direction des Études et synthèses Économiques g 2008 / 10
TRANSCRIPT
Direction des Études et Synthèses Économiques
G 2008 / 10
Who is confronted to insecure labor market histories?
Some evidence based on French labor market transitions
Magali BEFFY - Élise COUDIN - Roland RATHELOT
Document de travail
Institut National de la Statistique et des Études Économiques
INSTITUT NATIONAL DE LA STATISTIQUE ET DES ÉTUDES ÉCONOMIQUES
Série des documents de travail de la Direction des Études et Synthèses Économiques
NOVEMBRE 2008
We thank Thierry Kamionka and Denis Fougère for their precious help both on technical and practical issues. This work has also benefited from helpful comments from Didier Blanchet, Pascale Breuil, Bruno Crépon, Etienne Debauche, Hélène Erkel-Rousse, Pauline Givord, Guy Laroque, Thomas Le Barbanchon, Thierry Magnac, Antoine Magnier, Sébastien Roux, Béatrice Sédillot and Philippe Zamora. We also thank the participants at the INSEE-DEEE seminar (Malakoff, June 2007), AFSE conference (Paris, September 2007), the INSEE research seminar (Paris, February 2008), the INSEE-DSDS seminar (Paris, March 2008), the CREST-LMi seminar (Malakoff, March 2008), and the ESPE Conference (London, June 2008) for their stimulating comments. All remaining errors are ours. Any opinions expressed here are those of the authors and not of the INSEE.
_____________________________________________
* Direction de l’Évaluation, de la Prospective et de la Performance et Crest
** Faisait partie du Département de l’Emploi et des Revenus d’Activités/INSEE au moment de la rédaction de ce document et Crest
*** Faisait partie du Département des Études Économiques d’Ensemble/INSEE au moment de la rédaction de ce document et Crest
Département des Études Économiques d'Ensemble - Timbre G201 - 15, bd Gabriel Péri - BP 100 - 92244 MALAKOFF CEDEX - France - Tél. : 33 (1) 41 17 60 68 - Fax : 33 (1) 41 17 60 45 - CEDEX - E-mail : [email protected] - Site Web INSEE :
http://www.insee.fr
Ces documents de travail ne reflètent pas la position de l’INSEE et n'engagent que leurs auteurs. Working papers do not reflect the position of INSEE but only their author's views.
G 2008 / 10
Who is confronted to insecure labor market histories?
Some evidence based on French labor market transitions
Magali BEFFY* - Élise COUDIN** - Roland RATHELOT***
2
Who is confronted to insecure labor market histories? Some evidence based on the French labor market transitions
Abstract
This paper presents some empirical evidence on the French labor market focusing on transitions between stable jobs, temporary work, unemployment and non-participation. The model used is based on a Markov chain mixture which allows to distinguish labor market histories that are confined to contingent work and non-employment from the non-confined ones. This enables us to identify, quantify and characterize (conditional on observable characteristics) the workers who never accede to stable jobs and remain stuck to temporary jobs and non-employment spells. We consider quarterly labor market transitions, observed from 2003 to 2006 in the Labor Force survey (LFS). We find that on the whole, about 5% of the working age population experience confined transition dynamics: they cannot access to stable jobs. Confined workers are less educated and are more likely to live in distressed areas.
Keywords: labor market mobility, labor market transitions, mover-stayer models, Markov chains
Qui est confronté à des trajectoires précaires sur le marché du travail ?
Quelques enseignements tirés de l’analyse des transitions sur le marché du travail français
Résumé
Cette étude analyse les transitions sur le marché du travail entre quatre états : les périodes d’emploi stable, d’emploi instable, de chômage et d’inactivité. Le modèle utilisé repose sur un mélange de chaînes de Markov et permet de distinguer les individus dont les trajectoires sont confinées entre l’emploi instable et le non-emploi de ceux qui peuvent accéder à un emploi stable. Ceci nous permet de dénombrer et de caractériser en fonction de caractéristiques observables ces travailleurs qui n’accèderont jamais à l’emploi stable. L’estimation repose sur les données trimestrielles de l’enquête Emploi, de 2003 à 2006. Environ 5 % des 30-49 ans ont des trajectoires confinées entre des périodes d’emploi instable et des périodes de non-emploi et ne n’accèderont jamais à un emploi stable. Un très faible niveau d’éducation augmente les risques d’être confiné, habiter en ZUS également.
Mots-clés : mobilité sur le marché du travail, transitions sur le marché du travail, modèles mover-stayer, chaînes de Markov
Classification JEL : J21, J60, C33
Contents
1. Introduction 6
2. Data 9
2.1. Summary statistics and representativeness. . . . . . . . . . . . . . . . . . . . . . . . 10
2.2. Descriptive statistics on transitions. . . . . . . . . . . . . . . . . . . . . . . . . . . 12
3. Methodology: the conditional confined-unconfined worker model 13
4. Results 16
4.1. Duality in the labor market . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 16
4.1.1. The confined movers. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 17
4.1.2. The nonparticipant stayers. . . . . . . . . . . . . . . . . . . . . . . . . . . . 18
4.1.3. The stable-job stayers. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 19
4.2. Dynamics on the labor market. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 19
4.2.1. Confined and unconfined mover transitions. . . . . . . . . . . . . . . . . . . 19
5. Robustness analysis 21
5.1. Heckman-Singer approach. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 21
5.2. Specification tests: stability of transition matrices across gender. . . . . . . . . . . . 26
6. Conclusion 27
A. Stationary occupation probabilities 28
B. Darmois-type test for coefficient equality across subsamples 29
C. Detailed results 29
3
C.1. Women between 30 and 49. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 29
C.2. Men between 30 and 49. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 30
List of Tables
1 Descriptive statistics. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 11
2 Data description. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 12
3 Four-State Markov transition matrices. . . . . . . . . . . . . . . . . . . . . . . . . . 12
4 Marginal probabilities for each type, given gender and age. . . . . . . . . . . . . . . 16
5 Average Type-probabilities conditional on initial states. . . . . . . . . . . . . . . . . 17
6 Stationary equilibria. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 20
7 Types Distribution. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 23
8 Transition matrices according to types. . . . . . . . . . . . . . . . . . . . . . . . . . 25
9 Limiting Probabilities . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 26
10 Tests for stability of dynamics across gender (p-values).. . . . . . . . . . . . . . . . . 27
11 Data : women between 30 and 49. . . . . . . . . . . . . . . . . . . . . . . . . . . . 30
12 Coefficients: women between 30 and 49. . . . . . . . . . . . . . . . . . . . . . . . . 31
13 Transition matrices: women between 30 and 49. . . . . . . . . . . . . . . . . . . . . 32
14 Data : men between 30 and 49. . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 33
15 Coefficients: men between 30 and 49. . . . . . . . . . . . . . . . . . . . . . . . . . 34
16 Transition matrices: men between 30 and 49. . . . . . . . . . . . . . . . . . . . . . 35
List of Figures
1 Densities of individual probabilities of being stayer in 1, stayer in 4 and confined condi-
tional on initial states: women between 30 and 49. . . . . . . . . . . . . . . . . . . . 30
4
2 Densities of individual probabilities of being stayer in 1, stayer in 4 and confined condi-
tional on initial states: men between 30 and 49. . . . . . . . . . . . . . . . . . . . . . 33
5
1. Introduction
Flexible employment has drastically increased in France since the introductionof short-term contracts
(Contrats à Durée Déterminée, CDD) and temporary work (mission d’intérim) in the early 1980’s.
Short-term contracts represent 66% of hirings in 2005 while 60% of the transitions from employment to
non-employment concern a short-term job ending. These flexible devices, which may be justified by the
need to maintain the competitiveness of the firms, induce a higher frequency of labor market transitions.
The transition rate between employment and nonemployment has significantly increased between 1975
and 2000 (Behaghel, 2003). Risks of involuntary job loss were higher inthe 1990’s than in the 1980’s
(Givord and Maurin, 2003). In this context, studying transitions on the labor market and the distribution
of mobilities within the workforce is of first interest.
The nature of the job contract occupies an important place in the French debate on labor market and
labor legislation. The controversy on the "Contrat Unique", following those on the "Contrat Nouvelle
Embauche" (CNE) and the "Contrat Première Embauche" (CPE) in 2005 and 2006, stressed indeed that
the nature of the job contract is a crucial feature of job quality.1 Then, it is interesting to distinguish job
spells in long-term contract and job spells in short-term contract. Hence, Four states stand out on the
French labor market: stable employment, which contains long-term contract jobs and self-employed;
contingent work, which refers to short-term contracts and temporary orseasonal jobs; unemployment
and nonparticipation. The scope of this paper is to analyze and quantify thedifferent kinds of labor
market histories entailed by the transition dynamics between those states.
Short-term jobs may be a stepping stone in an integration process or a trap intoinsecurity. The
economic literature supports both aspects. On the one hand, theories of imperfect information (Spence,
1973), transaction costs (Williamson, 1975) and insider-outsider (Lindbeck and Snower, 1986, 2002)
give some explanations of a dual labor market which either rely on the heterogeneity of the labor supply
1On the one hand, the pros of a "Contrat Unique" advocate for standardizing the multiple kinds of jobs contract in a singleform. On the other hand, in recent years, the government made two attempts to introduce new forms of job contracts: the CNE,introduced in August 2005, was a long-term contract with simplified and lightened termination rules only available for firmswith at most 20 employees; the CPE was an attempt of generalization of the CNE available only for young workers (under26). These two attempts aborted: the CPE was canceled due to tough demonstrations in spring 2007; the CNE was declaredunconstitutional just two years after its introduction.
6
productivity or on the existence of negotiation power in a context of imperfect information.2 These
theories stress the role of signalling in perpetuating a vicious circle. Employers may consider a long
history through unemployment and contingent work as a bad signal on a worker’s ability and then
they may offer him or her insecure positions rather secure ones (Katz, 1986).3 Further, Cahuc and
Postel-Vinay (2002) and Blanchard and Landier (2002) relate the labormarket duality to the coexistence
of short-term jobs and highly protected long-term ones. Besides, theories of segmented labor markets
stress the outstanding role of firms in shaping the labor market duality with the existence of internal
labor markets and human resources’ management or human capital investment which differ according
to the job sector; see for instance the seminal work of Doeringher and Piore (1971). On the other
hand, temporary jobs, more precisely short job spells, can be viewed as opportunities, especially for
young workers, to accumulate general human capital. Temporary job spellsmay also provide a worker
with enough time and/or information to find out the best firm match; see Burdett (1978), Jovanovic
(1979a, 1979b), Mortensen (1988), Topel and Ward (1992).
Cross-sectional studies on labor market duality (see L’Horty, 2004, Gazier and Petit, 2007) do not
take into account the (complete) labor market histories. Here, we adopt a totally different strategy.
