ISA Research Committee on Social Stratification and Mobility (RC28)

Neuchâtel Conference, Switzerland, 7-9 May 2004

The Dynamics of Inequality of Educational Opportunity in France:

Change in the Association Between Social Background and Education

in Thirteen Five-Year Birth Cohorts (1908-1972)

Louis-André Vallet, National Center for Scientific Research (CNRS), France

Quantitative Sociology Laboratory, Center for Research in Economics and Statistics (CREST), Timbre J350, 3 avenue Pierre Larousse, 92245 Malakoff Cedex, France

louis-andre.vallet@wanadoo.fr or vallet@ensae.fr

- Last revision 26 April 2004 -

1

The Dynamics of Inequality of Educational Opportunity in France:

Change in the Association Between Social Background and Education

in Thirteen Five-Year Birth Cohorts (1908-1972)1

1 I warmly thank Claude Thélot with whom I worked in a preliminary collaborative project on

that topic. Earlier versions of this article were presented at the European Science Foundation

conference “Educational Differentiation in European Societies: Causes and Consequences”

(Giens, France, September 16-21, 2000), at the meeting of the Social Stratification Research

Committee (RC 28) in the XV ISA World Congress of Sociology (Brisbane, Australia, July 7-

13, 2002) and at the methodological conference of the research network on Changing

Economy, Unequal Life-Chances and Quality of Life (Nuffield College, Oxford, September

25-27, 2003). I thank Hanna Ayalon, Richard Breen, Robert Erikson, Adam Gamoran, Harry

Ganzeboom, Anthony Heath, Walter Müller and Yossi Shavit for stimulating and helpful

comments. Special thanks are due to John Goldthorpe and Mike Hout who independently

suggested investigating the link between the two-way contingency table approach and the

educational transition approach on an empirical basis.

2

Abstract: Despite the educational expansion in industrialized societies, the association

between social origin of adult men and women and the highest degree they got is generally

considered highly stable over time. However, that “persistent inequality” conclusion has been

challenged more recently for a few countries by using powerful log-multiplicative statistical

techniques. This study addresses the issue of social class inequalities in educational

attainment for France from early twentieth century. Based on data from seven nationally

representative surveys (N circa 240,000 cases), the analysis uses the log-multiplicative layer

effect model in full interaction and the more recent Goodman-Hout regression-type model

(Sociological Methodology 1998). In French society the association between social origin and

education has undergone a decline among cohorts born from early twentieth century. This

change has been more marked for women than for men and has mainly occurred among

cohorts born from the mid-thirties to the early fifties. As regards temporal dynamics, this

study therefore confirms and extends previous research published in France by Prost (1986)

and in the US by Smith and Garnier (1986). The results also show that, until cohorts born in

the mid-thirties, the level of association between social origin and education was distinctly

stronger for women than for men. Studying the sensitivity of estimated trends to use of

various indicators of social background suggests that cultural inequalities in education are

more resistant to change than socioeconomic inequalities. Finally, investigating the link

between the two-way contingency table approach and the educational transition approach,

this article demonstrates on an empirical basis that the general downward trend in origin –

education association is compatible with remarkably stable, or even increasing social origin

effects in transitions associated with advanced stages of the educational system, but is

generated by the consequences of declining origin effects in the most basic transitions.

3

The dynamics of socioeconomic inequality of educational opportunities in industrial or post-

industrial societies is a research question with a long-standing interest in sociology which has

been studied using various quantitative approaches. It is indeed worth emphasizing that its

statistical methodologies have been strongly reshaped over the last three decades. Till the

end of the 1970s the linear regression model of educational attainment was the unique

approach. Using a metric dependent variable to measure the final amount of schooling, the

first period answered the following research question: what has been the change over time in

the effect of social origin variables on the mean number of school years completed? The

empirical evidence was somewhat mixed. Reporting regressions of highest grade of school

completed on father’s occupation, parental income, father’s and mother’s schooling and

three other control variables for American white males born between 1907 and 1951, Mare

(1981) described little change in the educational attainment process: only the effect of

father’s occupational socioeconomic index had declined slightly from early cohorts to more

recent ones. More generally, summarizing the results of a comparative project on thirteen

countries, Shavit and Blossfeld (1993, p. 16) concluded that the effect of father’s education

declined over time in five countries and remained unchanged in the others (except for a

country where the effect first declined then increased), and that the effect of father’s

occupation remained unchanged in nine countries, declined in three and increased in one.

However, as emphasized by Treiman and Ganzeboom (2000), in six of the eight nations for

which linear regressions were reported cohort by cohort, a downward trend was apparent in

the proportion of variance explained by background variables, thereby suggesting a historical

decline in the dependence of educational attainment on social origins.

The second period of educational stratification research began with the proposal of the

sequential logistic regression model of educational transitions (Mare 1980). Decomposing

the intrinsically discrete and sequential nature of an educational career in a series of

successive branching points – an idea also outlined by Boudon (1974) – this model assesses

the net effect of social background variables on the odds of “surviving” each specific

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transition. With this model it has widely been observed that social origin effects decline

steadily from the earliest school transitions to the latest (Müller and Karle 1993; Shavit and

Blossfeld 1993; Rijken 1999). This result has often been attributed to a process of differential

selection: from the earliest to the latest school transitions, differential dropout rates

systematically reduce heterogeneity between children from different social origins on

unmeasured determinants of school continuation such as ability or motivation (Mare 1981, p.

82), and because of the correlation between these variables and social origins greater

homogeneity on unmeasured factors at higher levels of schooling reduces the effects of

observed social background variables.2 According to a related argument, as educational

expansion increases the proportion of the total population at risk at a given transition, its

heterogeneity on unmeasured determinants of school continuation grows and, as a

consequence, the effects of social background variables on the odds of surviving that

transition are likely to increase. However, as regards change over time, the empirical

evidence was rather mixed. In the aforementioned comparative project, only two out of

thirteen countries (the Netherlands and Sweden) experienced a decline in social origin

effects for transitions within secondary education. In the remaining countries, contrasts

between men and women from different social backgrounds were fairly stable over birth

cohorts for each transition, thus leading the editors to a general “persistent inequality”

2 However, more recent research which analyzed data on American cohorts who

experienced an environment of contracting public support for higher education has cast

some doubt on the selective attrition explanation (Lucas 1996). After a more advanced

discussion of the incidence of unobserved heterogeneity on modeling school continuation

decisions (Mare 1993), research by economists has also strongly criticized the dynamic

selection bias faced by the sequential logit model of educational transitions and, more

generally, has questioned the usefulness of this model (Cameron and Heckman 1998); see,

however, the vigorous response by Lucas (2001, pp. 1653-62).

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conclusion (Shavit and Blossfeld 1993).3 However, another cross-national and over-time

comparison based on a large-scale data set found both declining social inequalities over

cohorts in school continuation probabilities and an offsetting effect caused by an increase in

the percentage of students at risk which enhances the effects of father’s occupation (Rijken

1999). A suggestive conclusion of the latter study therefore is that the “persistent inequality”

conclusion emphasized by Shavit and Blossfeld would in fact be produced by the

combination of two contradictory trends.

As regards temporal change in inequality of educational opportunity, the linear regression

model of highest educational level attained and the sequential logistic regression model of

educational transitions may tell us different, albeit reconcilable stories. While the latter is only

sensitive to the relative allocation of schooling between social groups, the former is also

affected by the marginal distribution of schooling and change in it, notably increased average

educational level as a consequence of educational expansion (Mare 1981). For instance, for

American white males born between 1907 and 1951, quasi-stable linear effects of parental

socioeconomic characteristics on highest grade attained were produced by the combination

of inter-cohort increases in school continuation rates (which by themselves imply declining

background effects on educational attainment) and increasing social origin effects over time

on the odds of surviving some educational transitions. Conversely, in the Philippines, Smith

and Cheung (1986) demonstrated both declining social background effects in the linear

model of educational attainment and stable background effects on each of the educational

transitions. It might thus been said that, while the linear model provides a general picture of

temporal change in the dependence of educational attainment on social origins (De Graaf

and Ganzeboom 1993), the sequential logistic regression model yields a more structural or

3 More recently, a reanalysis of Italian data nonetheless revealed declining effects of father’s

education on the odds of completing the lower levels of the educational hierarchy (Shavit and

Westerbeek 1998).

6

“pure” measure of inequality of educational opportunity as it is unaffected by historical

change in the marginal distribution of schooling (Mare 1993). This provides an understanding

of its impressive centrality in comparative educational stratification research over the last two

decades which also comes from the fact that the discrete model corresponds to the way

persons accumulate formal schooling, namely, in a sequence of irreversible steps.

