem-ployment in UK, Germany, and France and low persistence in the US in the post-WorldWar II period which is consistent with other studies quoted above.
14 Other powerful tests such as Elliot et al. (1996) which is reported in Table 1 and Ng and Perron (2001)which is not reported also fail to reject the unit root null for all countries except the US.15 Results do not change qualitatively if we estimate an ARMA(2, 0) model with a time trend.
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Table 1 Unit root test results
Country ADF t stat PP t stat ERS t stat
US −3.12** −2.46 −2.77***
UK −2.12 −1.60 −0.93
Germany −1.45 −0.97 −1.43
France −0.82 −0.91 −0.56
* Denotes reject H0 at 10 % level, ∗∗ at the 5 % level and ∗∗∗ at the 1 % level. The critical values for the teststatistic (for ADF and PP test) are 3.56, 2.92, and 2.60 for 1, 5, and 10 % significance level, respectively.The critical values for Elliot et al. (1996) test are −2.61, −1.95, and −1.61 for 1, 5, and 10 % significancelevel, respectively
3.3 A time varying parameter model of the unemployment rate
As argued earlier, if one looks at the evolution of unemployment in the US and UK, onthe one side, and the major continental European countries on the other, the differencesstand out immediately. Parallel to these developments the institutional characteristicsof the labor market have undergone major changes in Europe over the last 50 years.Which institutions changed and by how much? Answering this question requires mod-elling the time variation in the parameters. In doing so, it is important to keep in mindthat these institutions have evolved gradually. This would suggest gradual and continu-ous drift in the model parameters rather than the discrete shift that is implicitly assumedin the literature (see Bianchi and Zoega 1998 and Papell et al. 2000, for example).
To this end, we estimate our reduced-form model for unemployment (2.11), treatingunemployment as an observable variable and the intercept and persistence parameteras an unobserved time-varying state variables. We estimate the following model,
ut = δt + ρt ut−1 + εt
δt = δt−1 + ξt
ρt = ρt−1 + ηt ,
where the first equation represents the measurement equation and the remaining twoequations are transition equations.16 The disturbances ξt and ηt are serially uncorre-
16 The theory underlying the model (Sect. 2 above) does not suggest a moving average (MA) component.However, time aggregation as well as other measurement errors could well introduce such a component,
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The European unemployment problem
Fig. 1 Estimates of the intercept with standard error bands
lated disturbances with zero mean and constant variances, and are assumed uncorre-lated with each other in all time periods. These equations represent a state space form,in which the unknown parameters of the model, including the variance of, ξt and ηt ,can be estimated jointly by (MLE) using the Kalman Filter algorithm. Provided withan estimate of the variance of, ξt and ηt , the time series of the parameters, δt and ρt ,can be obtained using the Kalman filter.
The estimated series (δ̂t ) for US, UK, Germany, and France is plotted as the solidline in Fig. 1 along with two root mean-square error bands (95 % confidence interval).
Footenote 16 continuedeven if the theory was true (Alogoskoufis and Manning 1988b). When we estimated the model with,additional MA terms not only were these terms found to be not significant but there was also a considerabledeterioration in the fit of the model.
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The estimates of the intercept indicate significant variation in all four countries overthe sample period. Specifically, the estimate of the intercept was very low to beginwith but started to drift up in the mid-1970s in all four countries. In the case of theUS and UK it rose steadily until the early 1980s, when it peaks. From then on, itdeclines continuously throughout the 1980s. After the recession of the early 1990s itonce again starts to drift down. In Germany and France, too, the intercept rose sharplyin the 1970s. However, rather than peaking and declining, it continued to rise.
So what explains the time-variation in the intercept term? In our model the interceptdepends on the labor force (n), the union’s real wage target (Ω) and the weight itplaces on the employment target relative to the real wage target (θ ). The later, asargued earlier, depends indirectly on the nature of labour market institutions. Forexample, the level and duration of unemployment benefits, minimum wage, taxes,social security payments will exert upward pressure on the trade union’s real wagetarget, Ω . These institutions have undergone major changes over the sample period,especially in Europe. In particular, a whole set of measures raised the reservationwage: the duration of unemployment benefits was partly increased; it was made easierto obtain such benefits; the relative distance between the lowest wage in the labourmarket and non-working income in welfare programs became more narrow; and theminimum wage, which is applied in some countries, was raised.17 All these changesoccurred more or less simultaneously in each of the countries (see Siebert 1997 andSaint-Paul 2004).
