1. Introduction
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Standard search theory predicts that the duration of an unemployment spell is increasing in the level of unemployment benefits, because the cost of rejecting a
job offer decreases. Furthermore, if benefits are cut after a fixed period, the reservation wage decreases, and the exit rate out of unemployment increases as the
Ž Ž
.. time of running out approaches Mortensen 1977 . Of these two hypotheses, the
first one is fairly well established empirically. A number of contributions deal with the connection between benefit size and unemployment duration,
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even though there are different results as to the magnitude of the effect. The second prediction
is less researched, and our purpose with the present paper is to add to the knowledge of the potential incentive effects of a fixed unemployment insurance
period. Doing so, we use two large samples of extensive Norwegian register data covering a period including a natural experiment — in 1992 a former rule of 13
weeks benefits withdrawal after 80 weeks was abolished. Thus, we can compare a group of unemployed who were affected by this policy change with another
consisting of individuals who were not.
Compared to most other European countries, the Norwegian unemployment rate has been remarkably low through most of the 1980s, mostly staying between 1.5
and 3 until 1988. Then, in 1989, unemployment increased sharply and stayed at approximately 5–6 of the labour force during the early 1990s. Unemployment
insurance is universal for all employees with earnings above a minimum level. Until 1991, the maximum entitlement period was 80 weeks, followed by 26 weeks
Ž .
without benefits. One could then receive unemployment benefits UB for a
second 80-week period. The alternative to UB for those who did not manage to get a job, would be means tested social benefits. In May 1991, as a response to the
increasing number of long-term unemployed, the length of the period without UB was reduced to 13 weeks. One year later, it was then decided that if the
unemployment agencies had not offered an individual a new job or a labour market programme after 80 weeks, benefits should no longer be withdrawn in the
13 weeks period. Thus, from May 1992, it became possible to receive UB for a continuous period of 186 weeks.
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As we have already suggested, the empirical literature concerning the effect of Ž
. fixed benefits periods is relatively scarce. Meyer 1990 and Katz and Meyer
Ž .
1990 find that spikes in the hazard out of unemployment may be explained by
2
Ž .
The data used in this paper are provided by The Norwegian Social Science Data Services NSD . NSD is not responsible for the authors’ analyses.
3
Ž .
Ž .
Surveys can be found in, e.g., Atkinson 1987 and Layard et al. 1991 .
4
The benefit level is adjusted after 80 weeks. Before the 1992 rule change, individuals with pre-unemployment earnings close to the minimum level for eligibility might not qualify for a second
period. After 1992, the UB level for the second period was set to 90 of the first UB period.
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Ž .
the end of benefits approaching. Fallick 1991 and Narendranathan and Stewart
Ž .
1993a , however, find that the effect of benefits decreases over time. These results are harder to reconcile with the prediction that the tendency to leave
unemployment increases at the end of the benefits period. Micklewright and Nagy Ž
. 1996 , using Hungarian post-transition data, find no rise in the hazard near the
Ž .
time of benefit exhaustion. Winter-Ebmer 1998 , using Austrian data, finds that males react to extended benefits duration but females do not.
Apart from the academic issue of whether the theory yields correct hypotheses, there are of course important policy implications. For instance, Layard et al.
Ž .
1991 advocate using labour market policies similar to those of Sweden to keep long term unemployment down: a fixed UB period combined with active man-
power policy. However, it is not clear that a policy that consists of labour market training programmes and relief jobs does not distort the potential incentives from a
Ž .
fixed benefit period. Carling et al. 1996 , using Swedish data, investigate that question and conclude that such distortions probably do not take place. Their
evidence of the effects of benefit exhaustion on the hazard into employment is, Ž
. however, only marginally statistically significant. Korpi 1995 , in a study of
youth unemployment, does not find benefits to have any significant effect on the rate of transition.
The Norwegian labour market policy includes a variety of training programmes and relief jobs, and thus resembles that of Sweden. Until 1992, these countries
also were similar in having a fixed benefits period, even though it was longer in Norway. Researching Norwegian data can therefore have bearing both on the
incentive and the labour market programme issues. In the present paper, we focus on the former. Recently extensive Norwegian register data have become available.
Using data from these registers, we analyse two random samples of the inflow to the unemployed population, before and after the 1992 changes in unemployment
Ž .
insurance rules. Hernæs and Strøm 1996 have used data from the same registers, but focus on duration dependence in general. To our knowledge, the present
analysis is the first that addresses the fixed UB period question with Norwegian data. Our data sets are large and utilise information from the unemployment
registers, the employers’ register, and other sources. Using a semiparametric proportional hazard model, we estimate exit rates into employment. Search theory
predicts, ceteris paribus, individuals with UB in the first sample to have a jump in the transition rate into employment when the first 80 weeks entitlement period has
expired. Since persons in the second sample would not be faced with a benefit withdrawal after 80 weeks, a corresponding jump therefore should not be observed
there. In addition, the prospect of benefit cuts should lead to increased search efforts in the entire unemployment period of the first sample compared to the
second one.
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Ž .
Atkinson and Micklewright 1991 survey this literature.
In both of our samples we expect to find different exit patterns between persons with and without entitlement to benefits. Furthermore, the 1992 rule changes are
expected to have no effect on unemployed individuals not receiving UB. The non-receivers therefore function as a control group.
We now proceed by presenting our model specification in Section 2. In Section 3 we describe our data sets and discuss the definition of unemployment spells. Our
results are presented and discussed in Section 4. It turns out that the analysis does not give empirical support to the notion that the 1992 reform produced significant
changes in the behaviour around 80 weeks of unemployment duration. However, the reform appears to have had effects on the earlier stages of the search process,
in that the first year’s exit rates for the second sample UB-receivers are lower than in the first sample. Section 5 contains a critical assessment of the findings.
2. Model specification
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