In our view and despite its intrinsic interest, the classical model of educational transitions

nonetheless encounters two limitations. Firstly, it assumes that individuals progress through

the educational system in a unilinear sequential mode whereas many school systems –

notably those of European societies – contain parallel branches of study that are most

fruitfully seen as qualitatively different alternative pathways with different probabilities of

school continuation attached to them. Within the Mare model it is therefore difficult to take

account of possibly existing second-order differences such as the decision between

vocational studies and academic studies within secondary education and this feature recently

led Breen and Jonsson (2000) to propose a multinomial transition model. Secondly, as it

closely parallels the continuation decision process along the educational career, the

sequential model of educational transitions provides us with local “pure” measures of social

origin effects, i.e. measures which are specific for each transition examined. But the

sequential model leaves the following question entirely unanswered: if, in a given country,

social origin effects decline over birth cohorts for some transitions, but remain stable or even

increase for some others, what is the final outcome as regards temporal dynamics in the

intrinsic association between highest educational level attained and social origins in that

country? Such a question may be answered using a statistical technique which is less

frequent in educational stratification studies, namely log-linear and log-multiplicative

modeling of a detailed three-way contingency table cross-classifying birth cohort, social

origins and highest educational level attained. Such an approach does not afford a

multivariate perspective able to separate the effects of multiple social background

determinants of educational attainment. As regards its micro-sociological foundations, a

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drawback of what we can call the contingency table approach to educational stratification

certainly is that it implicitly (and rather implausibly) assumes that the decision of how much

education to acquire is taken at the outset of the individual’s educational career (Breen and

Jonsson 2000). However, as a tool to discover what has occurred in society, this

methodological approach has certainly been revitalized over recent years, especially

because progress in log-multiplicative modeling – the “Unidiff” or log-multiplicative layer

effect model (Erikson and Goldthorpe 1992; Xie 1992) – now offers considerable statistical

power to discern even slow historical trends which would have gone undetected otherwise.

For instance, using this technique led Jonsson, Mills and Müller (1996) to demonstrate a

trend toward equalization in West Germany and to challenge the “persistent inequality”

conclusion previously reached for this country in the Shavit and Blossfeld’s comparative

project. Finally, a more flexible and more general log-multiplicative model has been recently

proposed to analyse how the association between two variables depends on a third variable

(Goodman and Hout 1998, 2001). It is likely that this model can be useful in comparative

educational stratification research and will prove a sensible instrument to trace change in

strength and pattern of association between social origins and highest qualification attained

over birth cohorts.

Following the same statistical approach, the aim of this paper is therefore to investigate

temporal dynamics in the association between social background and education in France

over thirteen five-year birth cohorts (1908-1972). As France was not represented among the

thirteen countries studied in Shavit and Blossfeld’s comparative project, we begin by briefly

reviewing existing literature on temporal trends in inequality of educational opportunity in

French society, then we detail the statistical models we will use paying special attention to

the most recent one, namely the Goodman and Hout’s regression-type model which will

prove able to fit French data remarkably well. Finally, after a presentation of the surveys and

variables we use, the empirical section will be devoted to an examination of the following

questions:

8

(i) to what extent does it exist a trend over time toward decreasing association between

social origins and educational destination in France and does its magnitude differ between

men and women as it has been observed in Sweden (Jonsson, Mills and Müller 1996)?

(ii) as regards temporal dynamics in inequality of educational opportunity in France, how can

we conciliate results provided by the contingency table approach with those yielded by the

sequential model of educational transitions? more specifically, is change in inequality of

educational opportunity in France compatible with the “maximally maintained inequality”

hypothesis (Hout, Raftery and Bell 1993) according to which the effects of social origin on

educational destination only decline at those levels of the educational system for which the

attendance rates of the privileged classes are saturated?

(iii) to what extent does the magnitude of the declining trend in inequality of educational

opportunity in France depend on the variable used for social background as it has been

observed in the Netherlands where the effect of father’s education has proved more resistant

to change than the effect of father’s occupation (De Graaf and Ganzeboom 1993)?

(iv) how much difference does the reduction in inequality of educational opportunity in France

make, i.e. how many children from the non privileged classes in the youngest (1968-1972)

birth cohort possess high educational qualifications which they would not have held if the

association between social origin and educational destination had remained the same as it

was in the oldest (1908-1912) birth cohort?

Previous research on temporal trends in inequality of educational opportunity in

France

Previous research which investigated the dynamics of inequality of educational opportunity in

France followed the three aforementioned statistical approaches and analyzed one or

several of the Formation-Qualification Professionnelle (FQP) surveys, a series of nationally

representative surveys conducted by the French National Institute of Statistics and Economic

Surveys (INSEE).

9

Using data from the 1977, 1985 and 1993 FQP surveys in order to distinguish between six

birth cohorts, Duru-Bellat and Kieffer (2000) estimated linear regression models of

educational attainment on five explanatory variables: father’s and mother’s socioeconomic

group, father’s and mother’s highest diploma and gender. A shortcoming of the analysis is

that the dependent variable is not measured as the actual number of years of education, but

on a somewhat arbitrary interval-level scale ranging from 1 for “without any diploma” to 7 for

“tertiary education diploma”. The results nonetheless clearly demonstrate a steady fall in the

explanatory power of the background variables: R2 decreases from 32.3% for men and

women born before 1939 to 20.3% for the most recent (1964-1973) birth cohort. However,

the decrease in the dependence of highest educational level attained on background

variables only applies to the effect of father’s and mother’s education while the effect of their

socioeconomic group remains much more stable over time.

Brauns (1998) applied the sequential logistic regression model of educational transitions to

French data from the 1985 FQP survey and highlighted a complex picture of educational

inequalities over the various levels of the educational system. As regards admission to

secondary school, she found significantly declining class effects mainly from the 1945-1949

birth cohort for children of farmers as well as skilled and unskilled manual workers. The

enlargement of access to secondary education which strongly accelerated from the early

1950s however resulted in an opposite change for the next transition: in the 1955-1959,

1960-1964 and 1965-1968 birth cohorts the odds of completing lower secondary education

worsened for children of skilled and unskilled manual workers who entered secondary school

as compared to corresponding odds for upper service class children.4 From completion of

lower secondary education to completion of upper secondary education, decreasing

inequality of educational opportunity again characterized skilled manual workers from the

1925-1934 birth cohort as well as non manual employees and the lower service class from

4 The same result was also found in independent analyses by Duru-Bellat and Kieffer (2000).

10

the 1935-1944 birth cohort. No trend at all was apparent as regards the last transitions

(completion of lower, respectively intermediate or upper, tertiary education). Estimating also

unconditional logit models for completion of each educational level, Brauns (1998, 1999)

finally demonstrated that declining class inequalities for admission to secondary education

also resulted in similar effects for completion of lower and upper secondary education as well

as lower and intermediate tertiary education.

It is worth emphasizing that these results are therefore somewhat at odds with a widely cited

paper by Garnier and Raffalovich (1984) whose conclusion stressed little change in the

pattern of association between social origins and educational certification in France. Using

the 1970 FQP survey to distinguish between six ten-year birth cohorts, five categories of

father’s occupation and seven educational categories (from “no degree at all” to “a university

degree”), the authors constructed a series of hierarchical log-linear models estimating the

odds of having obtained each degree versus all others as a function of father’s occupation

and cohort. After taking account of marginal shifts in the distributions of occupations and

diplomas over cohorts, they found that the constant association model was able to reproduce

the observed data rather faithfully and concluded that “the educational expansion that has

taken place over the twentieth century has not enormously altered differential access to

education based on social origins, except for sons of farmers and for women” (Garnier and

Raffalovich 1984, p. 9).

However, Smith and Garnier (1986) questioned the appropriateness of this analysis in a

subsequent but less widely cited paper.5 Recognizing that moving directly from the constant

association model to the saturated model has strong drawbacks, they argued that it is much

5 For instance, in their comparative project, Shavit and Blossfeld (1993, p. 4) commented on

Garnier and Raffalovich (1984) but did not introduce any reference to Smith and Garnier

(1986).

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more preferable to assess trends with intermediate models that do not require constant

association across cohorts, but that stop short of using all available degrees of freedom. The

authors thus applied Goodman’s log-multiplicative association model to the 1970 French

data for men and women born between 1920 and 1949 and demonstrated “a notable decline

among cohorts born in the 1940s in the strong association between father’s occupation and

highest degree obtained, and few sex differences in the trend in this association” (Smith and

Garnier 1986, p. 339).

Goux and Maurin (1995) also used hierarchical log-linear modeling on French data for young

people aged 25 to 34 in the 1970, 1977, 1985 and 1993 surveys as well as for men and

women aged 25 to 64 in the 1993 survey. However, following Garnier and Raffalovich’s

strategy, they also moved directly from the constant association model to the saturated one.

After an examination of three-way interaction parameters (social origin x educational

destination x survey (or birth cohort)), their conclusion which, for reasons underlined above,

may suffer from lack of statistical power was that no firm trend has existed toward increasing

or decreasing inequality of educational opportunity in French society.

Before concluding this review, it must be stressed that Smith and Garnier’s findings fit well

with French studies in history of education. Investigating all secondary schools in the Orléans

area, Prost (1986, 1990, 1992) systematically examined the transformations in the social

recruitment of pupils between 1945 and 1980 at various levels of the educational system.

According to his analysis, a process of democratization has been in train since the mid-

1940s. However, Prost also concluded that educational reforms introduced between the end

of the 1950s and the mid-1960s to provide children from all social backgrounds with

increased education and to promote equality of educational opportunity have paradoxically

introduced additional rigidities which have impeded the ongoing process of democratization.

In the following sections, we therefore aim to replicate and extend Smith and Garnier’s

12

analysis by applying recent progress in log-multiplicative contingency table analysis to

French data.

Statistical modeling

Let c be one of the examined birth cohorts (out of NC), o and o’ two different social origins

(out of NO), e and e’ two different educational destinations (out of NE) and odc the odds ratio

which measures, in birth cohort c, the intrinsic strength of the statistical association between

these social origins and these educational destinations:

)()()()()(elyalternativor '''''''

'ecocoeceooecc

ceoeco

coeoecc mLogmLogmLogmLogodLog

mmmm

od −−+==

We can analyze the contingency table cross-classifying social origin (O), educational

destination (E) and cohort (C) by means of four log-linear or log-multiplicative nested models.