The identification of the duality structure relies only on the observed transitions between the different
positions/states on the labor market. More precisely, we use a mover-stayerapproach (Blumen, Kogan,
and MacCarthy, 1955, Kamionka, 1996), which distinguishes workers who remain stuck to contingent
work (typically those alternating nonemployment spells with short-term jobs) from those who may
access to stable jobs and benefit in a sense from an integration process.The approach proposed is
conditional on individual characteristics, which extends Kamionka (1996). Hence, this method enables
us to separate labor market histories which are confined to contingent work and non-employment from
those which are not and to characterize the individuals who experience them.
The discrete time mover-stayer model was first introduced by Blumen, Kogan, and MacCarthy (1955)
to study industrial mobility on the labor market; see also Anderson and Goodman(1957), Goodman
(1961), Spilerman (1972), Singer and Spilerman (1976) and Frydman (1984). This model relies on a
2Lindbeck and Snower (1986, 2002) summarize theoretical breakthroughs and list key references.3Blanchard and Landier (2002) provide some semantic advice: whereas the French have a specific word designating a
succession of short-term jobs and unemployment spells (précarité), there does not exist an equivalent expression in English.We follow Blanchard and Landier’s suggestion to useinsecurityinstead.
7
mixture of Markov chains which accounts for different dynamic patterns among individuals. Its most
basic version assumes that two kinds of workers coexist on the labor market: while themoverscan move
from unemployment to employment, thestayersremain indefinitely in the state they initially occupy.4
In the version proposed by Kamionka (1996), some workers, namedunconfined movers, can have
access to any kind of jobs whereas some others, calledconfined movers, can only transit between
unemployment, short-term jobs and non-participation. The introduction of different individual types
allows one to account for the so-called partially observed heterogeneity.We use the same partition
but we explicitly let the mixture probabilities (being a mover, confined or unconfined, or being a
stayer) depend on observed characteristics (conditional confined-unconfined model). This allows us
to investigate which individual characteristics are correlated with specific dynamic patterns on the
labor market. In other words, the version that we propose enables us to highlight who the stayers, the
unconfined movers and the confined movers are. Further, the share ofunconfined moversin the economy
and amongst the movers may provide an indicator of the labor market duality level.
Apart from the mover-stayer models, labor market transitions are usually studied using discrete
choice models and/or duration models.Discrete-choice modelsexplain the individual status given
his/her past (and notably his/her past status) and covariates; see for instance Card and Sullivan (1988),
Magnac (2000) or Havet (2006).Duration modelsexplain the duration of a spell in a given state by
the past and a set of individual characteristics; see for example in the French labor market context,
Bonnal, Fougère, and Sérandon (1997) and Magnac and Robin (1994). Duration models capture state
and duration dependence whereas Markov-chain-based approaches account for state dependence and
partially observed heterogeneity. So our study completes previous studiesof the French labor market by
focusing on partially observed heterogeneity.
The discrete time mover-stayer-type model proposed in this paper aims to separate histories when
individuals never accede to stable jobs from histories when individuals have a potential access to both
unstable and stable jobs. The population who experiences confined moverhistories is of first interest
for policy concerns. Moreover, a statistical approachà la Heckman and Singer (1984), which does not
4Mover-stayer models have also been adapted to continuous time Fougèreand Kamionka (1992a, 1992b, 2003, 2008) suchas other Markov chain models Kalbfleish and Lawless (1985), Geweke,Marshall, and Zarkin (1986a, 1986b).
8
require toa priori impose the nature of types and zero-constraints on the transition matrix components,
does not reject the relevance of the partition postulated here. In this alternative approach, the form and
the number of the transition matrices are let free but more structure is imposed on the state dependence.
Transitions are actually modeled by a dynamic multinomial logit with unobserved heterogeneity - which
entails restrictions; see Magnac (2000) and Brodaty (2007).
The model is estimated by maximum likelihood on a sample composed of 30-49 yearsold people who
finished their studies. The data come from the French Labor Force Survey. We focus on middle-aged
people to avoid life-cycle effects that may violate stationarity requirements of Markov Chain models
(labor market entrance of youth, retiring). Our main findings are the following. Individuals trapped into
confined mover histories represent about5% of the total population. This is much less than the13%
computed in summary statistics, showing the relevance of our model to handle heavily censored data.
Individuals falling into theconfined-movercategory are more likely to be less educated, younger and
single. At stationary equilibrium,30% of them occupy unstable jobs, nearly one half are unemployed,
while the remaining do not participate.
The paper is organized as follows. The data and summary statistics are presented in section 2, and the
model in section 3. Section 4 is dedicated to the estimation results. Section 5 contains the results of the
Heckman-Singer approach and some specification tests. Section 6 concludes.
2. Data
The data come from the French Labor Force survey (LFS), 2003-2007, undertaken by Insee, the French
national statistical office. The LFS is a rolling panel in which individuals areinterviewed on their labor
market status, once per quarter, six times. This scheme enables one to construct individual labor market
histories over 15 months. Each quarter, one surveyed individual out of six is replaced. In this paper, we
use the LFS answers of the 30-49 years old individuals who entered the survey from 2003Q1 to 2005Q4,
who finished their studies and who were interviewed 6 times. This panel consists in 33,206 individuals.
The LFS contains information on labor market states - employment, unemploymentand nonparticipation
- as well as a detailed description of the job occupied by the employed. Long-term contracts, short-term
9
contracts, temporary jobs, internships are distinguished. In what follows, we consider four labor market
states: nonparticipation (NP), unemployment (U), unstable job state (UJ) which contains public and
private short-term contracts, temporary jobs and seasonal jobs, and astable job state (SJ) which contains
private long-term contracts, self-employed and civil-servant positions. Unemployment refers to the ILO
definition: unemployed are nonemployed, available to work within two weeks and actively search for a
job. So, non-employed who search for a job are classified as nonparticipants if they do not satisfy the
availability criterion.5 Finally, the panel contains information on individual characteristics, age,gender,
educational level, residential location, family characteristics, etc..
2.1. Summary statistics and representativeness
First, we briefly describe the current French labor market. In 2006, theaverage participation rate
amounted to69% for 15-64 years old,74.5% for men and63.8% for women; see Attal-Toubert and
Lavergne (2006). The French labor market is characterized by a weak participation rate of youth and the
oldest compared to other European countries. This feature is often linkedto the fact that in the 1980’s and
the 1990’s the government and social partners answered to a growing mass unemployment by promoting
early retirements and longer studies. In 2006, around10% of the 15-64 participants were unemployed,
nearly half of them having been unemployed for more than one year.6 Higher unemployment risk is
correlated with: a low level of education, youth, female gender, and blue-collar occupation.13.5%
of the employed occupied an unstable job that is training, apprenticeship, fixed-duration or temporary
contract jobs.
We focus on 30-49 years old people who finished their studies, becausewe are interested in rather
stable labor market histories, once integration is completed and before the retirement process begins.
Furthermore, we ensure stationarity of underlying processes by concentrating on individuals aged be-
tween 30 and 49. The descriptive statistics assessed on the panel data onthe one hand, and on the
pooled LFS 2003Q1-2007Q1 on the other hand, are quite close; see Table 1. Nonetheless, a slight under-
representation of men and of unemployed people can be observed in the balanced panel. This is due
5See Jones and Riddel (2006) for a deep analysis of the frontiers between nonparticipation and ILO unemployment; see alsoFlinn and Heckman (1983).
6The definition of unemployment and the estimation method of the unemployment rate changed in 2007 in France. Thedefinition used here is the one prevailing before 2007.
10
to attrition. Unemployed people usually move more often and they are less likely to be interviewed six
times. The results presented in the sequel are those holding for the panel.
Table 1. Descriptive statistics
Panel subsample LFS whole sample*(obs. 33,206× 6) (obs. 395,077)
(30-49 years old)(Studies finished)
Population% women 52.3 50.4
% men 47.8 49.6Participation rate %
Women 81.2 82.3Men 95.6 95.4Total 88.1 88.8
Unemployment rate %Women 7.5 7.8
Men 5.8 6.4Total 6.7 7.1
Employment rate %Women 73.7 74.5
Men 89.9 88.9Total 81.4 81.7
Share of long-term contracts %Women 86.2 86.2
Men 92.9 91.5Total 89.8 89.1
* pooled analysis.
11
2.2. Descriptive statistics on transitions
The transitions between nonparticipation, unemployment, unstable jobs and stable jobs observed in the
balanced panel are described in Table 2. First,75% of the sample sojourn within long-term jobs or do
not participate during the whole observation period, while only25% experience in-sample transitions.
Second, 19% of men and 27% of women experience one or more transition within the observation period.
Half of them accede to a long-term contract job and half of them transit without acceding to a CDI. So the
apparent ratio of individuals trapped into "contingent work" is 9% for menand 15% for women. Labor
market histories greatly differ between men and women. Women are more likely tobe nonparticipant
during the whole observation period than men (13% versus 20%) and men are more likely to occupy a
stable job during the whole observation period.
Table 2. Data description
Men WomenIndividuals... 15,847 100% 17,359 100%sojourning in long-term jobs 12,497 79% 10,495 60 %staying nonparticipants 382 2% 2,188 13%moving between long-term, short-term jobs and without job spells 1,571 10% 2,065 12%moving between short-term jobs and without job spells only 1,397 9% 2,611 15%
A simple Markov-chain model also provides some insightful summary statistics, see Table 3.
Table 3. Four-state Markov transition matrices
MenT → T + 1 SJ UJ U NPSJ 0.989 0.002 0.005 0.004
(0.000)* (0.000) (0.000) (0.000)UJ 0.084 0.759 0.128 0.029
(0.004) (0.008) (0.007) (0.003)U 0.067 0.147 0.696 0.090
(0.004) (0.006) (0.008) (0.004)NP 0.074 0.024 0.112 0.791
(0.005) (0.003) (0.006) (0.009)
WomenSJ UJ U NP0.983 0.003 0.005 0.009(0.001) (0.000) (0.000) (0.000)0.062 0.785 0.113 0.040(0.003) (0.007) (0.005) (0.002)0.059 0.136 0.669 0.137(0.003) (0.004) (0.006) (0.005)0.037 0.016 0.059 0.887(0.002) (0.001) (0.002) (0.003)
* bootstrapped standard deviations with 50 replicates.
Stable jobs and nonparticipation are the most persistent states:99% and79% of persistence within
three months for men, and98% and89% for women. Around75% of workers with unstable jobs and
12
66% of unemployed remain in the same state three months later.
The propensity to accede to stable jobs is more state-dependent for women than for men.6.7% of
male unemployed,8.4% of male temporary workers and7.4% of male nonparticipants obtain a stable
job within three months whereas5.9% of female unemployed,6.2% of female temporary workers and
only 3.7% of female nonparticipants obtain a stable job within three months. Female nonparticipants are
further away from the labor market than men are.