Nevertheless, since the 1980s it starts to decline in the case of the US and UK.For the US our estimates of the intercept reveals a hump-shaped pattern: it trended upfrom the 1960s until about 1980, then peaked and has declined since then. Estimatedmovements in the intercept over time naturally raise the question as to which factorscaused these movements. In seeking to explain the evolution of the US natural rate, anumber of authors point to the changing age composition of the labour force (Shimer1998). The proportion of the labour force aged 16–24 rose from 17 % in 1960 to24 % in 1978 as the baby boomers entered the labour force as young workers, and thispercentage fell to 16 % in 2000 as the boomers have aged.18 On the other hand, Kingand Morley (2007) attribute the shift to changes in the benefits, labour productivity,real wages and sectoral shifts in the labour market, with sectoral shifts having largestestimated impact on the natural rate. Our estimates for the US are consistent with theseinterpretations.
17 For example, in France, the minimum wage was raised in 1968, 1974, and 1981, rising from roughly40 % of the average monthly wage in the mid-1960s to 50 % by the late-1980s. Unemployment benefits wereraised in 1979, and guaranteed income benefits were offered in 1989. In Germany unemployment benefitsof various sorts were raised in 1975. In the UK, on the other hand, from the mid-1960s into the early1970s trade union pressure on wages increased which saw the number of industrial conflicts per employeealmost double. Unemployment benefits while modest in size could be claimed indefinitely provided onewas unemployed; there were no effective time limits on being unemployed in receipt of benefit (Minford1991). Notice that our estimates of the intercept indicate a shift in the mid-1970s, about the same time theinstitutional characteristics of the labour market underwent a change in all these economies.18 Gordon (1997) also makes this point but goes on to argue that the late 1960s in the US were a time oflabour militancy, relatively strong unions, a relatively high minimum wage. The 1990s have been a time oflabour peace, relatively weak unions, relatively low minimum wage.
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Turning to the UK, with the arrival of the Thatcher government in the 1980s, labormarket institutions were significantly overhauled. A series of laws throughout the1980s weakened union power by extending the grounds for refusing to join a union;enacting limits on picketing; strengthening employer power to get injunctions againststrikes; and extending individual rights to work against a union (see Minford 1991 andNickell 2001 for further details).19 Furthermore, the actual level of benefit relative toearnings has declined quite rapidly since the late 1970s. After 1986, under the ‘RestartProgram’ the unemployed had to undergo mandatory counselling after 6 months with-out work and prove their own efforts to get a new job. In 1996, the duration of unem-ployment benefits was reduced from one year to 6 months. All these changes have hada gradual impact on the intercept, which declines continuously throughout the 1980s.Conversely, countries that did not implement significant and widespread reforms haveexperienced no such shift. For example, between 1980 and 2000, benefit durationlengthened in Germany and France, union coverage went up in France.
The estimated persistence parameter (ρ̂t ) for these ountries is plotted as the solidline in Fig. 2 along with two root mean-square-error bands. There are three generalpoints to note. First, the absolute degree of unemployment persistence varies acrosscountries and our persistence ratings are consistent with other studies reported above.For example, US displays relatively lower level of persistence than the rest.20 Second,in all four countries the persistence estimates obtained conditional on an intercept shiftare substantially below those conditional on no shift.21 Finally, the general picture thatemerges is one of low variability in the persistence parameter.
For the US the point estimate of persistence is close to zero. This is consistent withthe widely held belief that US labour markets are highly flexible and that its unem-ployment is dominated by purely cyclical movements. In the case of UK and France,employment protection was low to start with, then increased in the late 1960s andearly 1970s.22 With the arrival of the Thatcher government in the 1980s, employmentprotection legislation was weakened. For example, a 1985 law said that to be protectedunder the laws against unfair dismissal, one had to have been employed for 2 years,not just one. In spite of these reforms, we only see a moderate drop in the persistenceparameter in the 1980s. On the other hand, the persistence estimate is roughly stablein Germany while it continues to drift up in the case of France. In France employ-ment protection became stricter between 1980 and 2000 (see Saint-Paul 2004). Ourpersistence estimates are consistent with these developments.
19 The Trade Union legislation of the 1980s moved the balance of power in disputes away from employeesand made it harder for unions to organise. This made it less easy and attractive to join a union. Second, theheavily unionised sectors of the economy (public sector) have been in relative decline since then.20 Bertola (1990) compares ten countries ranked in terms of job security; Italy comes at the top of the table(most restrictions on firing), whereas the US comes at the bottom (least restrictive practices). The UK andGermany come in between.21 Our results are consistent with Phelps and Zoega (1998) and Papell et al. (2000) who also report that theunemployment persistence parameter drops considerably if structural break in the intercept is accountedfor.22 In the 1960s and 1970s the UK passed several laws which made labour markets more rigid; for example,the Redundancy Payment Act of 1965; the Unfair Dismissal Law of 1971; and the Employment ProtectionConsolidation Act of 1978 (see Siebert 1997).