Null association model (1)

λλλλλλ EC

ec

OC

oc

C

c

E

e

O

ooecmLog +++++=)(

Estimated with NC(NO-1)(NE-1) degrees of freedom, Model 1 implies that 0)( =codLog or

1=cod . Assuming that social origin and educational destination are independent in each

birth cohort, it expresses the hypothesis of complete equality of educational opportunity and

provides us with a reference for assessing the extent to which more realistic models fit the

data more closely.

Constant association model (2)

λλλλλλλ OE

oe

EC

ec

OC

oc

C

c

E

e

O

ooecmLog ++++++=)(

Estimated with (NO-1)(NE-1)(NC-1) degrees of freedom, Model 2 implies:

λλλλOE

eo

OE

oe

OE

eo

OE

oecodLog''''

)( −−+=

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Assuming that all the odds ratios which measure the association between social origin and

educational destination are constant over birth cohorts, Model 2 expresses the hypothesis of

constant inequality of educational opportunity.

‘Multiplicative uniform cohort effect’ association model (3)

ψβλλλλλλ oec

EC

ec

OC

oc

C

c

E

e

O

ooecmLog ++++++=)(

Proposed by Xie (1992) and Erikson and Goldthorpe (1992), Model 3 is estimated with

(NO.NE-NO-NE)(NC-1) degrees of freedom and implies:

)()(''''

ψψψψβeooeeooeccodLog −−+=

Decomposing the origin – education association and its variation over birth cohorts as the

product of a common pattern (the ψoe

parameters) and a cohort-specific parameter (βc),

Model 3 is able to detect differences over cohorts in strength of association, i.e. in the

general level of inequality of educational opportunity. More precisely, assuming that β1 is

set at 1, estimating βc as less than 1 (respectively more than 1) for a subsequent cohort will

correspond to all estimated logged odds ratios moving towards 0 (respectively away from 0),

i.e. will correspond to the association becoming weaker (respectively stronger) than in the

first cohort. As it assumes that all odds ratios are moving in the same direction from one

cohort to another and expresses this variation with only one parameter, Model 3 is very

powerful to detect a dominant trend in the data but may be also rather crude to accurately

describe the change that occurs.

‘Regression-type cohort effect’ association model (4)

ϕγλλλλλλλ oec

OE

oe

EC

ec

OC

oc

C

c

E

e

O

ooecmLog +++++++=)(

Proposed by Goodman and Hout (1998, 2001) and estimated with (NO.NE-NO-NE)(NC-2)

degrees of freedom, Model 4 implies:

14

)()()('''''''' ϕϕϕϕγλλλλ eooeeooec

OE

eo

OE

oe

OE

eo

OE

oecodLog −−++−−+=

While the λOE

oe parameters establish the baseline (stable over birth cohorts) pattern of

association between social origin and educational destination, the ϕoe

parameters represent

the part of the association which varies over birth cohorts and the magnitude of the γc

parameter determines the strength of the adjustment of the association for cohort c. As a

consequence, Model 4 is able to detect differences over cohorts in both pattern and strength

of association. More precisely, examination of the ϕoe

parameters will allow us to highlight

the combined social origins and educational destinations for which change over cohorts has

been the most pronounced while examination of the γc parameters will demonstrate which

birth cohorts have been most affected by change in the association between social origin and

education. Without any loss of generality we will use the following identifying constraints:

0and1,0NC1

====== ∑∑∑∑ γγϕϕλλe

oeo

oee

OE

oeo

OE

oe

so that the λOE

oe parameters will represent the pattern of association in the last (youngest)

birth cohort and the ϕoe

parameters will represent the pattern of deviation characteristic of

the first (oldest) birth cohort as compared to the last one.6

Data and variables

To analyze change in the association between social origin and educational destination in

20th-century France over a large number of cohorts and with sufficient statistical power we

use a series of seven nationally representative surveys conducted by INSEE, namely the

1964, 1970, 1977, 1985 and 1993 Formation-Qualification Professionnelle (FQP) surveys

6 For estimation purposes we use the LEM software (version 1.0) developed by Jeroen K.

Vermunt (University of Tilburg, The Netherlands).

15

and the 1993 and 1997 Emploi surveys. Taking characteristics of the surveys into account

and restricting the analysis to French-born men and women aged at least 25 with information

available on their father’s occupation at the time they ceased attending school or university

on a regular basis,7 we can assess the association between social origin and educational

destination in nearly fifty two-way contingency tables which correspond to thirteen five-year

birth cohorts from 1908-1912 to 1968-1972 (Table 1). The total sample size is 240,367 cases

(or 244,591 cases if the first two cohorts in the 1970 survey are included).

Preliminary analyses which investigated differences in one-way margins and two-way

association between contingency tables corresponding to the same cohort did not detect

important and systematic departures from homogeneity across surveys. Most of the analysis

will therefore combine the surveys to consolidate all tables for the same cohort and will finally

use thirteen cohort-specific tables cross-classifying social origin and educational destination.

In that case, the sample size in the 1908-1912 birth cohort is 1,273 cases (or 3,577

depending on treatment of the 1970 survey). It amounts to 11,063 cases in the 1968-1972

birth cohort and 36,118 in the most numerous one, namely the 1948-1952 birth cohort.

Robustness of results will however be assessed by supplementary analyses investigating

either each survey separately or the whole set of basic contingency tables.

We define social origin as an eight-category variable on the basis of father’s occupation: (1)

farmers and smallholders; (2) artisans and shopkeepers; (3) higher-grade professionals and

managers; (4) teachers (in primary, secondary or tertiary education) and assimilated

7 If the father was unknown or deceased at that time, his occupation was replaced by

mother’s or guardian’s occupation in the Emploi surveys. In the 1964 FQP survey father’s

occupation was only asked to men and women born after 1917. In the 1970 FQP survey

educational qualifications were differently and less accurately collected among people born

before 1918 than among the youngest.

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occupations; (5) lower-grade professionals and technicians; (6) routine non manual workers;

(7) foremen and skilled manual workers; (8) agricultural and unskilled manual workers.

Figure 1 shows how the relative importance of the various social origins has evolved in

France from the early decades of the twentieth century. Children of farmers and smallholders

experienced the most dramatic change, falling from nearly one third in the 1908-1912 birth

cohort to less than 6% in the 1968-1972 cohort. On the other hand the decrease was rather

limited for the offspring of the self-employed petty bourgeoisie (artisans and shopkeepers).

While the proportion of men and women originating from the agricultural and unskilled

working class remained more or less stable until the 1948-1952 birth cohort, it has declined

in the subsequent ones. Conversely, an almost continuous increase characterized the skilled

fraction of the working class so that one out of four men and women in the last cohort is the

child of a foreman or skilled manual worker. Finally, the growth of occupations in the tertiary

sector resulted in steadily expanding non manual classes: higher-grade professionals and

managers (from 7% to 11%), teachers and assimilated occupations (from 1% to 4%), lower-

grade professionals and technicians (from 2% to 10%) and routine non manual workers (from

9% to 16%).

We define educational destination as a seven-category variable on the basis of highest

degree obtained8: (1) no diploma (or no information); (2) primary education certificate

(Certificat d’Études Primaires); (3) lower secondary education diploma (without vocational

qualification) (Brevet élémentaire, BEPC); (4) lower vocational education diploma (Certificat

d’Aptitude Professionnelle, Examen de Fin d’Apprentissage Artisanal); (5) upper secondary

or technical education diploma (Baccalauréat or assimilated diploma); (6) lower tertiary

8 In the 1970, 1977, 1985 and 1993 FQP surveys we only consider degrees obtained in initial

schooling including apprenticeship, i.e. without taking post-school training or in-service

training into account. Unfortunately, a similar distinction between initial and further schooling

cannot be implemented in the 1964 FQP survey, nor in the 1993 and 1997 Emploi surveys.

17

education degree (one or two years after Baccalauréat); (7) upper tertiary education degree

(at least three years after Baccalauréat). Over more than sixty years educational expansion

has been dramatic in French society (Figure 2). A majority of men and women in the 1908-

1912 birth cohort did not possess any diploma and a further third held no more than a

primary education certificate, but corresponding percentages in the 1968-1972 birth cohort

were respectively 15% and 1%. While elementary qualification virtually disappeared over the

period, it is noticeable that the decrease in the proportion of men and women without any

qualification leveled off from the 1943-1947 cohort. The expansion of lower vocational

education was tremendous after the 1923-1927 birth cohort: the incumbents of such a

diploma reached a peak at 34% of the 1963-1967 generation, then declined rather sharply in

the very last birth cohort. Conversely, lower secondary education diplomas never exceeded

10% of a generation. But the most clear consequence of educational expansion has

expressed itself in a continuous rise in the relative size of advanced qualifications, either

incumbents of Baccalauréat-level diplomas (from 3% to 18%) or incumbents of post-

Baccalauréat diplomas (from 1% to 16% for degrees involving one or two further years of

education and from 2% to 16% for University or Grandes Écoles degrees).

In the first 1908-1912, the median 1938-1942 and the last 1968-1972 birth cohort,

educational destination strongly depends on social origin, and in essentially the same way

(Table 2). For instance, in each generation, men and women with origins in the “teachers

and assimilated occupations” category were the most advantaged, as indicated by the

percentage of those who reached a lower or upper tertiary degree. Using the same criterion,

children of higher-grade professionals and managers, then children of lower-grade

professionals and technicians were the second and third groups in each generation again.