About 75% of unemployed individuals, whether male or female, transit to employment via a tempo-
rary job (15% versus6% for stable jobs). On the one hand, this underlines the potentially integrating
nature of temporary jobs. Before finding a long-term job, a large part ofthe unemployed go through
temporary jobs. On the other hand, this may also suggest a dual labor market. Unemployed people have
more frequently access to unstable jobs rather than to stable positions. The relationship between unem-
ployment and non-participation is asymmetric for women:14% of the unemployed leave the labor force
each quarter, whereas only6% of the non-participants become unemployed. For men, these proportions
are quite the same:9% of male unemployed exit the labor force,11% of male nonparticipants become
unemployed.
3. Methodology: the conditional confined-unconfined worker model
The former Markov-chain model assumes that labor market transitions aregenerated by the same
underlying process for all individuals. This approach is restrictive in that it does not provide information
on coexisting different dynamic processes. To cover a potential labor market heterogeneity, we turn
to mover-stayer-like models. Mover-stayer models rely on a mixture of Markov chains; see Blumen,
Kogan, and MacCarthy (1955), Goodman (1961), Spilerman (1972), Singer and Spilerman (1976),
Frydman (1984). The model developed in this section extends the version of Kamionka (1996)
Let us considerN individualsi = 1, . . . , N , observed at datest = 0, . . . , T . These individuals can
transit betweenK states relating to their labor market situation (K = 4 in what follows) -stable jobs(1),
short-term jobs(2), unemployment(3) andnonparticipation(4). The individuali experiences a sequence
of states denoted by theT -vector(ei0, . . . , eiT ). Ci denotes the kind of dynamic process generating the
13
transitions experienced by individuali. Four dynamic processes are assumed to exist:stable-job stayer
(S1), nonparticipant stayer(SK), unconfined mover(M ), confined mover(I). The twostayerprocesses
generate histories sojourning indefinitely in the same state and the twomoverprocesses generate histories
with transitions.
• The unconfined-moverprocess corresponds to labor-market histories where individuals canac-
cess to any of theK states, and in particular to stable jobs. Those histories are associated to an
unconstrained Markov chain with transition matrixM = {mij}.
• The confined-moverprocess corresponds to labor-market histories where workers cannot have
access to stable jobs. Formally, the underlying stochastic process is a degenerated Markov chain
with transition matrixQ = {qij}, in which the row and the column components related to the
stable-job state are set to zero.
Furthermore, individuali is endowed with characteristicsXi. The dynamic heterogeneity which is
taken into account by the random variableCi is not observed but is assumed to depend on observables.7
Then,
• pS1(Xi) is the probability to be a stayer in stable jobs (state 1), conditional on starting in state1
and covariatesXi;
• pSK(Xi) is the probability to be a stayer out of the labor market (stateK) , conditional on starting
in stateK and covariatesXi;
• pI(Xi) is the probability to be a confined mover, conditional on not starting in state1 and covariates
Xi, i.e., whether the individual starts in state2, 3, . . . , or K.
The contribution of individuali to the likelihood conditional on the initial state depends on the ob-
served history.
1. When individuali is observed to start in state1, stable job, alternatives cases may occur.
If a transition is observed during the observation period, individuali is, for sure, an unconfined
7Kamionka (1996) describes this dynamic heterogeneity as a partially observed heterogeneity since the labor market histo-ries, partially observed, provide information on the individual types, in contrast with other unobserved individual heterogeneitymethods.
14
mover. His or her contribution to the likelihood is thus:
(
1 − pS1(Xi)
)
T∏
t=1
mit−1it .
If no transition is observed, individuali may either be a stayer in state1 or an unconfined mover.
His/her contribution is:
pS1(Xi) +
(
1 − pS1(Xi)
)
T∏
t=1
mit−1it .
2. When individuali starts in states2 or 3, there are also two options.
If individual i occupies a stable job at least once, then he or she is an unconfined mover. His or her
contribution is:
(1 − pI(Xi))T
∏
t=1
mit−1it .
If individual i does not occupy a stable job during the observation period, then he or she may either
be a confined mover or an unconfined mover. His or her contribution is:
pI(Xi)T
∏
t=1
qit−1it + (1 − pI(Xi))T
∏
t=1
mit−1it .
3. When individuali starts by a nonparticipation spell, three cases may occur.
If individual i occupies once a stable job, then he or she is an unconfined mover. His or her
contribution is:
(1 − pSK(Xi))(1 − pI(Xi))
T∏
t=1
mit−1it .
When individuali does not occupy a stable job during the period, then he or she may be a confined
mover or an unconfined mover. His or her contribution is:
(1 − pSK(Xi))
[
pI(Xi)T
∏
t=1
qit−1it + (1 − pI(Xi))T
∏
t=1
mit−1it
]
.
If individual i remains in stateK, then he or she may be stayer, confined mover or unconfined
mover. His or her contribution is:
pSK(Xi) + (1 − pSK
(Xi))
[
pI(Xi)T
∏
t=1
qit−1it + (1 − pI(Xi))T
∏
t=1
mit−1it
]
.
15
Finally, the conditional likelihood is the product of theN individual contributions. The model is
identified if the number of periods of observation is at least3. The identification relies on the fact that
the stayer transition matrices are set to be the identity matrix and that the individuals who move at least
once in the stable job state are known to be unconfined movers. If they are observed at least three times,
they are supposed to experience the4×4 kinds of transitions, which enables the identification. The model
is estimated by a standard maximum likelihood method. For a more detailed discussionon identification
and consistency of ML estimators, see Kamionka (1996) and Frydman (1984). In practice, the model
is reparameterized to take into account that, in the transition matrices, the exit probabilities belong to
[0, 1] and sum to one by row. The conditional probabilities of being of a given type are modeled by logit
models.
4. Results
The conditional confined-unconfined model is estimated separately on men and women, in order to take
into account gender heterogeneity of labor market dynamics. This approach is justified by a specification
analysis presented in section 5.2.8 The covariates used to explain the conditional probabilities of being
of a given kind are the following: age, marital status, having children, education, residence location (in
Paris region vs. outside, in a distressed area (ZUS) vs. outside). In what follows, a discussion of the
main results is presented. The detailed figures are reported in Appendix C (see Tables 11-15 and Figures
1 and 2).
4.1. Duality in the labor market
First, table 4 reports the probabilities that a worker is of one of the four types.
Table 4. Marginal probabilities for each type, given gender and age.
stayer unconfined confined stayer outin stable job mover mover of the labor market
Women 30-49 51.0% 32.1% 5.2% 11.7%Men 30-49 70.6% 22.5% 4.3% 2.5%
8In section 5.2, we investigate whether the labor market dynamics can be modeled by the same processes for both gender.Tests confirm that transition matrices differ across gender groups.
16
Around 63% of women and 73% of men are stayers, either in stable jobs or out of the labor market,
while the remaining are movers. Confined movers are around 5% of the wholepopulation, but around
15% of the movers.
The 5% figure of confined movers has to be compared to the empirical ratio ofconfined movers found
without accounting for truncation, which amounts to 13% [Tables 11 and 14].9 So the model structure is
successful in controlling for the truncation induced by the 15 months of observation.
Table 5. Average Type-probabilities conditional on initial states
confined stayer in nonparticipant % confinedmover stable job stayer movers|2,3,4
Women 30-49 0.153 0.771 0.610 23%(0.033) (0.014) (0.012)
Men 30-49 0.285 0.832 0.546 36%(0.043) (0.012) (0.018)
* st. errors obtained by bootstrap, using 100 sample replicates.
Table 5 reports the marginal probabilities of being of a given type conditional on the initial state. This
table sums up the last rows of tables 12 and 15 for readability.
4.1.1. The confined movers
The confined-mover population keeps on alternating nonemployment spells with short-term jobs
without being able to accede to a stable position. The confined-mover labor-market histories concern
approximately 29% of the 30-49 men and 15% of the 30-49 women who do not start in a stable job
[Table 5].
Figures 1 and 2 report the densities of the individual probabilities of beingof a given type estimated
on the sample. Their spread provides some insight about the way the included covariates explain the
propensity of being of that type. In a sense it gives an indication on the goodness-of-fit of the model.
When the covariates are poor predictors, the distributions of the predictedprobabilities are expected to
peak around the mean value. Here, on the contrary, they stretch over[0, 1], which indicates that a notable
part of the heterogeneity is explained by the observables.
9i.e., the share of people that never reach stable jobs during the 15 months observed.
17
The effects of the covariates on the probability of being a confined mover are reported in the first
column of Tables 13 and 16. Education is the only relevant variable we find toexplain females probability
to be confined movers: lower degrees tend to be correlated with higher probabilities. Not surprisingly,
for males, education is relevant as well. But family variables also enter significantly: men with lower
probabilities to be confined movers are more likely to be married and to have children. They are also less
likely to live in more distressed areas (ZUS).
4.1.2. The nonparticipant stayers
55% of men who were initially out of the labor market are nonparticipant stayers versus62% for women
(table 5). This illustrates the fact that French women are further from the labor market than are men.
After a nonparticipation spell, women are more likely to stay nonparticipant thanmen; after a long-term
job spell, they are more likely to move to short-term jobs or nonemployment.
The effects of the covariates on the probability of being stayers out of thelabor market are reported in
the third column of tables 13 and 16. Non participants stayers are rather older (being over than forty is
significant for both gender groups), and less educated. However, the degree stratification does not look
the same across groups. Among men, the distinction is between having a degree or no degree at all: the
quality of the degree is not correlated with the probability to stay out of the labor force. Among women,
on the other hand, there seems to exist a strict hierarchy in degrees: women with university degree are
less likely to be non participant than high school graduates, who themselvesare less likely than women
with some elementary or no degree at all. For men, having children is correlated with lower probabilities
to be non participants. The effect of children is more complex for women. Ofcourse, having a 0 to 3
year old child is correlated with higher probabilities to stay our of the labor market. However, having
a 4 to 6 years old child is correlated with lower probabilities to stay our of the labor market. The fact
of being married and living outside the Paris regions are two characteristicsof women who are further
away from the labor market. These results illustrate that family variables affect the female labor market
histories and dynamics whereas their impacts are smaller and more subtle on male ones. They directly
refer to the traditional separation of roles between men and women.
18
4.1.3. The stable-job stayers
Between 30 and 49,83% of men starting in stable jobs are stayers in stable jobs versus77% of women.
The education level has a noticeable impact both for women and for men. Having no degree or a basic
vocational degree seriously reduces the chances of being a stayer in stable jobs. Then, age has a strong
positive effect, indicating that older workers enjoy more stable histories. Adistressed local labor market
has a significant negative impact: living in a ZUS reduces the probability of being a stayer in a stable job
both for 30-49 men and women. Finally, family variables have some, yet less important than for other
probabilities, impact on the probability of being a stayer in stable jobs. Being married is more frequent
for men who are stayers in stable jobs. As expected, having a child aged 0 to6 is correlated with not
being a stayer in a stable jobs.