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Fig. 2 Estimates of the unemployment persistence coefficient with standard error bands
Figure 3 plots our estimates of the natural rate [E(u) = (δt/1 − ρt )] for these fourcountries. The standard errors are derived using the delta method. The standard errorbands are sufficiently wide suggesting considerable uncertainty surrounding theseestimates. Nevertheless, the estimates of the natural rate broadly mirror the estimatedevolution of the intercept term. The rise in the natural rate occurred in the mid-1970s inall four countries—about the same time as the possible regime change discussed above.It starts to decline in the US and UK in the early 1980s. On the other hand, countriessuch as Germany and France that did not reform their labour markets have experiencedno such shift. Rather than peaking and declining, the natural rate continued to rise inthese countries.
To illustrate how well the model characterizes the behavior of the unemploymentrate in these four countries, Fig. 4 plots our estimates of the natural rate alongsidethe actual unemployment rate. The moving natural rate model tracks the actual unem-ployment path quite well, predicting a rise in the mid-1970s and a fall in the US and
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The European unemployment problem
Fig. 3 Estimates of the natural rate with standard error bands
UK in the early 1980s. For France and Germany it predicts no such shift. Moreover,the estimated natural rate leads the actual unemployment rate over the business cyclein all the countries, which is inconsistent with the hysteresis view. On the basis ofour results it appears that persistence in the determinants of the natural rate of unem-ployment might have been a more important reason for the differential unemploymentexperience of these countries than pure state dependence in unemployment. Amongthe determinants of the natural rate labour market institutions that affect the shapeof the supply curve appear to have been more important than institutions that affectlabour demand.
We would like to end by briefly commenting on the policy implications of ourmain empirical findings. Our analysis suggests that structural factors that affect theshape of the supply curve can best account for the upward “drift” in unemploymentobserved in Germany and France since the mid-1970s. By implication the appropriate
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Fig. 4 Estimates of the natural rate along with actual unemployment rate
policy response is structural reform aimed at reversing the factors which caused therise. Specifically, emphasis should be placed on reforming institutions that alter labourmarket incentives for workers (i.e., minimum wage, the level and duration of unem-ployment benefits, taxes, social security payments, etc.) in order to insure a lastingreduction in unemployment.
4 Conclusion
The high and persistent levels of unemployment experienced by European countriessince the mid-1970s have led to a major reconsideration of the natural rate paradigm.This paper addresses two broad issues. (i) Is there clear evidence of hysteresis effectsor high persistence in unemployment? (ii) Why is the European experience of high
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The European unemployment problem
and persistent unemployment so strikingly different from that of the US and the UK?To this end, we estimate a univariate time series model of the unemployment rate forthe US, UK, Germany and France, allowing for time-variation in both the interceptand the slope coefficient. The approach we employ is sufficiently flexible to permit theproperties of unemployment to change between integrated and mean-reverting processover the sample, unlike the traditional unit-root literature of classifying time seriesas I(0) or I(1). Our results suggest that the unemployment rate is a trend stationaryprocess in all four countries. Moreover, the general picture that emerges is one oflow persistence. The differential unemployment experience of these countries can beattributed to changes in the natural rate. Specifically, government policies that affectthe shape of the supply curve appear to have been more important than institutionalfactors that affect labour demand. While the empirical results are suggestive of the roleof the institutions, they are best regarded as exploratory at this stage. This is becausewe have not empirically established the explanatory power of these institutions forlong-run unemployment. Current and future research by the authors seeks to addressthese issues.
From a broader perspective, our results are also relevant to the literature on theempirical time-series properties of inflation and unemployment. Our findings contrastwith the viewpoint that unemployment has a unit root, as argued by Ireland (1999)and Doyle and Falk (2008). These authors go on to test for a cointegrating relationshipbetween inflation and unemployment, based on the assumption that the unemploymentrate (and inflation) has a unit root. Our point estimates of the persistence parameterare less than one, implying that unemployment is mean reverting (trend stationary)process in all these countries.
Appendix
This appendix discusses the derivation of Eq. (2.11) in the text. Since the government’sobjective function (2.10) is quadratic and its constraints (2.9) are linear, it is possible toguess that linear-decision rules to solve the government’s optimization problem. Thegovernment bases its decisions at time t solely on the state variables while inflationexpectations are left to be determined by a rational expectations condition. We formthe government’s Lagrangian as:
Substituting Eqs. 4.5, 4.7, and 4.8 into the Euler Eq. 4.4 above, taking expectations,and equating constant terms and coefficients on the states yields values for φi ’s interms of the underlying parameters of the model. Substituting the φi ’s in 4.7 yields:
ut = (1 − ρ)un + ρut−1 +( 1 − βρ2
1 − βρ2 + λ
)εt . (4.9)
Finally, substituting for un from (2.9) yields equation (2.11) in the text.
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