Conversely, children of farmers and smallholders and children of agricultural and unskilled

manual workers were equally disadvantaged in the 1908-1912 birth cohort: the percentage

distributions were very close and in each case about two thirds did not get any diploma. In

the 1938-1942 birth cohort the offspring of the two social groups were again rather close and

18

still appeared to be the most disadvantaged considering their educational qualifications. But

children of farmers and smallholders strongly improved their relative position between the

1938-1942 and 1968-1972 cohorts. At the end of the period their educational destinations

were considerably more favorable than those of children of agricultural and unskilled manual

workers. They were also clearly better than those of children of foremen and skilled manual

workers and slightly better than those of routine non manual workers. The examination of

simple row percentages therefore suggests that despite strong inertia in the association

between social origin and educational destination in France some change has occurred from

the early decades of the twentieth century in which children of farmers and smallholders

played a significant part.

An empirical analysis of the dynamics of inequality of educational opportunity in

France

In order to analyze change in statistical association net of educational expansion and

marginal shifts in the distribution of social origins we apply the aforementioned log-linear and

log-multiplicative models to the three-way contingency table cross-classifying social origin (8

categories), educational destination (7 categories) and birth cohort (13 categories) (Table 3).

Beginning our analysis with the whole sample of all men and women (first panel), the

constant association model which imposes cohort invariance on all the odds ratios in the

origin – education table appears to have strong potential for describing French society over

thirteen five-year birth cohorts. Although it is rejected by a conventional statistical test as a

consequence of the extremely large sample size, this model has to be preferred to the

saturated model on the basis of the Bayesian Information Criterion (BIC) statistic, it

misclassifies only 4.1% of the total sample involved and eliminates 91.8% of the distance

which separates the data from the baseline model – that of statistical independence in each

cohort.

19

However, the ‘multiplicative uniform cohort effect’ association model which estimates twelve

supplementary parameters improves on the constant association model very significantly9

and is, according to the BIC statistic, also preferable to the latter. It may thus be concluded

that significant variations occurred over birth cohorts in the general strength of the origin –

education association. Finally, the ‘multiplicative uniform cohort effect’ association model is

superseded by the ‘regression-type cohort effect’ association model which affords the best fit

as indicated by the likelihood-ratio chi-square and BIC statistics. The latter model indeed fits

the data remarkably well as it misclassifies only 1.7% of the total sample involved and

eliminates 98.4% of the distance between the data and the null association hypothesis. It

must therefore be concluded that significant variations occurred over birth cohorts not only in

strength, but also in pattern of association between social origin and educational destination.

The third and fourth panels of Table 3 separately replicate the same analysis with similar

results on the male and female samples. Finally, as the previous section strongly suggests

that children of farmers and smallholders played a prominent part in changing origin –

education association, the second panel again applies the same models to the whole

sample while ignoring this social group in the analysis (using structural zeros). The constant

association model fits better than in the general analysis, but the ‘multiplicative uniform

cohort effect’ association model significantly improves on it again and the ‘regression-type

cohort effect’ association model still proves to be the best-fitting model. The second panel

therefore demonstrates that the dynamics of origin – education association did not only result

from change in the relative position of the offspring of farmers10. We go on with the analysis

9 The difference in the likelihood-ratio chi-square statistic is 665.7 for 12 degrees of freedom.

10 On the basis of the G2 statistics displayed in the first and second panels, it is possible to

compute the part of total lack of fit of any model which is attributable to children of farmers

and smallholders: 62.2% for the constant association and ‘multiplicative uniform cohort effect’

association models, but only 12.8% for the null association model and 27.1% for the

‘regression-type cohort effect’ association model.

20

by considering parameters which depict the extent of change under the simplest log-

multiplicative model, then we will examine parameters from the more complex ‘regression-

type cohort effect’ association model.

Change in the general strength of association between social origin and educational

destination

Figure 3 highlights the downward trend in inequality of educational opportunity in France,

estimated assuming a stable pattern of association. Fixed at 1 in the 1908-1912 cohort, the

log-multiplicative parameter is estimated at 1.123 in the next one, then decreases until 0.648

in the 1968-1972 cohort, thereby demonstrating a fall of more than 35% in the logged odds

ratios. While it is nearly monotonic, the downward trend cannot be accurately summarized

with a linear trend as a substantial part of the change occurred between the 1933-1937 and

1948-1952 birth cohorts, i.e. for generations who could enter secondary school before the

major educational reforms (Prost 1990). Change largely leveled off in the three subsequent

birth cohorts characterized by post-reform secondary school system, but took off again in the

very last cohort. After omitting children of farmers and smallholders in the analysis, a mainly

similar dynamics reappears though on a reduced scale as the parameter declines from 1.061

in the 1913-1917 cohort to 0.716 in the last one. This simultaneously demonstrates that the

improvement of educational opportunities among sons and daughters of farmers played a

significant part in accentuating the equalization trend but was not the only factor in creating

it.11

When it is separately applied to the seven surveys the ‘multiplicative uniform cohort effect’

association model significantly improves on the constant association model in each case and

Figure 4 displays the estimated log-multiplicative parameters. The 1970, 1977 and 1985

11 The same result has already been documented for Germany and Sweden (Jonsson, Mills

and Müller 1996, pp. 194-5).

21

FQP surveys are remarkably consistent in demonstrating a sustained trend toward

equalization between the 1933-1937 and 1938-1942 birth cohorts. This both confirms Smith

and Garnier’s previous findings and Prost’s historical study according to which a reform

promulgated in 1941 by the conservative Minister of Education Jérôme Carcopino to

integrate the Écoles Primaires Supérieures in the secondary school system had positive

effects and resulted in declining inequality of educational opportunity. Despite some

irregularities, notably for the last birth cohorts in the 1993 FQP survey, the survey-specific

analysis replicates the downward trend demonstrated in the pooled data set. This is also the

same when the ‘multiplicative uniform cohort effect’ association model is simultaneously

estimated on the whole set of basic contingency tables (Figure 5), though some variance is

especially visible in the 1918-1922 to 1933-1937 birth cohorts.12

Comparison between males and females in strength and trend of origin – education

association

Research on Swedish data which applied the same statistical technique demonstrated an

equalization trend for both males and females and produced consistent results regarding

strength of association across the gender variable: Jonsson, Mills and Müller (1996, p. 194)

observed that “amongst the oldest cohorts in Sweden, class differences are greater for

females than for males, but amongst the youngest cohorts this is reversed and class

differences are stronger for men” and Jonsson and Erikson (2000, p. 371) concluded that “for

women the association between social origin and attainment of upper secondary and

university education were somewhat stronger than for men in older cohorts”. Figure 6

12 Research which investigated measurement error in social class variables concluded that

reliability is rather low for reports of class origin for older respondents (Breen and Jonsson

1997). This may provide a partial explanation for the variance in the 1918-1922, 1923-1927

and 1928-1932 birth cohorts as the estimates for the 1985 FQP survey (which, according to

Table 1, comprises the oldest respondents in each of these cohorts) are outliers.

22

investigates the same research question for France by applying the ‘multiplicative uniform

cohort effect’ association model to a set of 26, i.e. 2 sexes x 13 cohorts, contingency tables,

assuming a common pattern of association and fixing the log-multiplicative parameter at 1 for

males in the first birth cohort.

For both males and females, the estimates clearly show a downward trend in strength of

association between social origin and educational destination, but the equalization trend is

more marked among women. This corresponds to the level of association being distinctly

stronger for women than for men until cohorts born in the mid-1930s. Then the difference

progressively disappears and it is even reversed in the 1968-1972 cohort. An examination of

observed data confirms the latter result: for instance, the odds of getting an upper tertiary

education degree rather than no diploma are 71.4 times higher for daughters of teachers

born around 1970 than for daughters of agricultural and unskilled manual workers, but the

same odds ratio amounts to 129.0 for sons. Finally, sex-specific analyses also confirm the

more pronounced equalization trend among women as the log-multiplicative parameter fixed

at 1 in the first birth cohort peaks at 1.196 in the next one, then declines till 0.647 in the last

one while corresponding estimates for men are respectively 0.978 and 0.608.

Reconciling results from the contingency table approach with the educational transition

approach and investigating the “maximally maintained inequality” hypothesis

The key proposition in this hypothesis states that “transition rates and odds ratios between

social origins and educational transitions remain the same from one cohort to another unless

forced to change by increasing enrollments” (Hout, Raftery and Bell 1993, p. 25). More

specifically, “if demand for a given level of education is saturated for the upper classes, that

is, if some origin-specific transition rates approach or reach 100%, then the odds ratios

decrease (that is, association between social origin and education is weakened)”. Testing

this hypothesis implies that we must leave the general contingency table approach to adopt

an educational transition approach and perform “surviving” analyses in the Mare’s tradition

23

on the same data. Starting from our seven-category variable describing final educational

destination, we thus assume that it results from a sequence of school continuation decisions

which can be described by five educational (or certification) transitions:

- getting any diploma v. no diploma at all (first transition);

- among those who succeeded in any diploma, getting at least a lower secondary or lower

vocational education diploma v. getting only a primary education certificate (second

transition);

- among those who survived the second transition, getting at least an upper secondary or

technical education diploma v. getting only a lower secondary or lower vocational education

diploma (third transition);

- among those who survived the third transition, getting a tertiary education degree v. getting

only an upper secondary or technical education diploma (fourth transition);

- finally, among those who succeeded in a tertiary education degree, getting an upper as

opposed to a lower tertiary education degree (fifth transition).

