Department of Psychiatry and Neuropsychology, Maastricht University, European Graduate School of Neuroscience, The Netherlands
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Correspondence: Dr J. van Os, Maastricht University, Department of Psychiatry and Neuropsychology, European Graduate School of Neuroscience, PO Box 616, 6200 MD Maastricht, The Netherlands. Tel: +31 43 3299783; Fax: +31 43 3299708
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ABSTRACT |
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Aims To examine these issues in a multi-level model of schizophrenia incidence.
Method Cases of schizophrenia, incident between 1986 and 1997, were identified from the Maastricht Mental Health Case Register. A multi-level analysis was conducted to examine the independent effects of individual-level and neighbourhood-level variables in 35 neighbourhoods.
Results Independent of individual-level single and divorced marital status, an effect of the proportion of single persons and proportion of divorced persons in a neighbourhood was apparent (per 1% increase respectively: RR=1.02; 95% CI 1.00-1.03; and RR=1.12, 95% CI 1.04-1.21). Single marital status interacted with the neighbourhood proportion of single persons, the effect being stronger in neighbourhoods with fewer single-person households.
Conclusions The neighbourhood environment modifies the individual risk for schizophrenia. Premorbid vulnerability resulting in single marital status may be more likely to progress to over disease in an environment with a higher perceived level of social isolation.
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INTRODUCTION |
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METHOD |
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Maastricht is a relatively small city (population 120 000), located in the extreme south of The Netherlands in the province of Limburg. There are strong local traditions and Limburg has its own, officially recognised, dialect. The neighbourhoods of Maastricht represent traditional and sociologically meaningful entities, not arbitrary administrative subdivisions. Compared with the densely populated and more industrialised areas of the north-west of the country, levels of immigration of foreign nationals over the past decades have been low. A national insurance scheme covers mental health services and referral by a general practitioner is not necessary for attending the CMHC. Access to mental health services does not depend on the neighbourhood level of deprivation.
The period of investigation for the present study was 1981-1997. The case sample was defined by four criteria: (a) age 15-64 years; (b) having been coded as living in the city of Maastricht; (c) and ICD-9 diagnosis of schizophrenia and related disorders (ICD-295.x and 297.x; World Health Organization, 1978), recorded at least once during a psychiatric career; (d) in order to skew the sample towards true incident cases of schizophrenia, subjects registered in the first five years of the register (many of whom would have been prevalent cases who were in treatment when the register opened) were excluded, leaving subjects registered during the period 1986-1997. By confining the analyses to the city of Maastricht, and excluding the surrounding villages, any effect of distance to psychiatric services was minimised, as within the city of Maastricht all distances to mental health services can easily be covered by bicycle.
Individual-level variables and their neighbourhood-level
equivalents
The register routinely collects information on age, gender, marital status
and neighbourhood. Four-dimensional population data in the age range 15-64
years (age, gender, marital status and neighbourhood) for this period were
obtained from the municipal authorities for each of the years of the period
under investigation, allowing us to express the variables age, gender and
marital status of the population aged 15-64 years at the neighbourhood level.
Thus, for each neighbourhood we calculated, for the period 1986-1997, the
proportion of men, the proportion who were single, divorced and widowed (if
these three are known, the proportion of married persons can be derived,
therefore the proportion of married persons was not analysed separately) the
proportion aged under 25 years and the proportion aged 55 years and older
(Table 1).
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Neighbourhood-level measures of deprivation
(Table 1)
In order to classify neighbourhood according to level of deprivation, we
requested from the municipal authority information on seven socio-economic
variables to characterise the neighbourhoods of Maastricht over the period
1981-1997. Over this period, the mean values were calculated for:
Very small neighbourhoods or neighbourhoods consisting mainly of industrial compounds (n=6) were excluded, leaving 35 neighbourhoods with a median yearly total population size of 2804 over the period of investigation (interquartile range: 1718-4383).
Analyses
For descriptive purposes, adjusted incidence rates were calculated for each
neighbourhood over the period 1986-1997, using the ISTDIZE procedure in the
STATA statistical program (StataCorp,
1999). Indirect standardisation was used to adjust for age
(10-year age groups), gender and marital status (married, single, divorced,
widowed). The standardisation used the stratum-specific rates of the standard
population (the total population over the period 1986-1997) to calculate the
expected number of cases in the study populations, which were then used to
calculate adjusted rates. The standardised incidence ratio (SIR) is the ratio
of the observed over the expected number of cases in the study
populations.