4.2. Dynamics on the labor market
Four different processes generating labor market transitions are estimated. Two of them are stayer
processes. People experiencing them remain indefinitely in their initial state,i.e. nonparticipation or
long-term job. The two other processes generate labor market transitions. The unconfined-mover process
generates histories in which individuals can access to the four states without restriction. The confined-
mover process generates histories in which individuals cannot access tostable jobs. In this section, the
estimated dynamics are compared.
4.2.1. Confined and unconfined mover transitions
The unconfined-mover-transition process and the confined-mover-transition process clearly describe
different labor market histories. The unconfined-mover-transition process generates histories which refer
much more often to employment states than the confined-mover-transition one. This holds whatever the
gender category.
Table 6 reports the stationary occupation probabilities for each state depending on the underlying
dynamic. This table sums up the results of tables 13 and 16.
A given woman (resp. man) in unconfined-mover dynamics is in employment atthe stationary
19
Table 6. Stationary equilibria
Unconfined equilibrium Confined equilibriumSJ UJ U UNP UJ U UNP
Women 30-49 0.585 0.143 0.134 0.1380.299 0.403 0.297(0.023) (0.009) (0.011) (0.013) (0.059) (0.044) (0.069)
Men 30-49 0.680 0.143 0.122 0.055 0.289 0.541 0.169(0.026) (0.012) (0.014) (0.006) (0.054) (0.042) (0.032)
*bootstrap standard errors using 100 sample replicates
equilibrium with a probability of 73% (resp. 82%). For a woman (resp. man) inconfined-dynamics,
this probability is only 30% (resp. 29%). Therefore, being employed is twiceas likely for individuals
in unconfined dynamics than for those in confined ones. Furthermore, theunconfined-mover-transition
process generates histories which refer slightly more often to participation than the confined-mover one.
At equilibrium, unconfined males (resp. females) are 94% (resp. 86%) to participate, versus 83% (resp.
70%) of confined males (females).
These results suggest that the main part of the difference between the unconfined and the confined-
mover dynamics cannot be explained by an underlying difference in participation behaviors. This
difference is rather explained by the fact that people in confined-moverdynamics more often experience
difficult episodes on the labor market such as unemployment than people in unconfined-mover dynamics.
This is obvious when the unemployment probability is examined (around 12% for unconfined movers
versus 40% to 54% for confined movers).
The parameters in transition matrices stress the unemployment risk faced by individuals with
confined-mover histories. Around 30% of individuals initially in unstable jobsand with a confined-
mover dynamic would experience unemployment three months later versus around 7% of those with
unconfined-mover dynamics. 38% of nonparticipant men with a confined-mover history would become
unemployed three months later versus 20% of those with a unconfined-moverhistory. These remarks
hold also for women.
The male and the female unconfined-mover dynamics are significantly different, as shown by speci-
fication tests. Men in confined dynamics tend to be less mobile than women and display, for example,
20
higher persistence in unemployment. The picture is rather different for unconfined dynamics. Uncon-
fined males are more mobile than women and they more often get access to employment, both to stable
and unstable jobs.
5. Robustness analysis
5.1. Heckman-Singer approach
In mover-stayer-type models, the form of the heterogeneity is imposedex anteby the model (i.e., stayers,
unconfined movers, confined movers). In this section, we adopt an alternative model, which does not
require to fixa priori the nature of types and constraints on transition matrices, in order to see whether
the entailed partition shares common features with the one we proposed. We follow the approach of
Brodaty (2007) which is inspired by Magnac (2000) and Heckman and Singer (1984). Transitions are
modeled by a dynamic multinomial logit with unobserved heterogeneity. Letyit denote the labor market
state occupied by the individuali at datet, then
yit = k if only if y∗ikt = maxj=1,...,4
(y∗ijt), ∀(i, t), (5.1)
where
y∗ijt =
4∑
j=1
δjkIyi,t−1=j + αik + ǫikt, ∀(i, t). (5.2)
δjk’s account for dependence from the lag state (state dependence),αik’s is the unobserved hetero-
geneity term. This unobserved propensity to move from one state to another istype-specific,i.e.
αi = (αi1, . . . , αi4)′ can takeT different values,α1, . . . , αT , whereT is the number of types present in
the model. In order to identify the model, some parameters are set to 0:
δj4 = 0,∀j = 1, . . . , 4,
δ4j = 0,∀j = 1, . . . , 4,
αl4 = 0,∀l = 1, . . . , T.
Further, theǫ’s are type-I extreme value distributed, independent across alternatives, individuals, time
and independent of theα’s. Then, the probability that individuali goes to statek at timet conditional on
21
being in statej at t − 1 is given by
P (yit = k|yit−1 = j, αi = αl) =exp(δjk + αik)
∑4m=1 exp(δjm + αim)
. (5.3)
It depends on his unobserved typel. Contrary to mover-stayer model, noa priori 0-constraints are
imposed on the transition matrix components, but the dynamic multinomial logit model implies that the
odds ratios satisfy the following constraints:
P(SJ |State=k,Type=l)P(SJ |State=j,Type=l)
P(UJ |State=k,Type=l)P(UJ |State=j,Type=l)
=
P(SJ |State=k,Type=l′)P(SJ |State=j,Type=l′)
P(UJ |State=k,Type=l′)P(UJ |State=j,Type=l′)
, ∀(j, k, l, l′). (5.4)
Finally, there is no reason why the dynamic multinomial logit model would be more orless flexible than
the mover-stayer approach.
This model is estimated sequentially. The first step consists in a conditional maximum likelihood
estimation that yields consistent estimates of the state dependence parameters.The type-specific terms
are estimated in a second step using an EM algorithm, given the first-stage estimates. The number of
types is determined iteratively. The initial condition problem is tackled by using alikelihood conditional
on initial states (Brodaty, 2007). Hence, the probability of being of typer depends on the individual
initial state. The number of types is determined iteratively.
For men as well as for women, the iterative procedure suggests to retain a partition in five categories.
Table 7 details the probabilities of these five types conditional on the four possible initial states. Women
who are initially out of the labor force have a high probability to be type-1 individuals (60%) or type-5
individuals (28%). Probabilities for men are not as clear cut: when they start out of the labor force,
they tend to be type-1 (57%) and type-5 (18%), but also of type-2 and type-4 (11% each). Almost all
individuals starting in stable jobs are type-2. Conditional on starting in unstable jobs, men are mainly
type-3 (84%) and type-4 (13%). Women, apart from type-3 (83%) andtype-4 (10%) are, in fewer cases,
type-5 (6%). Finally, the types of men and women starting in unemployment are similarly distributed:
mainly type-4 (around 65%), then type-3 (around 20%) and type-5 (around 15%).
Table 8 contains the estimates of the transition matrices associated to each type. These matrices and
22
Table 7. Types Distribution
Initial StateWomen SJ UJ U NPType 1 0.002
(0.001)0.000(0.003)
0.009(0.007)
0.601(0.024)
Type 2 0.907(0.009)
0.019(0.008)
0.011(0.007)
0.044(0.005)
Type 3 0.020(0.004)
0.831(0.018)
0.220(0.017)
0.024(0.003)
Type 4 0.018(0.003)
0.094(0.013)
0.611(0.021)
0.056(0.007)
Type 5 0.053(0.010)
0.056(0.013)
0.148(0.020)
0.275(0.024)
Initial StateMen SJ UJ U NP
Type 1 0.000(0.001)
0.010(0.004)
0.000(0.004)
0.572(0.031)
Type 2 0.925(0.013)
0.020(0.009)
0.000(0.002)
0.106(0.018)
Type 3 0.025(0.004)
0.839(0.052)
0.207(0.026)
0.030(0.010)
Type 4 0.015(0.004)
0.131(0.017)
0.657(0.037)
0.114(0.025)
Type 5 0.035(0.012)
0.000(0.049)
0.136(0.029)
0.179(0.044)
Note: Each column sums to one. Standard er-rors are reported in parentheses.
the heterogeneity distribution have to be analyzed together. They are usedto give an interpretation of
the individual types that were found and to compare them with the types imposed in the mover-stayer
approach. Standard errors are obtained by bootstrap using 70 replicates. Table 9 contains the stationary
occupation probabilities for each type.
• Type-1 transition matrix exhibits high transition probabilities to non-participation.Moreover, in-
dividuals who are initially out of the labor market are mainly of type-1. Thesetwo points enable
us to interpret quite unambiguously type-1- individual as “stayers out ofthe labor market”.
• The same kind of argument can be used to assert that individuals followingtype-2 process are
“stayers in stable jobs”.
• The three last types are more intricate. Individuals of type-3 are often in unstable jobs. They have
low conditional probabilities to accede to a stable job whatever their departing state: less than
10% when they have an unstable job the period before, and even lower for women (6%) than for
men (9%); between 8% and 13% when they come from unemployment or non-participation. Thus,
individuals of type-3 are close to confined movers.
23
• Type-4 individuals are mainly unemployed. When they get exit unemployment,they are more
likely to find unstable jobs if they are male (10%) and to exit the labor market if they are female
(8%). Their probability to exit unemployment obtaining a stable job is very low around 5%, both
for men and for women. Further, obtaining a stable job is unlikely, whatever the departing state:
the transition probabilities never exceed 10%. And this holds also for men andwomen. Type 4 is
the closest to the confined mover-type.
• Finally, type-5 individuals have relatively strong probabilities to accede to astable job, whatever
the state they start in (around 25% for men, and 15% for women). However, they almost never pass
through unstable job or unemployment spells. They obviously can be interpreted as unconfined
movers.