As Table 2 shows that observed rates characterizing the two most advantaged classes were

already close to saturation from the 1938-1942 birth cohort for the first two transitions 13 but

not the subsequent ones, a marked equalization trend should only have existed in the former

case in order to provide support for the “maximally maintained inequality” hypothesis. To

examine this prediction Table 4 contrasts the goodness of fit provided by the constant

association model and the ‘multiplicative uniform cohort effect’ association model for each of

the five transitions. The latter model very significantly improves on the former model in the

first and second transitions, but the improvement in fit is barely significant or not significant in

the third, fourth and fifth transitions. Moreover, by displaying the temporal pattern of log-

13 In that cohort, 93.2% of children of teachers (and assimilated occupations) and 85.7% of

children of higher-grade professionals and managers got a lower secondary or lower

vocational education diploma at least.

24

multiplicative parameters estimated for each of the five transitions, Figure 7 affords

consistent empirical support to the “maximally maintained inequality” hypothesis in the

French case. As regards the first transition, a downward trend in the general strength of

social origin effects clearly appears from the early decades of twentieth century. A decline is

also visible in the second transition from the 1938-1942 birth cohort to the last one – the

parameter falls from 0.997 till 0.742. On the contrary, remarkably constant social origin

effects characterize the third transition between the 1928-1932 and 1968-1972 birth cohorts

– all the estimated parameters lie between 1.196 and 1.124. Finally, a slow but nearly

monotonic increase in social origin effects appears for the fourth transition from the 1938-

1942 birth cohort14 and there is also a sharp rise over the last three cohorts in the fifth

transition.15 As mentioned in the introduction, these upward trends in social origin effects

may be related to the educational expansion which, by increasing the proportion of the total

population at risk at a given transition, also increases its heterogeneity on unmeasured

determinants of school continuation.

However, it must be emphasized that “maximally maintained inequality” is not synonymous

with “absolutely persistent inequality” because declining trends in social origin effects in early

transitions involve consequences for (unconditional) odds ratios associated with attainment

of upper diplomas. This result can be highlighted by collapsing all educational destinations in

only two categories, then comparing different dichotomies, or cut-off points, in the

educational system. We examined three dichotomies: at least lower secondary or lower

vocational education diploma v. less than that; at least upper secondary or technical

14 The estimated parameter is 0.837 in the 1938-1942 birth cohort, but 1.009 in the 1963-

1967 birth cohort and 0.967 in the 1968-1972 birth cohort. 15 As regards the very last transition, constraining the log-multiplicative parameter to be

exactly the same for the first four five-year birth cohorts seems insufficient to get reliable

estimates of the temporal pattern over the whole period (see Figure 7). However, such a

limitation does not impede comparability of parameters estimated for the most recent

cohorts.

25

education diploma v. less than that; at least lower tertiary education degree v. less than that.

Figure 8 displays trends independently estimated for each educational dichotomy using the

‘multiplicative uniform cohort effect’ association model. Although the first dichotomy is the

only one specifically associated with the first and second transitions characterized by

declining social origin effects, equalization trends also appear between the 1938-1942 and

1968-1972 cohorts for the second and third dichotomies, though on a slightly reduced scale

as compared to trend for the first dichotomy. This analysis therefore highlights to what extent

conditional and unconditional analyses can lead to different results on the same topic and the

same data.

Sensitivity of estimated trends to use of various indicators of social background

Research on Dutch data which investigated both the determination of final education level

and educational transitions concluded that the effect of father’s education did not decrease

as dramatically over time as the effect of father’s occupation, i.e. that “father’s education is

more resistant to change than father’s occupation [which is a pattern which] confirms the

expectations of cultural reproduction theory rather than those of modernization theory” (De

Graaf and Ganzeboom 1993, p. 97). In order to assess the sensitivity of estimated trends to

the definition of social background in the French case, we restricted the analysis to the 1977,

1985 and 1993 FQP surveys which permit to observe social background both in the

socioeconomic and cultural dimension.16

Figure 9 presents trends estimated over twelve five-year birth cohorts using the

‘multiplicative uniform cohort effect’ association model and three different definitions of social

background: the combination of father’s occupation and mother’s occupation (10 categories)

16 In that case, we do not use the original file for the 1977 FQP survey (with N=28,105 as

indicated in Table 1), but another file which includes the recoding of father’s occupation in

the new French socio-occupational classification introduced in 1982 (with N=28,113).

26

only involves the socioeconomic dimension; parents’ highest diploma (6 categories) only

involves the cultural dimension; the combination of father’s occupation and mother’s highest

diploma (10 categories) is a mixture of both dimensions. The ‘multiplicative uniform cohort

effect’ association model significantly improves on the constant association model in each

case, but the former model is not preferable to the latter on the basis of the BIC statistic when

social background is approached by parents’ highest diploma. This corresponds to the

estimated downward trend being less marked with this variable than with the two others.

Such a result is also confirmed by linearly restricting the log-multiplicative trend in the

estimation process: the annual pace of change amounts to -0.86% with the combination of

father’s occupation and mother’s occupation, -0.80% with the combination of father’s

occupation and mother’s highest diploma, but -0.49% only with parents’ highest diploma. In

the French case the existence of a decline in inequality of educational opportunity therefore

does not depend on the variable used to define social background, but the scale of the

decline is sensitive to the chosen indicator and the results strongly suggest that cultural

inequalities are more resistant to change than socioeconomic inequalities.

Examining the parameters of the best-fitting model

All the analyses above derive from the simplest log-multiplicative model which assumes that

change only occurred in strength of the origin – education association. However, the

‘regression-type cohort effect’ association model has proved the best-fitting model, which

implies that significant variations over birth cohorts occurred not only in strength, but also in

pattern of association between social origin and educational destination. Table 5 therefore

presents the corresponding parameters estimated on all men and women.

As a consequence of identifying constraints for the γc parameters, the set of λ

OE

oe

parameters describes the origin – education association in the last (1968-1972) birth cohort

while the set of ϕoe

parameters expresses the pattern of deviation characteristic of the first

27

(1908-1912) birth cohort as compared to the last one. Apart from an exception between the

1918-1922 and 1923-1927 cohorts the strength of the adjustment of the association declines

monotonically over the period, but change is especially marked between the 1933-1937 and

1948-1952 birth cohorts.17

In the 1968-1972 cohort, the highest diplomas (upper and lower tertiary degrees) are strongly

and positively associated with the most advantaged classes of origin (teachers and

assimilated occupations, then higher-grade professionals and managers). Conversely, they

are strongly and negatively associated with origins in the most disadvantaged classes

(agricultural and unskilled manual workers, then foremen and skilled manual workers, and

farmers and smallholders for upper tertiary degree only). An opposite pattern characterizes

the association between these social groups and the “no diploma” and “primary education

certificate” destinations. Finally, getting a lower vocational education diploma is positively

associated with origins in the class of farmers and smallholders and the class of agricultural

and unskilled manual workers, and negatively associated with the two most advantaged

classes.

Taking the absolute value of ϕoe

parameters and computing their average by row and

column highlights those classes of origin whose relative position has been transformed most

and those educational destinations with the most important contributions to historical change

in the pattern of association. It is especially visible that, in the past as compared to the 1968-

1972 cohort, the “no diploma” category was more strongly positively associated with

disadvantaged classes (especially farmers and smallholders) and more strongly negatively

associated with the two most advantaged classes and also the class of lower-grade

17 It is certainly difficult to provide a firm explanation for the irregularity in the 1923-1927

cohort, but it is worth noting that it was the first cohort affected by the 1933-1934 reform

which introduced the obligation of passing an examination to enter secondary education.

28

professionals and technicians and the class of routine non manual workers. Examining the

profile of ϕoe

parameters by row reveals that, as regards their educational destinations,

sons and daughters of farmers and smallholders considerably improved their relative position

over sixty years while the situation of children of artisans and shopkeepers remained much

more stable. In the early decades of the twentieth century, men and women originating in

four classes were characterized by more favorable (relative) educational trajectories than

they are in the most recent cohorts: these are the classes of teachers and assimilated

occupations, higher-grade professionals and managers, lower-grade professionals and

technicians, and routine non manual workers. Finally, historical change in the educational

destinations of the offspring of skilled and unskilled fractions of the working class appears

more complex. Children of foremen and skilled manual workers have improved their relative

position more than children of agricultural and unskilled manual workers. In the first birth

cohorts as compared to the last ones, the former were more positively associated with the

“no diploma” and “lower vocational diploma” categories and more negatively associated with

the “upper tertiary degree” category, but association did not change much between the

offspring of the unskilled fraction of the working class and the highest educational

destinations (upper secondary diploma, lower and upper tertiary degrees).

How much difference does the reduction in inequality of educational opportunity make?

To assess the concrete effects of change in inequality of educational opportunity, we start

with the origin – education tables estimated using the best-fitting model, i.e. the ‘regression-

type cohort effect’ association model. These tables have been constructed from a sample.

Firstly, by multiplying by an appropriate factor we can adjust the estimated table for the last

cohort in accordance with the margins that characterize French society in this generation, i.e.

the social origins and educational destinations of all French-born men and women born

between 1968 and 1972. We shall refer to this table as T. We now turn to the estimated

origin – education table for the first (1908-1912) birth cohort. If we apply the margins which

29

characterize French society in the last birth cohort to this we obtain a counterfactual table

showing what would have been the case in the 1968-1972 cohort if the association between

social origin and educational destination had remained exactly the same as it was in early

20th-century France. In order to do this we simply use the Deming-Stephan algorithm whose

essential feature is that it conserves all the odds ratios of the initial table. If we refer to the

resulting table as T’ we finally compute for each cell the difference T – T’ in order to show the

“reallocation” which occurred in the 1968-1972 cohort solely as a result of change in

inequality of educational opportunity over sixty years. These differences, rounded to the

nearest thousand, are given in Table 6 which also includes a second assessment of the

effect of change in the origin – education association based on the median (1938-1942) birth

cohort.

From the first table, it can be concluded that around 415,000 men and women, i.e. about

11% of all members in the 1968-1972 cohort, have educational destinations which differ from

those they would have held were the origin – education association strictly constant over

sixty years. Change is considerable for children of farmers and smallholders: 51% of them

hold different – and, in that case, better – educational destinations as a result of changing

association and equalization trend. This is the case for 7% of children of higher-grade

professionals and managers, of lower-grade professionals and technicians, and of

agricultural and unskilled manual workers; for 8% of children of artisans and shopkeepers;

and for 10% of children of teachers and assimilated occupations, of routine non manual

workers, and of foremen and skilled manual workers. Finally, it can be computed that change

in the association between social origin and educational destination over six decades

resulted in 101,000 “additional” men and women originating from disadvantaged classes, i.e.

the peasantry or the skilled or unskilled fractions of the working class, with diplomas in the

upper secondary, lower tertiary or upper tertiary categories; they represent nearly 6% of all

men and women in the 1968-1972 cohort with background in these social groups.

30

From the second table, it can similarly be concluded that changing origin – education

association over thirty years resulted in a reallocation of 216,000 men and women, i.e. about

6% of the last cohort, and that 55,000 “additional” men and women with background in the

three aforementioned disadvantaged classes rejoined the three highest educational

categories as a consequence of the democratization process per se.

Discussion and conclusion

In the conclusion of the article about France in which they demonstrated “a notable decline

among cohorts born in the 1940s in the strong association between father’s occupation and

highest degree obtained, and few sex differences in the trend in this association”, Smith and

Garnier (1986, pp. 339-40) stated: “We must temper these substantive findings by pointing

out the following. First, because our view of educational attainment is truncated beginning

with the birth cohort of 1950, we are unable to tell whether trends beginning with the cohorts

born during the 1940s were continued. Our suspicion is that an extension of this analysis to

more recent cohorts would confirm departures from the exceedingly high background-degree

association among individuals born during the 1920s and 1930s; however, we can hardly

rule out the possibility that what appears to us to be the onset of some diminution in this

association is in longer perspective a transitory phenomenon”.

In this article we have used seven nationally representative French surveys from 1964 to

1997 with a total sample of more than 240,000 cases as well as the most recent

methodologies in log-multiplicative contingency table analysis to replicate and extend Smith

and Garnier’s study on thirteen five-year cohorts of French-born men and women born

between 1908 and 1972. Our results undoubtedly corroborate Smith and Garnier’s previous

findings. In French society the general strength of the association between social origin and

educational destination has declined by more than 35% (in the logged odds ratios) over sixty

years – a result which parallels recent research about France which demonstrated increasing

intergenerational social fluidity between the 1950s and the 1990s (Vallet 2001). While it has

31

been nearly monotonic, change in the origin – education association was especially sharp

between the 1933-1937 and 1948-1952 birth cohorts, then largely leveled off in the three

subsequent cohorts, but took off again in the very last one. The decline in origin – education

association in France therefore seems largely independent of major secondary school

reforms explicitly introduced from the late 1950s to promote equality of educational

opportunity. The downward trend was more pronounced among women than men, especially

because the former were characterized by stronger origin – education association until

cohorts born in the mid-1930s. It seems to conform to predictions derived from the

“maximally maintained inequality” hypothesis and reveals itself as quite robust as its

existence does not depend on the variable used to define social background – though

change in origin – education association appears more resistant to cultural inequalities than

to socioeconomic inequalities. We have finally demonstrated that the improvement of

educational trajectories of children of farmers and smallholders played a major part in

accentuating the trend though not in creating it, and we have also assessed how much

difference in society the reduction in inequality of educational opportunity has made.

From a comparative perspective France therefore appears as the fourth national case – after

Sweden, the Netherlands and Germany – to challenge the “persistent inequality” conclusion

in education which was reached in the major comparative project directed by Shavit and

Blossfeld (1993). This strongly suggests that, in the current state of the art of educational

stratification studies, two research enterprises are entirely legitimate and have to be pursued:

first explaining, in the line of, e.g., Breen and Goldthorpe (1997), why class or socioeconomic

differentials in educational attainment exhibit a so strong inertia over time; second explaining

why, despite this considerable inertia, long-term investigations based on very large

representative samples and sufficiently powerful statistical techniques, are nonetheless able

to demonstrate a steady trend toward decreasing inequality which seems largely

independent of major educational reforms. Finally, it seems that, to pursue the second

enterprise, it is especially valuable to reconcile the global contingency table approach of

32

educational stratification with the more local educational transition approach which has long

been the dominant perspective in historical studies. While the educational transition

approach is based on a series of successive and conditional analyses which take account of

progressively reduced fractions of the entire population, the contingency table approach is

based on a single and unconditional analysis of the whole population. And, as we have

demonstrated for French society in this article, the general downward trend in origin –

education association is in fact compatible with remarkably stable, or even increasing social

origin effects in transitions associated with advanced stages of the educational system, but is

generated by the consequences of declining origin effects in the most basic transitions,

notably that one associated with the odds of getting any qualification and that one associated

with the odds of getting more than an elementary qualification.

The general title of this paper can therefore be understood in two different meanings. The

dynamics of inequality of educational opportunity primarily refers to the historical general

downward trend in origin – education association that we have documented for French

society. But the same expression secondarily refers to the contrasted and progressively

changing trends which characterize the association between social origins and completion of

successive educational transitions: as education expands over time, the earliest transitions

reveal a progressively decreasing association with social origins while the latest ones exhibit

a progressively increasing association with the same variable. On the basis of that result,

one might argue that the general downward trend in origin – education association is

essentially an artefact which only comes from the fact that each educational transition is

equally weighted over the entire covered period or birth cohorts. Such an argument would be

clearly related to the conception of education as a “positional good” (with, for instance, the

idea that holding a university degree is more important today for the future of individuals than

it was some decades ago). There is probably some truth in such an argument, but what

prevents us to accept it fully is that, even with the educational expansion, the earliest

educational transitions have not lost their own importance: in French society there is still a

33

significant fraction of the 1968-1972 birth cohort which has left the educational system

without holding any diploma.

34

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38

TABLE 1

OVERALL DESIGN TO STUDY THE ASSOCIATION BETWEEN SOCIAL ORIGIN AND EDUCATIONAL

DESTINATION OVER THIRTEEN FIVE-YEAR BIRTH COHORTS USING SEVEN SURVEYS

Birth cohort FQP 1964 FQP 1970 FQP 1977 FQP 1985 FQP 1993 Emploi 1993

Emploi 1997 Sample size

1908-1912 aged 58-62 [ (2,304) ]

aged 65-69 (1,273) 1,273

[ 3,577 ]

1913-1917 aged 53-57 [ (1,920) ]

aged 60-64 (1,372) 1,372

[ 3,292 ]

1918-1922 aged 42-46 (2,219)

aged 48-52 (2,756)

aged 55-59 (2,218)

aged 63-67 (1,264) 8,457

1923-1927 aged 37-41 (2,635)

aged 43-47 (3,561)

aged 50-54 (3,361)

aged 58-62 (2,073) 11,630

1928-1932 aged 32-36 (2,622)

aged 38-42 (3,630)

aged 45-49 (3,477)

aged 53-57 (2,729) 12,458

1933-1937 aged 27-31 (2,349)

aged 33-37 (3,369)

aged 40-44 (3,351)

aged 48-52 (2,953)

aged 56-60 (1,505)

aged 56-60 (8,258) 21,785

1938-1942 aged 28-32 (3,394)

aged 35-39 (3,050)

aged 43-47 (2,813)

aged 51-55 (1,437)

aged 51-55 (7,467)

aged 55-59 (7,332) 25,493

1943-1947 aged 30-34 (4,872)

aged 38-42 (3,767)

aged 46-50 (1,850)

aged 46-50 (9,778)

aged 50-54 (9,078) 29,345

1948-1952 aged 25-29 (5,131)

aged 33-37 (5,044)

aged 41-45 (2,107)

aged 41-45 (12,239)

aged 45-49 (11,597) 36,118

1953-1957 aged 28-32 (5,212)

aged 36-40 (2,068)

aged 36-40 (11,928)

aged 40-44 (11,481) 30,689

1958-1962 aged 31-35 (1,942)

aged 31-35 (11,689)

aged 35-39 (11,725) 25,356

1963-1967 aged 26-30 (1,904)

aged 26-30 (11,803)

aged 30-34 (11,621) 25,328

1968-1972 aged 25-29 (11,063) 11,063

Sample size 9,825 16,710 [ 20,934 ] 28,105 25,855 12,813 73,162 73,897 240,367

[ 244,591 ]

In the 1964 FQP survey the contingency table cross-classifying social origin and educational

destination can be built on a sample of 2,219 French-born men and women in the 1918-1922 birth

cohort, i.e. aged between 42 and 46 at the time of the survey.