24
Table 8. Transition matrices according to types
Men
First type
0.243(0.129)
0.006(0.002)
0.010(0.004)
0.741(0.129)
0.031(0.028)
0.074(0.028)
0.024(0.013)
0.871(0.046)
0.015(0.013)
0.019(0.007)
0.034(0.019)
0.931(0.029)
0.006(0.005)
0.005(0.002)
0.009(0.004)
0.980(0.007)
Second type
0.998(0.000)
0.000(0.000)
0.001(0.000)
0.001(0.000)
0.950(0.018)
0.032(0.016)
0.011(0.004)
0.008(0.004)
0.933(0.024)
0.017(0.009)
0.033(0.015)
0.017(0.008)
0.924(0.031)
0.011(0.006)
0.020(0.009)
0.044(0.023)
Third type
0.856(0.022)
0.079(0.016)
0.046(0.007)
0.019(0.004)
0.086(0.009)
0.810(0.008)
0.086(0.006)
0.018(0.003)
0.103(0.016)
0.534(0.028)
0.316(0.025)
0.048(0.009)
0.132(0.029)
0.454(0.046)
0.254(0.029)
0.161(0.028)
Fourth type
0.710(0.046)
0.027(0.006)
0.213(0.039)
0.050(0.010)
0.089(0.017)
0.346(0.029)
0.507(0.026)
0.058(0.012)
0.046(0.007)
0.097(0.007)
0.790(0.013)
0.067(0.008)
0.058(0.013)
0.083(0.013)
0.635(0.039)
0.224(0.039)
Fifth type
0.902(0.033)
0.002(0.018)
0.029(0.013)
0.068(0.016)
0.399(0.083)
0.083(0.113)
0.243(0.041)
0.275(0.061)
0.221(0.052)
0.025(0.067)
0.411(0.048)
0.343(0.062)
0.159(0.039)
0.012(0.056)
0.185(0.039)
0.645(0.072)
Women
First type
0.000(0.065)
0.000(0.001)
0.005(0.003)
0.995(0.066)
0.000(0.010)
0.000(0.008)
0.009(0.007)
0.991(0.019)
0.000(0.005)
0.000(0.002)
0.014(0.010)
0.986(0.015)
0.000(0.002)
0.000(0.000)
0.003(0.002)
0.997(0.004)
Second type
0.998(0.001)
0.000(0.000)
0.000(0.000)
0.001(0.000)
0.956(0.016)
0.021(0.013)
0.010(0.004)
0.014(0.005)
0.940(0.019)
0.009(0.006)
0.026(0.011)
0.025(0.009)
0.921(0.024)
0.006(0.004)
0.014(0.006)
0.058(0.021)
Third type
0.804(0.029)
0.108(0.020)
0.060(0.009)
0.028(0.004)
0.059(0.004)
0.828(0.007)
0.089(0.004)
0.024(0.002)
0.082(0.011)
0.512(0.022)
0.344(0.019)
0.062(0.006)
0.105(0.014)
0.457(0.027)
0.244(0.016)
0.194(0.019)
Fourth type
0.726(0.032)
0.023(0.004)
0.196(0.026)
0.054(0.008)
0.094(0.014)
0.312(0.024)
0.513(0.023)
0.082(0.010)
0.052(0.004)
0.077(0.006)
0.788(0.009)
0.006(0.028)
0.070(0.010)
0.072(0.007)
0.585(0.022)
0.273(0.023)
Fifth type
0.830(0.020)
0.007(0.002)
0.030(0.004)
0.133(0.018)
0.223(0.028)
0.194(0.038)
0.164(0.013)
0.418(0.043)
0.145(0.017)
0.056(0.016)
0.296(0.021)
0.503(0.034)
0.093(0.011)
0.025(0.009)
0.104(0.009)
0.778(0.023)
Note: bootstrapped standard errors are reported in parentheses.
25
Table 9 Limiting Probabilities
Men SJ UJ U NP
Type 1 0.008(0.014)
0.006(0.002)
0.009(0.005)
0.977(0.017)
Type 2 0.998(0.001)
0.000(0.000)
0.001(0.000)
0.001(0.000)
Type 3 0.386(0.049)
0.492(0.040)
0.097(0.010)
0.025(0.004)
Type 4 0.154(0.042)
0.114(0.012)
0.658(0.038)
0.075(0.010)
Type 5 0.650(0.107)
0.007(0.088)
0.108(0.041)
0.234(0.056)
Women SJ UJ U NP
Type 1 0.000(0.003)
0.000(0.001)
0.003(0.002)
0.997(0.005)
Type 2 0.998(0.001)
0.000(0.000)
0.001(0.000)
0.001(0.000)
Type 3 0.247(0.035)
0.600(0.031)
0.117(0.007)
0.036(0.003)
Type 4 0.176(0.025)
0.087(0.008)
0.640(0.022)
0.097(0.009)
Type 5 0.384(0.049)
0.025(0.009)
0.096(0.009)
0.495(0.051)
Note: each row sums to one. Bootstrapped standard errors are reported in parentheses.
This alternative analysis does support the relevance of the mover-stayer-confined partition. First,
nonparticipant stayers and stayers in stable jobs appear clearly. Resultsare less clear-cut for the mover
categories, since individuals of all categories have a chance to obtain a stable job. This could be a
consequence of the structure imposed by the model. But for two types, probabilities to obtain a stable
job turns out to be rather small. Hence the results obtained here underline that clear differences exist in
the transition dynamics, and that splitting movers into confined/unconfined categories is relevant.
5.2. Specification tests: stability of transition matricesacross gender
In section 4 we focused on the results of separate estimations on sub-samples by gender. This was
justified by the results of the present section, in which we test whether, once controlling for conditional
heterogeneity, the transition dynamics on the labor market are the same for menand women. To do this,
we consider three testing hypotheses.
• H10 : both confined and unconfined-mover transition matrices are stable across gender,
• H20 : the unconfined-mover transition matrix is stable across gender,
• H30 : the confined-mover transition matrix is stable across gender.
H10 can be tested by a classical LR test: we estimate the model on men and women separately (M1)
and simultaneously with adequate covariates (M0), and compute a LR statistic. For testingH20 andH3
0 ,
we use aχ2-statistic (denotedDA, hereafter) based on the difference of the estimates between the two
26
Table 10. Tests for stability of dynamics across gender (p-values).
Null hypothesis H10 H2
0 H30
30-49 0.000 0.000 0.513Note: Test statistics LR is the first column and DA for the second and the thirdones. p-values are computed usingχ2 distributions. Degrees of freedom areresp. 18 (24 and 12) for the first (second and third) columns.
groups which are assumed to be independent (the method is described in details in appendix B). Results
are reported in Table 10. The stability of the labor market dynamics across gender is rejected due to
different unconfined-mover dynamics whereas the stability of the confined-mover dynamics cannot be
rejected. For the latter, labor market histories differences can be explained conditionally, by differences
in covariates.
6. Conclusion
The model used in this paper is based on a Markov-chain mixture of four types of transition dynamics:
the stayers in stable-jobs, the stayers in nonparticipation, the unconfined movers, and the individuals
stuck onconfinedstates and who cannot accede to stable jobs. This partition enables us to specifically
account for heterogenous abilities to accede to stable jobs. The probabilities of being of a given type also
depend on observable individual characteristics. The data come from the French Labor Force Survey, the
model is estimated on the 30-49 years old individuals.
The main results are the following. Individuals who are trapped in confinedmover histories represent
around 5% of the 30-49 years old population (versus 13% apparently observed). At equilibrium, partic-
ipation rates of the confined and the unconfined populations are similar but an individual whose labor
market history is generated by the confined mover process has between 3and 4 times more chances to be
unemployed than a confined mover. The probability to be a confined mover decreases with the education
level. Male confined movers are also more likely to be single, and to live in a distressed area whereas
for women, only the education seems to matter. Finally, unconfined-mover dynamics depend on gender,
whereas male and female confined mover dynamics are not significantly different.
27
APPENDIX
A. Stationary occupation probabilities
Confined-unconfined models, just like mover-stayer models, satisfy the Markov assumption conditional
on the initial state. The stationary occupation probability vector represents the probabilities associated
to each state once the process converged to the steady state and can be defined for any Markov-chain
process. Let us consider a Markov-chain process with transition matrixA. The stationary occupation
probability vector, denoteda∗, is defined such that it is invariant by pre-multiplication by the transition
matrix:
A′a∗ = a∗. (A.5)
Moreover, it is a vector of probabilities. Hence, its components remain in[0, 1] and sum to one. The
stationary occupation probability vector is a useful tool to describe how much labor market histories
generated by a given dynamic are confined in some states.
Stationary occupation probabilities (conditional on the initial values) are easily extended to mixtures
of Markov chains by:
pMm∗ + pQq∗ + pS1s∗1 + pSK s∗K ,
wherem∗, q∗, s∗1 and s∗K are the stationary probability vectors (as defined in A.5) relating to each
elementary Markov chain, andpM , pQ, pS1 , andpSK , the mixture coefficients relating to each elementary
Markov chain. In the conditional confined-unconfined model, sample stationary occupation probability
vector can be estimated by the sample average of the weighted sum of the stationary probability vectors
of each elementary Markov process composing the mixture.
1
N
N∑
i=1
pMi m∗ + p
Qi q∗ + pS1
i s∗1 + pSK
i s∗K ,
wherem∗, q∗, s∗1 and s∗K are the stationary probability vectors (as defined in A.5) relating to each
elementary Markov chain, andpMi , p
Qi , pS1
i , andpSK
i , the individual probabilities of following each
elementary Markov chain. Note thatpMi , p
Qi , pS1
i , pSK
i sum to one.
28
B. Darmois-type test for coefficient equality across subsamples
The idea is the same as the the classical Darmois test for testing the equality of themeans in two sub-
samples with unknown different variances; see Darmois (1954). Data is composed of two samples:
sample 1, withn1 observations{Y1i }i=1,...,n1
whose distribution is function of the parameter of in-
terestβ1 ∈ Rk; and sample 2, withn2 observations{Y2
i }i=1,...,n2whose distribution is function of
β2 ∈ Rk. {Y1
i }i=1,...,n1and{Y2
i }i=1,...,n2are independent and both composed ofi.i.d. observations.̂β1
(resp. β̂2) denotes the estimate ofβ1 (resp. β2) based on sample 1 (resp. sample 2). Consider testing
H0 : β1 = β2 = β0 againstH1 : β1 6= β2. Assume that CLT theorems apply forβ1 andβ2, i.e. under
H0:
√n1(β̂1 − β0) → N (0, Vas(β̂1)) (B.6)
√n2(β̂2 − β0) → N (0, Vas(β̂2)) (B.7)
andβ̂1 andβ̂2 are independent. Hence, it follows that underH0,
DA = (β̂1 − β̂2)′
(
1
n1Vas(β̂1) +
1
n2Vas(β̂2)
)−1
(β̂1 − β̂2) → χ2(2k). (B.8)
A test forH0 with asymptotic levelα rejectsH0 whenDA > c1−α, wherec1−α is the 1-α quantile of a
χ2 distribution with2k degrees of freedom.
C. Detailed results
C.1. Women between 30 and 49
Women between 30 and 49, 4 states : out of the labor market (4), unemployed (3), short-term contract (2),
long-term contract (1). Asymptotic standard-errors are obtained by bootstrap (design matrix bootstrap
centered around the sample estimate) with 100 sample replicates.
Table 11 describes the observed histories.
29
Table 11. Data : women between 30 and 49.Individuals... 17,359 100%staying in 1 10,495 60 %staying in 4 2,188 13%moving between 1, 2, 3 and 4 2,065 12%moving between 2, 3 and 4 only 2,611 15%
Figure 1. Densities of individual probabilities of being stayer in 1, stayer in4 and confined conditionalon initial states
0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 10
1
2
3
4
5
Bandwiths: kst1
=kst4
=kins
=0.02
Den
sitie
s
stayer 1 (| state1=1) stayer 4 (| state1=4) insecure (| state1=2,3,4)
C.2. Men between 30 and 49
Men between 30 and 49, 4 states : out of the labor market (4), unemployed(3), short-term contract (2),
long-term contract (1). Asymptotic standard-errors are obtained by bootstrap (design matrix bootstrap
centered around the sample estimate) with 100 sample replicates.
Table 14 describes the observed histories.