As each survey used a complex sampling design, each contingency table was computed using

appropriate reweighting, then downscaled without altering its structure to reflect the exact sample

surveyed, e.g. 2,219 for the 1918-1922 birth cohort in the 1964 FQP survey. The same process

was also replicated separately on tables for males and tables for females.

The 1970 FQP survey data for the 1908-1912 and 1913-1917 birth cohorts cannot be used in all

the analyses as they do not distinguish between lower and upper tertiary education.

39

TABLE 2

EDUCATIONAL DESTINATIONS FOR EACH CATEGORY OF SOCIAL ORIGINS IN THE 1908-1912 BIRTH

COHORT (N=3,577), THE 1938-1942 BIRTH COHORT (N=25,493) AND THE 1968-1972 BIRTH COHORT

(N=11,063)

Birth cohort No diploma Primary

education certificate

Lower secondary diploma

Lower vocational diploma

Upper secondary diploma

Lower/upper tertiary degree

Total

Farmers and smallholders 1908-1912 66.1 28.4 1.3 2.3 1.1 0.9 100

1938-1942 28.0 40.2 4.6 18.0 4.5 4.6 100

1968-1972 9.6 0.8 2.3 33.3 21.1 32.9 100

Artisans and shopkeepers 1908-1912 38.2 45.1 5.6 6.2 3.5 1.4 100

1938-1942 14.2 24.9 10.2 24.9 12.4 13.5 100

1968-1972 12.8 1.4 5.6 31.4 15.8 33.1 100

Higher-grade professionals 1908-1912 19.7 24.9 12.3 12.5 16.0 14.6 100

and managers 1938-1942 7.1 7.3 8.3 12.8 20.5 44.0 100

1968-1972 4.9 0.1 3.0 8.7 18.6 64.8 100

Teachers and assimilated 1908-1912 17.1 25.7 8.6 7.3 21.6 19.8 100

occupations 1938-1942 4.9 2.0 7.2 11.3 18.9 55.7 100

1968-1972 4.2 0.3 2.5 8.0 15.6 69.4 100

Lower-grade professionals 1908-1912 15.2 35.1 15.6 16.5 12.4 5.2 100

and technicians 1938-1942 9.6 14.0 10.9 24.6 18.3 22.5 100

1968-1972 7.4 0.3 4.4 18.3 20.4 49.3 100

Routine non manual 1908-1912 39.1 38.1 5.5 10.3 4.1 2.9 100

workers 1938-1942 15.4 21.7 9.4 28.3 12.6 12.6 100

1968-1972 14.5 0.7 5.4 31.2 19.5 28.6 100

Foremen and skilled 1908-1912 45.9 37.6 3.6 9.3 2.3 1.3 100

manual workers 1938-1942 20.8 30.1 5.6 29.1 8.3 6.1 100

1968-1972 19.1 0.8 5.5 35.2 18.1 21.4 100

Agricultural and unskilled 1908-1912 65.2 27.8 1.1 4.8 0.8 0.3 100

manual workers 1938-1942 30.2 33.4 4.7 23.4 4.7 3.6 100

1968-1972 27.3 1.7 6.6 38.2 14.1 12.2 100

Total 1908-1912 51.5 32.7 3.8 6.2 3.4 2.4 100 1938-1942 20.8 28.1 6.7 23.3 9.5 11.6 100

1968-1972 15.0 0.8 5.0 28.6 17.7 32.9 100

40

TABLE 3

RESULTS OF FITTING THE NULL ASSOCIATION, CONSTANT ASSOCIATION, ‘MULTIPLICATIVE UNIFORM

COHORT EFFECT’ ASSOCIATION AND ‘REGRESSION-TYPE COHORT EFFECT’ ASSOCIATION MODELS TO

CONTINGENCY TABLES CROSS-CLASSIFYING SOCIAL ORIGIN (8 CATEGORIES) AND EDUCATIONAL

DESTINATION (7 CATEGORIES) OVER THIRTEEN FIVE-YEAR BIRTH COHORTS

Model G2 df DI rG2 Bic All men and women (N=240,367) 1. Null association 50,388.8 546 15.9 - 43,623.9 2. Constant association 4,107.8 504 4.1 91.8 -2,136.7 3. Multiplicative uniform cohort effect 3,442.1 492 3.9 93.2 -2,653.7 4. Regression-type cohort effect 799.2 451 1.7 98.4 -4,788.6 Men and women, disregarding children of farmers and smallholders (N=200,550) 1. Null association 43,928.6 468 16.6 - 38,214.9 2. Constant association 1,552.7 432 2.8 96.5 -3,721.5 3. Multiplicative uniform cohort effect 1,301.5 420 2.7 97.0 -3,826.2 4. Regression-type cohort effect 582.7 385 1.6 98.7 -4,117.7 Men only (N=127,229) 1. Null association 28,039.4 546 16.4 - 21,621.9 2. Constant association 2,587.4 504 4.6 90.8 -3,336.5 3. Multiplicative uniform cohort effect 2,190.6 492 4.3 92.2 -3,592.2 4. Regression-type cohort effect 647.8 451 2.1 97.7 -4,653.1 Women only (N=113,138) 1. Null association 23,773.5 546 16.1 - 17,420.1 2. Constant association 2,000.8 504 4.1 91.6 -3,864.0 3. Multiplicative uniform cohort effect 1,685.6 492 3.9 92.9 -4,039.5 4. Regression-type cohort effect 580.1 451 2.0 97.6 -4,667.9

41

TABLE 4

RESULTS OF FITTING THE CONSTANT ASSOCIATION AND ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’

ASSOCIATION MODELS TO FIVE SUCCESSIVE EDUCATIONAL TRANSITIONS OVER THIRTEEN FIVE-YEAR

BIRTH COHORTS

Model G2 df DI rG2 Bic First transition (N=240,367) 2. Constant association 1,094.2 84 1.8 90.5 53.5 3. Multiplicative uniform cohort effect 920.9 72 1.8 92.0 28.9 Second transition (N=189,603) 2. Constant association 1,050.8 84 2.1 92.1 30.0 3. Multiplicative uniform cohort effect 963.8 75 2.2 92.8 52.3 Third transition (N=142,588) 2. Constant association 209.4 84 1.1 98.9 -787.5 3. Multiplicative uniform cohort effect 191.1 75 1.0 99.0 -699.0 Fourth transition (N=63,739) 2. Constant association 140.0 84 1.5 95.5 -789.3 3. Multiplicative uniform cohort effect 127.4 75 1.4 95.9 -702.3 Fifth transition (N=38,065) 2. Constant association 117.6 84 1.7 93.5 -768.4 3. Multiplicative uniform cohort effect 98.4 75 1.4 94.6 -692.7

For the second, third, fourth, and fifth transitions, the log-multiplicative parameter has been

constrained to be the same for the first four five-year birth cohorts in order to get more reliable

estimates of the temporal pattern. Such a constraint does not result in a significant worsening of

the model fit at the 5% level for three degrees of freedom.

42

TABLE 5

PARAMETERS OF THE ‘REGRESSION-TYPE COHORT EFFECT’ ASSOCIATION MODEL APPLIED TO ALL MEN

AND WOMEN (N=240,367) – BASELINE PATTERN OF ASSOCIATION ( λOE

oe PARAMETERS), PATTERN OF

DEVIATION (ϕoe

PARAMETERS) AND STRENGTH OF DEVIATION (γc

PARAMETERS) OVER BIRTH

COHORTS

λOE

oe parameters No diploma

Primary education certificate

Lower secondary diploma

Lower vocational diploma

Upper secondary diploma

Lower tertiary degree

Upper tertiary degree

Farmers and smallholders -0.190 0.492 -0.021 0.529 0.033 -0.132 -0.711 Artisans and shopkeepers -0.036 0.042 0.032 0.125 -0.036 -0.054 -0.073 Higher-grade professionals

and managers -0.479 -1.064 -0.089 -0.640 0.284 0.650 1.339 Teachers and assimilated

occupations -0.583 -1.446 -0.201 -0.589 0.421 0.797 1.602 Lower-grade professionals

and technicians -0.271 -0.319 0.033 -0.203 0.123 0.294 0.342 Routine non manual workers 0.215 0.327 0.035 0.083 -0.072 -0.212 -0.375 Foremen and skilled manual

workers 0.508 0.821 0.031 0.268 -0.281 -0.501 -0.847 Agricultural and unskilled manual

workers 0.836 1.146 0.181 0.428 -0.472 -0.842 -1.277

ϕoe

parameters No diploma Primary

education certificate

Lower secondary diploma

Lower vocational diploma

Upper secondary diploma

Lower tertiary degree

Upper tertiary degree

Farmers and smallholders 2.325 0.883 -0.153 -0.828 -0.915 -1.006 -0.307 Artisans and shopkeepers 0.001 0.170 0.142 -0.032 -0.017 -0.235 -0.029 Higher-grade professionals

and managers -0.709 0.135 0.032 0.067 0.297 0.205 -0.027 Teachers and assimilated

occupations -1.333 0.227 0.544 -0.490 0.302 0.579 0.170 Lower-grade professionals

and technicians -0.847 -0.203 0.189 0.315 0.279 0.095 0.172 Routine non manual workers -0.531 -0.399 0.107 0.282 0.008 0.279 0.256 Foremen and skilled manual

workers 0.320 -0.400 -0.207 0.518 0.058 -0.021 -0.268 Agricultural and unskilled manual

workers 0.774 -0.414 -0.653 0.169 -0.013 0.104 0.033

Birth cohort 1908-1912 1913-1917 1918-1922 1923-1927 1928-1932 1933-1937 1938-1942

γc

parameters 1 0.891 0.758 0.867 0.772 0.673 0.519

Birth cohort 1943-1947 1948-1952 1953-1957 1958-1962 1963-1967 1968-1972

γc

parameters 0.357 0.241 0.200 0.175 0.101 0

The λOE

oe and ϕ

oe parameters are estimated using effect coding identifying constraints. The

software used does not compute the standard error on the parameters of a log-multiplicative

model.

43

TABLE 6

TWO ASSESSMENTS OF THE CONCRETE EFFECTS OF CHANGE IN INEQUALITY OF EDUCATIONAL

OPPORTUNITY ON MEMBERS OF THE LAST BIRTH COHORT (FRENCH-BORN MEN AND WOMEN BORN

BETWEEN 1968 AND 1972)

Effect of change between the 1908-1912 birth cohort and the 1968-1972 birth cohort

No diploma Primary

education certificate

Lower secondary diploma

Lower vocational diploma

Upper secondary diploma

Lower tertiary degree

Upper tertiary degree

Total (thousands)

Farmers and smallholders -107 -4 0 +57 +27 +23 +4 (219) Artisans and shopkeepers +13 -2 -10 +10 -10 +9 -10 (410) Higher-grade professionals

and managers +14 0 -1 +6 -18 -12 +11 (428) Teachers and assimilated

occupations +5 0 -2 +8 -1 -13 +4 (166) Lower-grade professionals

and technicians +23 0 -4 -1 -14 +4 -8 (376) Routine non manual

workers +59 +1 -7 -10 +4 -24 -22 (608) Foremen and skilled

manual workers +39 +2 +8 -90 +5 +15 +21 (921) Agricultural and unskilled

manual workers -44 +2 +17 +20 +7 -1 0 (616)

Total (thousands) (561) (31) (185) (1,071) (664) (614) (617) (3,744)

In the population of French-born men and women born between 1968 and 1972, 219,000 are

children of farmers and smallholders. As a result of change in the association between social origin

and educational destination over 60 years, 107,000 do not belong to the “no diploma” category and

4,000 do not hold a primary education certificate; 57,000 of these hold a lower vocational diploma.

Effect of change between the 1938-1942 birth cohort and the 1968-1972 birth cohort

No diploma Primary

education certificate

Lower secondary diploma

Lower vocational diploma

Upper secondary diploma

Lower tertiary degree

Upper tertiary degree

Total (thousands)

Farmers and smallholders -45 -2 -2 +27 +12 +11 0 (219) Artisans and shopkeepers +5 -1 -5 +5 -4 +5 -4 (410) Higher-grade professionals

and managers +8 0 -1 +3 -10 -7 +6 (428) Teachers and assimilated

occupations +4 0 -1 +5 -1 -7 +1 (166) Lower-grade professionals

and technicians +14 0 -2 -2 -8 +1 -4 (376) Routine non manual

workers +34 0 -4 -8 +2 -13 -11 (608) Foremen and skilled

manual workers +12 +1 +4 -43 +4 +9 +12 (921) Agricultural and unskilled

manual workers -31 +1 +10 +13 +5 +1 +1 (616)

Total (thousands) (561) (31) (185) (1,071) (664) (614) (617) (3,744)

44

In the population of French-born men and women born between 1968 and 1972, 219,000 are

children of farmers and smallholders. As a result of change in the association between social origin

and educational destination over 30 years, 45,000 do not belong to the “no diploma” category,

2,000 do not hold a primary education certificate and 2,000 do not hold a lower secondary diploma;

27,000 of these hold a lower vocational diploma.

45

FIGURE 1

TRENDS IN THE DISTRIBUTION OF SOCIAL ORIGINS (FATHER’S OCCUPATION) OVER THIRTEEN FIVE-

YEAR BIRTH COHORTS (N=244,591)

0

10

20

30

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Prop

ortio

n (%

)

Farmers and smallholders Artisans and shopkeepersHigher-grade professionals and managers Teachers and assimilated occupations

Lower-grade professionals and technicians Routine non manual workersForemen and skilled manual workers Agricultural and unskilled manual workers

46

FIGURE 2

TRENDS IN THE DISTRIBUTION OF EDUCATIONAL DESTINATIONS (HIGHEST DEGREE OBTAINED) OVER

THIRTEEN FIVE-YEAR BIRTH COHORTS (N=240,367)

0

10

20

30

40

50

60

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Prop

ortio

n (%

)

No diploma (or no information) Primary education certificateLower secondary education diploma Lower vocational education diploma

Upper secondary or technical education diploma Lower tertiary education degreeUpper tertiary education degree

47

FIGURE 3

CHANGE IN THE GENERAL STRENGTH OF ASSOCIATION BETWEEN SOCIAL ORIGIN AND EDUCATIONAL

DESTINATION – ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’ ASSOCIATION MODEL ESTIMATED OVER

THIRTEEN FIVE-YEAR BIRTH COHORTS (N=240,367)

0,500

0,600

0,700

0,800

0,900

1,000

1,100

1,200

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Log

-mul

tiplic

ativ

e pa

ram

eter

All social origins Disregarding farmers and smallholders

48

FIGURE 4

CHANGE IN THE GENERAL STRENGTH OF ASSOCIATION BETWEEN SOCIAL ORIGIN AND EDUCATIONAL

DESTINATION – ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’ ASSOCIATION MODEL ESTIMATED

SEPARATELY ON EACH SURVEY

0,500

0,600

0,700

0,800

0,900

1,000

1,100

1,200

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Log

-mul

tiplic

ativ

e pa

ram

eter

FQP 1964 (N=9,825) FQP 1970 (N=16,710) FQP 1977 (N=28,105) FQP 1985 (N=25,855)FQP 1993 (N=12,813) Emploi 1993 (N=73,162) Emploi 1997 (N=73,897)

For the purpose of comparison the first log-multiplicative parameter for each survey is adjusted to

reproduce its estimated value in the 1977 FQP survey exactly; the subsequent parameters are

adjusted accordingly.

49

FIGURE 5

CHANGE IN THE GENERAL STRENGTH OF ASSOCIATION BETWEEN SOCIAL ORIGIN AND EDUCATIONAL

DESTINATION – ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’ ASSOCIATION MODEL ESTIMATED

SIMULTANEOUSLY ON THE WHOLE SET OF BASIC CONTINGENCY TABLES (N=240,367)

0,500

0,600

0,700

0,800

0,900

1,000

1,100

1,200

1,300

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Log

-mul

tiplic

ativ

e pa

ram

eter

FQP 1964 FQP 1970 FQP 1977 FQP 1985 FQP 1993 Emploi 1993 Emploi 1997

50

FIGURE 6

CHANGE IN THE GENERAL STRENGTH OF ASSOCIATION BETWEEN SOCIAL ORIGIN AND EDUCATIONAL

DESTINATION – ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’ ASSOCIATION MODEL ESTIMATED

SIMULTANEOUSLY ON TABLES FOR MALES AND FOR FEMALES

0,500

0,600

0,700

0,800

0,900

1,000

1,100

1,200

1,300

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Log

-mul

tiplic

ativ

e pa

ram

eter

Men (N=127,229) Women (N=113,138)

51

FIGURE 7

CHANGE IN THE GENERAL STRENGTH OF ASSOCIATION BETWEEN SOCIAL ORIGIN AND SURVIVING IN AN

EDUCATIONAL TRANSITION – ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’ ASSOCIATION MODEL

ESTIMATED CONSIDERING FIVE SUCCESSIVE EDUCATIONAL TRANSITIONS

0,500

0,750

1,000

1,250

1,500

1,750

2,000

2,250

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Log

-mul

tiplic

ativ

e pa

ram

eter

First transition Second transition Third transition Fourth transition Fifth transition

52

FIGURE 8

CHANGE IN THE GENERAL STRENGTH OF ASSOCIATION BETWEEN SOCIAL ORIGIN AND EDUCATIONAL

DESTINATION – ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’ ASSOCIATION MODEL ESTIMATED

INDEPENDENTLY CONSIDERING THREE DIFFERENT EDUCATIONAL DICHOTOMIES

0,500

0,600

0,700

0,800

0,900

1,000

1,100

1,200

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67 68-72

Birth cohort 19 ...

Log

-mul

tiplic

ativ

e pa

ram

eter

At least lower secondary or lower vocational education diploma v. less than that (N=244,591)

At least upper secondary or technical education diploma v. less than that (N=244,591)At least lower tertiary education degree v. less than that (N=244,591)

53

FIGURE 9

CHANGE IN THE GENERAL STRENGTH OF ASSOCIATION BETWEEN SOCIAL BACKGROUND AND

EDUCATIONAL DESTINATION – ‘MULTIPLICATIVE UNIFORM COHORT EFFECT’ ASSOCIATION MODEL

ESTIMATED INDEPENDENTLY CONSIDERING THREE DEFINITIONS OF SOCIAL BACKGROUND

0,500

0,600

0,700

0,800

0,900

1,000

1,100

1,200

08-12 13-17 18-22 23-27 28-32 33-37 38-42 43-47 48-52 53-57 58-62 63-67

Birth cohort 19 ...

Log

-mul

tiplic

ativ

e pa

ram

eter

Combination of father's occupation and mother's highest diploma (N=66,781)

Combination of father's occupation and mother's occupation (N=66,781)Parents' highest diploma (N=66,781)