30
Table 12. Coefficients: women between 30 and 49.Estimates - MLE
Covariates confined mover stayer in stable job stayer in non-participationintercept -1.244 0.931 -1.187
(1.409) (0.359) (0.626)30-39 - - -
- - -40-49 0.124 0.640 0.576
(0.179) (0.101) (0.108)married -0.234 0.125 0.627
(0.188) (0.096) (0.096)university degree 0.386 0.172 -0.453(bac+3 and more) (0.357) (0.218) (0.211)college degree or more 0.141 0.174 -0.229(bac+2 and more) (0.475) (0.172) (0.213)completed high school - - -(bac) - - -basic vocational degree 0.622 -0.359 0.096
(0.299) (0.142) (0.141)elementary high school 0.236 0.054 0.504
(0.380) (0.211) (0.165)no degree 1.081 -0.994 0.835
(0.326) (0.139) (0.154)ZUS 0.262 -0.421 0.190
(0.385) (0.231) (0.121)Paris -0.416 -0.153 -0.312
(0.288) (0.135) (0.135)one 0-18 year-old child or more 0.258 -0.154 -0.395
(0.205) (0.104) (0.111)one 3- 6 year-old child or more -0.038 -0.601 0.607
(0.643) (0.155) (0.137)one 0- 3 year-old child or more -0.114 -0.282 -0.365
(0.252) (0.126) (0.121)Experience above 7 years -0.636 0.397 0.905
(1.336) (0.307) ( 0.632)Average conditional 0.153 0.771 0.610probability (0.033) (0.014) (0.012)
Asymptotic standard errors estimates are obtained by design matrix bootstrap centered around the sample estimate with 100
sample replicates.
31
Table 13. Transition matrices: women between 30 and 49.
Unconfined transition matrix:
T → T + 1 SJ UJ U NPSJ (k = 1) 0.931 0.011 0.023 0.036
(0.005) (0.001) (0.002) (0.003)UJ (k = 2) 0.079 0.836 0.061 0.023
(0.006) (0.033) (0.024) (0.008)U (k = 3) 0.082 0.099 0.720 0.099
(0.009) (0.025) (0.030) (0.019)NP (k = 4) 0.133 0.029 0.113 0.724
(0.014) (0.008) (0.016) (0.019)boot. st. err. : 100 replicates
Confined transition matrix:
T → T + 1 SJ UJ U NPSJ (k = 1) 0.000 0.000 0.000 0.000
0.000 0.000 0.000 0.000UJ (k = 2) 0.000 0.600 0.302 0.098
0.000 (0.090) (0.074) (0.025)U (k = 3) 0.000 0.227 0.540 0.233
0.000 (0.076) (0.070) (0.058)NP(k = 4) 0.000 0.094 0.320 0.586
0.000 (0.080) (0.063) (0.096)boot. st. err. : 100 replicates
Stationary equilibria:
SJ UJ U UNPUnconfined equilibrium 0.585 0.143 0.134 0.138
(0.023) (0.009) (0.011) (0.013)Confined equilibrium 0.000 0.299 0.403 0.297
(0.000) (0.059) (0.044) (0.069)Total equilibrium 0.698 0.062 0.064 0.177
(0.004) (0.002) (0.002) (0.003)boot. st. err. : 100 replicates
32
Table 14. Data : men between 30 and 49.Individuals... 15,847 100%staying in 1 12,497 79%staying in 4 382 2%moving between 1, 2, 3 and 4 1,571 10%moving between 2, 3 and 4 only 1,397 9%
Figure 2. Densities of individual probabilities of being stayer in 1, stayer in4 and confined conditionalon initial states
0 0.1 0.2 0.3 0.4 0.5 0.6 0.7 0.8 0.9 10
1
2
3
4
5
6
7
Bandwiths: kst1
=kst4
=kins
=0.03
Den
sitie
s
stayer 1 (| state1=1) stayer 4 (| state1=4) insecure (| state1=2,3,4)
33
Table 15. Coefficients: men between 30 and 49.Estimates - MLE
Covariates confined mover stayer in stable job stayer in non-participationintercept -0.480 1.141 0.892
(3.666) (0.391) (0.826)30-39 - - -
- - -40-49 0.028 0.475 0.832
(0.233) (0.094) (0.182)married -0.870 0.347 0.038
(0.245) (0.096) (0.209)university degree -0.019 0.089 -0.146(bac+3 and more) (1.930) (0.175) (0.415)college degree or more 0.270 0.306 -0.281(bac+2 and more) (0.484) (0.243) (0.446)completed high school - - -(bac) - - -basic vocational degree 0.361 -0.145 0.179
(0.447) (0.146) (0.306)elementary high school 1.257 -0.049 0.481
(0.616) (0.220) (0.353)no degree 1.391 -0.448 0.784
(0.486) (0.162) (0.314)ZUS 0.872 -0.644 0.312
(0.311) (0.211) (0.285)Paris -0.564 -0.123 -0.210
(0.494) (0.132) (0.332)one 0-18 year-old child or more -0.957 -0.062 -1.053
(0.253) (0.108) (0.225)one 0- 3 year-old child or more 0.069 0.057 -0.312
(0.374) (0.138) (0.408)Experience above 7 years -0.187 0.216 -1.108
(3.533) (0.370) (0.747)Average conditional 0.285 0.832 0.546probability (0.043) (0.012) (0.018)
Asymptotic standard errors estimates are obtained by design matrix bootstrap centered around the sample estimate with 100
sample replicates.
34
Table 16. Transition matrices: men between 30 and 49.
Unconfined transition matrix:
T → T + 1 SJ UJ U NPSJ (k = 1) 0.934 0.014 0.029 0.023
(0.005) (0.001) (0.002) (0.002)UJ (k = 2) 0.115 0.800 0.069 0.016
(0.011) (0.032) (0.028) (0.005)U (k = 3) 0.118 0.136 0.670 0.076
(0.012) (0.025) (0.024) (0.009)NP (k = 4) 0.256 0.045 0.197 0.502
(0.023) (0.012) (0.020) (0.030)boot. st. err.: 100 replicates
Confined transition matrix:
T → T + 1 SJ UJ U NPSJ (k = 1) 0.000 0.000 0.000 0.000
(0.000) (0.000) (0.000) (0.000)UJ (k = 2) 0.000 0.650 0.286 0.064
(0.000) (0.098) (0.091) (0.016)U (k = 3) 0.000 0.163 0.729 0.108
(0.000) (0.052) (0.044) (0.018)NP(k = 4) 0.000 0.077 0.378 0.545
(0.000) (0.037) (0.065) (0.079)boot. st. err.: 100 replicates
Stationary equilibria:
SJ UJ U UNPUnconfined equilibrium 0.680 0.143 0.122 0.055
(0.026) (0.012) (0.014) (0.006)Confined equilibrium 0.000 0.289 0.541 0.169
(0.000) (0.054) (0.042) (0.032)Total equilibrium 0.860 0.045 0.051 0.045
(0.003) (0.002) (0.002) (0.002)boot. st. err.: 100 replicates
35
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39
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reexamination on matched worker-firm panel data
G 9506 S. JACOBZONE Les liens entre RMI et chômage, une mise en perspective NON PARU - article sorti dans Economie et Prévision n° 122 (1996) - pages 95 à 113
G 9507 G. CETTE - S. MAHFOUZ Le partage primaire du revenu Constat descriptif sur longue période
G 9601 Banque de France - CEPREMAP - Direction de la Prévision - Erasme - INSEE - OFCE Structures et propriétés de cinq modèles macro-économiques français
G 9602 Rapport d’activité de la DESE de l’année 1995
G 9603 J. BOURDIEU - A. DRAZNIEKS L’octroi de crédit aux PME : une analyse à partir d’informations bancaires
G 9604 A. TOPIOL-BENSAÏD Les implantations japonaises en France
G 9605 P. GENIER - S. JACOBZONE Comportements de prévention, consommation d’alcool et tabagie : peut-on parler d’une gestion globale du capital santé ? Une modélisation microéconométrique empirique
G 9606 C. DOZ - F. LENGLART Factor analysis and unobserved component models: an application to the study of French business surveys
G 9607 N. GREENAN - D. GUELLEC La théorie coopérative de la firme
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G 9608 N. GREENAN - D. GUELLEC Technological innovation and employment reallocation
G 9609 Ph. COUR - F. RUPPRECHT L’intégration asymétrique au sein du continent américain : un essai de modélisation
G 9610 S. DUCHENE - G. FORGEOT - A. JACQUOT Analyse des évolutions récentes de la producti-vité apparente du travail
G 9611 X. BONNET - S. MAHFOUZ The influence of different specifications of wages-prices spirals on the measure of the NAIRU : the case of France
G 9612 PH. COUR - E. DUBOIS, S. MAHFOUZ, J. PISANI-FERRY The cost of fiscal retrenchment revisited: how strong is the evidence ?
G 9613 A. JACQUOT Les flexions des taux d’activité sont-elles seule-ment conjoncturelles ?
G 9614 ZHANG Yingxiang - SONG Xueqing Lexique macroéconomique Français-Chinois
G 9701 J.L. SCHNEIDER La taxe professionnelle : éléments de cadrage économique
G 9702 J.L. SCHNEIDER Transition et stabilité politique d’un système redistributif
G 9703 D. GOUX - E. MAURIN Train or Pay: Does it Reduce Inequalities to En-courage Firms to Train their Workers?
G 9704 P. GENIER Deux contributions sur dépendance et équité
G 9705 E. DUGUET - N. IUNG R & D Investment, Patent Life and Patent Value An Econometric Analysis at the Firm Level
G 9706 M. HOUDEBINE - A. TOPIOL-BENSAÏD Les entreprises internationales en France : une analyse à partir de données individuelles
G 9707 M. HOUDEBINE Polarisation des activités et spécialisation des départements en France
G 9708 E. DUGUET - N. GREENAN Le biais technologique : une analyse sur données individuelles
G 9709 J.L. BRILLET Analyzing a small French ECM Model
G 9710 J.L. BRILLET Formalizing the transition process : scenarios for capital accumulation
G 9711 G. FORGEOT - J. GAUTIÉ Insertion professionnelle des jeunes et processus de déclassement
G 9712 E. DUBOIS High Real Interest Rates: the Consequence of a Saving Investment Disequilibrium or of an in-sufficient Credibility of Monetary Authorities?
G 9713 Bilan des activités de la Direction des Etudes et Synthèses Economiques - 1996
G 9714 F. LEQUILLER Does the French Consumer Price Index Over-state Inflation?
G 9715 X. BONNET Peut-on mettre en évidence les rigidités à la baisse des salaires nominaux ? Une étude sur quelques grands pays de l’OCDE
G 9716 N. IUNG - F. RUPPRECHT Productivité de la recherche et rendements d’échelle dans le secteur pharmaceutique français
G 9717 E. DUGUET - I. KABLA Appropriation strategy and the motivations to use the patent system in France - An econometric analysis at the firm level
G 9718 L.P. PELÉ - P. RALLE Âge de la retraite : les aspects incitatifs du régime général
G 9719 ZHANG Yingxiang - SONG Xueqing Lexique macroéconomique français-chinois, chinois-français
G 9720 M. HOUDEBINE - J.L. SCHNEIDER Mesurer l’influence de la fiscalité sur la locali-sation des entreprises
G 9721 A. MOUROUGANE Crédibilité, indépendance et politique monétaire Une revue de la littérature
G 9722 P. AUGERAUD - L. BRIOT Les données comptables d’entreprises Le système intermédiaire d’entreprises Passage des données individuelles aux données sectorielles
G 9723 P. AUGERAUD - J.E. CHAPRON Using Business Accounts for Compiling National Accounts: the French Experience
G 9724 P. AUGERAUD Les comptes d’entreprise par activités - Le pas-sage aux comptes - De la comptabilité d’entreprise à la comptabilité nationale - A paraître
G 9801 H. MICHAUDON - C. PRIGENT Présentation du modèle AMADEUS
G 9802 J. ACCARDO Une étude de comptabilité générationnelle pour la France en 1996
G 9803 X. BONNET - S. DUCHÊNE Apports et limites de la modélisation « Real Business Cycles »
G 9804 C. BARLET - C. DUGUET - D. ENCAOUA - J. PRADEL The Commercial Sucess of Innovations An econometric analysis at the firm level in French manufacturing
G 9805 P. CAHUC - Ch. GIANELLA - D. GOUX - A. ZILBERBERG Equalizing Wage Differences and Bargaining Power - Evidence form a Panel of French Firms
G 9806 J. ACCARDO - M. JLASSI La productivité globale des facteurs entre 1975 et 1996
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G 9807 Bilan des activités de la Direction des Etudes et Synthèses Economiques - 1997
G 9808 A. MOUROUGANE Can a Conservative Governor Conduct an Ac-comodative Monetary Policy ?
G 9809 X. BONNET - E. DUBOIS - L. FAUVET Asymétrie des inflations relatives et menus costs : tests sur l’inflation française
G 9810 E. DUGUET - N. IUNG Sales and Advertising with Spillovers at the firm level: Estimation of a Dynamic Structural Model on Panel Data
G 9811 J.P. BERTHIER Congestion urbaine : un modèle de trafic de pointe à courbe débit-vitesse et demande élastique
G 9812 C. PRIGENT La part des salaires dans la valeur ajoutée : une approche macroéconomique
G 9813 A.Th. AERTS L’évolution de la part des salaires dans la valeur ajoutée en France reflète-t-elle les évolutions individuelles sur la période 1979-1994 ?
G 9814 B. SALANIÉ Guide pratique des séries non-stationnaires
G 9901 S. DUCHÊNE - A. JACQUOT Une croissance plus riche en emplois depuis le début de la décennie ? Une analyse en compa-raison internationale
G 9902 Ch. COLIN Modélisation des carrières dans Destinie
G 9903 Ch. COLIN Evolution de la dispersion des salaires : un essai de prospective par microsimulation
G 9904 B. CREPON - N. IUNG Innovation, emploi et performances
G 9905 B. CREPON - Ch. GIANELLA Wages inequalities in France 1969-1992 An application of quantile regression techniques
G 9906 C. BONNET - R. MAHIEU Microsimulation techniques applied to inter-generational transfers - Pensions in a dynamic framework: the case of France
G 9907 F. ROSENWALD L’impact des contraintes financières dans la dé-cision d’investissement
G 9908 Bilan des activités de la DESE - 1998
G 9909 J.P. ZOYEM Contrat d’insertion et sortie du RMI Evaluation des effets d’une politique sociale
G 9910 Ch. COLIN - Fl. LEGROS - R. MAHIEU Bilans contributifs comparés des régimes de retraite du secteur privé et de la fonction publique
G 9911 G. LAROQUE - B. SALANIÉ Une décomposition du non-emploi en France
G 9912 B. SALANIÉ Une maquette analytique de long terme du marché du travail
G 9912 Ch. GIANELLA
Bis Une estimation de l’élasticité de l’emploi peu qualifié à son coût
G 9913 Division « Redistribution et Politiques Sociales » Le modèle de microsimulation dynamique DESTINIE
G 9914 E. DUGUET Macro-commandes SAS pour l’économétrie des panels et des variables qualitatives
G 9915 R. DUHAUTOIS Evolution des flux d’emplois en France entre 1990 et 1996 : une étude empirique à partir du fichier des bénéfices réels normaux (BRN)
G 9916 J.Y. FOURNIER Extraction du cycle des afffaires : la méthode de Baxter et King
G 9917 B. CRÉPON - R. DESPLATZ - J. MAIRESSE Estimating price cost margins, scale economies and workers’ bargaining power at the firm level
G 9918 Ch. GIANELLA - Ph. LAGARDE Productivity of hours in the aggregate production function: an evaluation on a panel of French firms from the manufacturing sector
G 9919 S. AUDRIC - P. GIVORD - C. PROST Evolution de l’emploi et des coûts par quali-fication entre 1982 et 1996
G 2000/01 R. MAHIEU Les déterminants des dépenses de santé : une approche macroéconomique
G 2000/02 C. ALLARD-PRIGENT - H. GUILMEAU - A. QUINET The real exchange rate as the relative price of nontrables in terms of tradables: theoretical investigation and empirical study on French data
G 2000/03 J.-Y. FOURNIER L’approximation du filtre passe-bande proposée par Christiano et Fitzgerald
G 2000/04 Bilan des activités de la DESE - 1999
G 2000/05 B. CREPON - F. ROSENWALD Investissement et contraintes de financement : le poids du cycle Une estimation sur données françaises
G 2000/06 A. FLIPO Les comportements matrimoniaux de fait
G 2000/07 R. MAHIEU - B. SÉDILLOT Microsimulations of the retirement decision: a supply side approach
G 2000/08 C. AUDENIS - C. PROST Déficit conjoncturel : une prise en compte des conjonctures passées
G 2000/09 R. MAHIEU - B. SÉDILLOT Equivalent patrimonial de la rente et souscription de retraite complémentaire
G 2000/10 R. DUHAUTOIS Ralentissement de l’investissement : petites ou grandes entreprises ? industrie ou tertiaire ?
G 2000/11 G. LAROQUE - B. SALANIÉ Temps partiel féminin et incitations financières à l’emploi
G2000/12 Ch. GIANELLA Local unemployment and wages
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G2000/13 B. CREPON - Th. HECKEL - Informatisation en France : une évaluation à partir de données individuelles - Computerization in France: an evaluation based on individual company data
G2001/01 F. LEQUILLER - La nouvelle économie et la mesure de la croissance du PIB - The new economy and the measure ment of GDP growth
G2001/02 S. AUDRIC La reprise de la croissance de l’emploi profite-t-elle aussi aux non-diplômés ?
G2001/03 I. BRAUN-LEMAIRE Evolution et répartition du surplus de productivité
G2001/04 A. BEAUDU - Th. HECKEL Le canal du crédit fonctionne-t-il en Europe ? Une étude de l’hétérogénéité des comportements d’investissement à partir de données de bilan agrégées
G2001/05 C. AUDENIS - P. BISCOURP - N. FOURCADE - O. LOISEL Testing the augmented Solow growth model : An empirical reassessment using panel data
G2001/06 R. MAHIEU - B. SÉDILLOT Départ à la retraite, irréversibilité et incertitude
G2001/07 Bilan des activités de la DESE - 2000
G2001/08 J. Ph. GAUDEMET Les dispositifs d’acquisition à titre facultatif d’annuités viagères de retraite
G2001/09 B. CRÉPON - Ch. GIANELLA Fiscalité, coût d’usage du capital et demande de facteurs : une analyse sur données individuelles
G2001/10 B. CRÉPON - R. DESPLATZ Evaluation des effets des dispositifs d’allégements de charges sociales sur les bas salaires
G2001/11 J.-Y. FOURNIER Comparaison des salaires des secteurs public et privé
G2001/12 J.-P. BERTHIER - C. JAULENT R. CONVENEVOLE - S. PISANI Une méthodologie de comparaison entre consommations intermédiaires de source fiscale et de comptabilité nationale
G2001/13 P. BISCOURP - Ch. GIANELLA Substitution and complementarity between capital, skilled and less skilled workers: an analysis at the firm level in the French manufacturing industry
G2001/14 I. ROBERT-BOBEE Modelling demographic behaviours in the French microsimulation model Destinie: An analysis of future change in completed fertility
G2001/15 J.-P. ZOYEM Diagnostic sur la pauvreté et calendrier de revenus : le cas du “Panel européen des ménages »
G2001/16 J.-Y. FOURNIER - P. GIVORD La réduction des taux d’activité aux âges extrêmes, une spécificité française ?
G2001/17 C. AUDENIS - P. BISCOURP - N. RIEDINGER Existe-t-il une asymétrie dans la transmission du prix du brut aux prix des carburants ?
G2002/01 F. MAGNIEN - J.-L. TAVERNIER - D. THESMAR Les statistiques internationales de PIB par habitant en standard de pouvoir d’achat : une analyse des résultats
G2002/02 Bilan des activités de la DESE - 2001
G2002/03 B. SÉDILLOT - E. WALRAET La cessation d’activité au sein des couples : y a-t-il interdépendance des choix ?
G2002/04 G. BRILHAULT - Rétropolation des séries de FBCF et calcul du
capital fixe en SEC-95 dans les comptes nationaux français
- Retropolation of the investment series (GFCF) and estimation of fixed capital stocks on the ESA-95 basis for the French balance sheets
G2002/05 P. BISCOURP - B. CRÉPON - T. HECKEL - N. RIEDINGER How do firms respond to cheaper computers? Microeconometric evidence for France based on a production function approach
G2002/06 C. AUDENIS - J. DEROYON - N. FOURCADE L’impact des nouvelles technologies de l’information et de la communication sur l’économie française - un bouclage macro-économique
G2002/07 J. BARDAJI - B. SÉDILLOT - E. WALRAET Évaluation de trois réformes du Régime Général d’assurance vieillesse à l’aide du modèle de microsimulation DESTINIE
G2002/08 J.-P. BERTHIER Réflexions sur les différentes notions de volume dans les comptes nationaux : comptes aux prix d’une année fixe ou aux prix de l’année précédente, séries chaînées
G2002/09 F. HILD Les soldes d’opinion résument-ils au mieux les réponses des entreprises aux enquêtes de conjoncture ?
G2002/10 I. ROBERT-BOBÉE Les comportements démographiques dans le modèle de microsimulation Destinie - Une comparaison des estimations issues des enquêtes Jeunes et Carrières 1997 et Histoire Familiale 1999
G2002/11 J.-P. ZOYEM La dynamique des bas revenus : une analyse des entrées-sorties de pauvreté
G2002/12 F. HILD Prévisions d’inflation pour la France
G2002/13 M. LECLAIR Réduction du temps de travail et tensions sur les facteurs de production
G2002/14 E. WALRAET - A. VINCENT - Analyse de la redistribution intragénérationnelle dans le système de retraite des salariés du privé - Une approche par microsimulation - Intragenerational distributional analysis in the french private sector pension scheme - A microsimulation approach
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G2002/15 P. CHONE - D. LE BLANC - I. ROBERT-BOBEE Offre de travail féminine et garde des jeunes enfants
G2002/16 F. MAUREL - S. GREGOIR Les indices de compétitivité des pays : inter-prétation et limites
G2003/01 N. RIEDINGER - E.HAUVY Le coût de dépollution atmosphérique pour les entreprises françaises : Une estimation à partir de données individuelles
G2003/02 P. BISCOURP et F. KRAMARZ Création d’emplois, destruction d’emplois et internationalisation des entreprises industrielles françaises : une analyse sur la période 1986-1992
G2003/03 Bilan des activités de la DESE - 2002
G2003/04 P.-O. BEFFY - J. DEROYON - N. FOURCADE - S. GREGOIR - N. LAÏB - B. MONFORT Évolutions démographiques et croissance : une projection macro-économique à l’horizon 2020
G2003/05 P. AUBERT La situation des salariés de plus de cinquante ans dans le secteur privé
G2003/06 P. AUBERT - B. CRÉPON Age, salaire et productivité La productivité des salariés décline-t-elle en fin de carrière ?
G2003/07 H. BARON - P.O. BEFFY - N. FOURCADE - R. MAHIEU Le ralentissement de la productivité du travail au cours des années 1990
G2003/08 P.-O. BEFFY - B. MONFORT Patrimoine des ménages, dynamique d’allocation et comportement de consommation
G2003/09 P. BISCOURP - N. FOURCADE Peut-on mettre en évidence l’existence de rigidités à la baisse des salaires à partir de données individulles ? Le cas de la France à la fin des années 90
G2003/10 M. LECLAIR - P. PETIT Présence syndicale dans les firmes : quel impact sur les inégalités salariales entre les hommes et les femmes ?
G2003/11 P.-O. BEFFY - X. BONNET - M. DARRACQ-PARIES - B. MONFORT MZE: a small macro-model for the euro area
G2004/01 P. AUBERT - M. LECLAIR La compétitivité exprimée dans les enquêtes trimestrielles sur la situation et les perspectives dans l’industrie
G2004/02 M. DUÉE - C. REBILLARD La dépendance des personnes âgées : une projection à long terme
G2004/03 S. RASPILLER - N. RIEDINGER Régulation environnementale et choix de localisation des groupes français
G2004/04 A. NABOULET - S. RASPILLER Les déterminants de la décision d’investir : une approche par les perceptions subjectives des firmes
G2004/05 N. RAGACHE La déclaration des enfants par les couples non mariés est-elle fiscalement optimale ?
G2004/06 M. DUÉE L’impact du chômage des parents sur le devenir scolaire des enfants
G2004/07 P. AUBERT - E. CAROLI - M. ROGER New Techlologies, Workplace Organisation and the Age Structure of the Workforce: Firm-Level Evidence
G2004/08 E. DUGUET - C. LELARGE Les brevets accroissent-ils les incitations privées à innover ? Un examen microéconométrique
G2004/09 S. RASPILLER - P. SILLARD Affiliating versus Subcontracting: the Case of Multinationals
G2004/10 J. BOISSINOT - C. L’ANGEVIN - B. MONFORT Public Debt Sustainability: Some Results on the French Case
G2004/11 S. ANANIAN - P. AUBERT Travailleurs âgés, nouvelles technologies et changements organisationnels : un réexamen à partir de l’enquête « REPONSE »
G2004/12 X. BONNET - H. PONCET Structures de revenus et propensions différentes à consommer - Vers une équation de consommation des ménages plus robuste en prévision pour la France
G2004/13 C. PICART Évaluer la rentabilité des sociétés non financières
G2004/14 J. BARDAJI - B. SÉDILLOT - E. WALRAET Les retraites du secteur public : projections à l’horizon 2040 à l’aide du modèle de microsimulation DESTINIE
G2005/01 S. BUFFETEAU - P. GODEFROY Conditions de départ en retraite selon l’âge de fin d’études : analyse prospective pour les générations 1945 à1974
G2005/02 C. AFSA - S. BUFFETEAU L’évolution de l’activité féminine en France : une approche par pseudo-panel
G2005/03 P. AUBERT - P. SILLARD Délocalisations et réductions d’effectifs dans l’industrie française
G2005/04 M. LECLAIR - S. ROUX Mesure et utilisation des emplois instables dans les entreprises
G2005/05 C. L’ANGEVIN - S. SERRAVALLE Performances à l’exportation de la France et de l’Allemagne - Une analyse par secteur et destination géographique
G2005/06 Bilan des activités de la Direction des Études et Synthèses Économiques - 2004
G2005/07 S. RASPILLER La concurrence fiscale : principaux enseigne-ments de l’analyse économique
G2005/08 C. L’ANGEVIN - N. LAÏB Éducation et croissance en France et dans un panel de 21 pays de l’OCDE
G2005/09 N. FERRARI Prévoir l’investissement des entreprises
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Un indicateur des révisions dans l’enquête de conjoncture sur les investissements dans l’industrie.
G2005/10 P.-O. BEFFY - C. L’ANGEVIN Chômage et boucle prix-salaires : apport d’un modèle « qualiifés/peu qualifiés »
G2005/11 B. HEITZ A two-states Markov-switching model of inflation in France and the USA: credible target VS inflation spiral
G2005/12 O. BIAU - H. ERKEL-ROUSSE - N. FERRARI Réponses individuelles aux enquêtes de conjoncture et prévision macroéconomiques : Exemple de la prévision de la production manufacturière
G2005/13 P. AUBERT - D. BLANCHET - D. BLAU The labour market after age 50: some elements of a Franco-American comparison
G2005/14 D. BLANCHET - T. DEBRAND - P. DOURGNON - P. POLLET L’enquête SHARE : présentation et premiers résultats de l’édition française
G2005/15 M. DUÉE La modélisation des comportements démogra-phiques dans le modèle de microsimulation DESTINIE
G2005/16 H. RAOUI - S. ROUX Étude de simulation sur la participation versée aux salariés par les entreprises
G2006/01 C. BONNET - S. BUFFETEAU - P. GODEFROY Disparités de retraite de droit direct entre hommes et femmes : quelles évolutions ?
G2006/02 C. PICART Les gazelles en France
G2006/03 P. AUBERT - B. CRÉPON -P. ZAMORA Le rendement apparent de la formation continue dans les entreprises : effets sur la productivité et les salaires
G2006/04 J.-F. OUVRARD - R. RATHELOT Demographic change and unemployment: what do macroeconometric models predict?
G2006/05 D. BLANCHET - J.-F. OUVRARD Indicateurs d’engagements implicites des systèmes de retraite : chiffrages, propriétés analytiques et réactions à des chocs démographiques types
G2006/06 G. BIAU - O. BIAU - L. ROUVIERE Nonparametric Forecasting of the Manufacturing Output Growth with Firm-level Survey Data
G2006/07 C. AFSA - P. GIVORD Le rôle des conditions de travail dans les absences pour maladie
G2006/08 P. SILLARD - C. L’ANGEVIN - S. SERRAVALLE Performances comparées à l’exportation de la France et de ses principaux partenaires Une analyse structurelle sur 12 ans
G2006/09 X. BOUTIN - S. QUANTIN Une méthodologie d’évaluation comptable du coût du capital des entreprises françaises : 1984-2002
G2006/10 C. AFSA L’estimation d’un coût implicite de la pénibilité du travail chez les travailleurs âgés
G2006/11 C. LELARGE Les entreprises (industrielles) françaises sont-elles à la frontière technologique ?
G2006/12 O. BIAU - N. FERRARI Théorie de l’opinion Faut-il pondérer les réponses individuelles ?
G2006/13 A. KOUBI - S. ROUX Une réinterprétation de la relation entre productivité et inégalités salariales dans les entreprises
G2006/14 R. RATHELOT - P. SILLARD The impact of local taxes on plants location decision
G2006/15 L. GONZALEZ - C. PICART Diversification, recentrage et poids des activités de support dans les groupes (1993-2000)
G2007/01 D. SRAER Allègements de cotisations patronales et dynamique salariale
G2007/02 V. ALBOUY - L. LEQUIEN Les rendements non monétaires de l’éducation : le cas de la santé
G2007/03 D. BLANCHET - T. DEBRAND Aspiration à la retraite, santé et satisfaction au travail : une comparaison européenne
G2007/04 M. BARLET - L. CRUSSON Quel impact des variations du prix du pétrole sur la croissance française ?
G2007/05 C. PICART Flux d’emploi et de main-d’œuvre en France : un réexamen
G2007/06 V. ALBOUY - C. TAVAN Massification et démocratisation de l’enseignement supérieur en France
G2007/07 T. LE BARBANCHON The Changing response to oil price shocks in France : a DSGE type approach
G2007/08 T. CHANEY - D. SRAER - D. THESMAR Collateral Value and Corporate Investment Evidence from the French Real Estate Market
G2007/09 J. BOISSINOT Consumption over the Life Cycle: Facts for France
G2007/10 C. AFSA Interpréter les variables de satisfaction : l’exemple de la durée du travail
G2007/11 R. RATHELOT - P. SILLARD Zones Franches Urbaines : quels effets sur l’emploi salarié et les créations d’établissements ?
G2007/12 V. ALBOUY - B. CRÉPON Aléa moral en santé : une évaluation dans le cadre du modèle causal de Rubin
G2008/01 C. PICART Les PME françaises : rentables mais peu dynamiques
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G2008/02 P. BISCOURP - X. BOUTIN - T. VERGÉ The Effects of Retail Regulations on Prices Evidence form the Loi Galland
G2008/03 Y. BARBESOL - A. BRIANT Économies d’agglomération et productivité des entreprises : estimation sur données individuelles françaises
G2008/04 D. BLANCHET - F. LE GALLO Les projections démographiques : principaux mécanismes et retour sur l’expérience française
G2008/05 D. BLANCHET - F. TOUTLEMONDE Évolutions démographiques et déformation du cycle de vie active : quelles relations ?
G2008/06 M. BARLET - D. BLANCHET - L. CRUSSON Internationalisation et flux d’emplois : que dit une approche comptable ?
G2008/07 C. LELARGE - D. SRAER - D. THESMAR Entrepreurship and Credit Constraints - Evidence from a French Loan Guarantee Program
G2008/08 X. BOUTIN - L. JANIN Are Prices Really Affected by Mergers?
G2008/09 M. BARLET - A. BRIANT - L. CRUSSON Concentration géographique dans l’industrie manufacturière et dans les services en France : une approche par un indicateur en continu
G2008/10 M. BEFFY - É. COUDIN - R. RATHELOT Who is confronted to insecure labor market histories? Some evidence based on the French labor market transition