Two types of neighbourhood effects were examined: (a) the neighbourhood random effect relates to the question: are neighbourhoods different with regard to schizophrenia incidence? (b) Neighbourhood fixed effects relate to the question: what makes neighbourhoods different (i.e. do certain neighbourhood characteristics, such as the proportion of unemployed or proportion of single persons have an effect on schizophrenia incidence)? Individual-level variables in this study all represent fixed effects. The effect of individual-level and neighbourhood-level characteristics was estimated using the multi-level Poisson regression procedure of the MLwiN program (Goldstein et al, 1998). In this analysis, the coefficient of any explanatory variable may be random at the two levels of the hierarchy. Further, at each level, the random coefficients may have any pattern of variances and covariances. Count variables (in this case the incidence of schizophrenia) do not have a normal distribution. Because of this, using ordinary least squares regression with a count as the dependent variable is not appropriate. We used the log of the counts and estimated the regression parameters using the Poisson maximum-likelihood algorithm, adjusting for person-years of observation. Effect sizes for fixed effects were expressed as incidence rate ratios (RR). In view of gender differences in the effect of age and marital status on the incidence rate of schizophrenia (Riecher Rossler et al, 1992; Tien & Eaton, 19992; Jablensky & Cole, 1997) interaction terms for gender-by-age and gender-by-marital-status were fitted into the models.
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RESULTS |
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Random neighbourhood effect
The incidence, standardised with respect to age, gender and marital status,
varied from 0 to 51 per 100 000 person-years in different neighbourhoods.
There was wide variation in the SIR, but only in one neighbourhood was it
significantly higher than the rate for the standard population
(Table 2). The multi-level
model without any fixed-effect explanatory variable
(Table 3) showed a level-2
variance (representing the random neighbourhood effect) of 0.14 (95% CI
0.00-0.29; P=0.055), constituting 12% of the total variance
(0.14/{1+0.14}).
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Individual-level variables and their neighbourhood-level
equivalents
The level-2 variance was minimally, or not at all, reduced after adjustment
for individual-level age and gender, even though these factors had a
significant effect on rates. Thus, for women (RR=0.62; 95% CI 0.47-0.82;
P=0.001) and older people (RR=0.90; 95% CI 0.82-0.99;
P=0.038) the rates were lower. The level-2 variance was reduced by
more than 60% after adjustment for individual-level marital status
(Table 3). Single persons
(RR=3.95, 95% CI 2.86-5.45), and divorced persons (RR=3.31, 95% CI 2.01-5.43)
were at greater risk than married persons. There was a strong interaction
between age and gender (P=0.008), in that schizophrenia was
associated with younger age in men (RR=0.80, 95% CI 0.70-0.91,
P=0.001) but not in women (RR=1.05, 95% CI 0.90-1.22,
P=0.57). There was also an interaction between gender and single
marital status (P=0.001), such that the size of the effect of being
single was greater in men (RR=6.54, 95% CI 4.09-10.46, P<0.001)
than in women (RR=2.04, 95% CI 1.27-3.28, P=0.003). Marital status
expressed at the neighbourhood level (but not age and gender) also affected
the incidence of schizophrenia in the same direction as the individual-level
variable, even after adjustment for the individual-level variables and their
interactions (Table 3). Thus,
after adjustment for individual-level age, gender, marital status and the
gender-by-age and gender-by-marital-status interactions, the risk of
schizophrenia was increased with the proportion of divorced persons (RR=1.12
per 1% increase, 95% CI 1.04-1.21, P=0.003) and the proportion of
single persons (RR=1.02 per 1% increase, 95% CI 1.00-1.03,
P=0.013).
Neighbourhood-level deprivation variables
The effect of the seven neighbourhood-level deprivation variables was in
the direction of increased incidence of schizophrenia; with the exception of
number of new houses built since the Second World War, which had a protective
effect (Table 3). The effect of
most variables was statistically significant. After adjustment for
individual-level age, gender, marital status and the age-by-gender and
gender-by-marital-status interactions, only the effects of being foreign-born,
unemployed and dependent on welfare remained significant
(Table 3).
Independence of neighbourhood-level variables
In order to assess their independence of each other, the proportion of
persons who were divorced, foreign-born, unemployed and on welfare were
entered simultaneously in the unadjusted model. The effects of the proportion
of single (RR=1.02, 95% CI 1.01-1.04, P=0.005) and the proportion of
divorced persons (RR=1.15, 95% CI 1.01-1.32, P=0.040) remained, but
not the effects of the proportion of unemployed (RR=1.00, 95% CI 0.80-1.24,
P=0.98), the proportion on welfare (RR=0.94, 95% CI 0.79-1.11,
P=0.48), and the proportion foreign-born (RR=1.13, 95% CI 0.93-1.39,
P=0.22).
Interaction between neighbourhood-level variables and
individual-level equivalents
The independent effect of the proportion of persons living alone was
modified by its individual-level equivalent (single marital status) in the
model adjusted for individual-level age, gender, marital status and the
age-by-gender and gender-by-marital-status interactions (P<0.001).
Thus, in neighbourhoods where the proportion of persons living alone was below
the Maastricht mean (Table 1),
the effect of single marital status in the adjusted model was more than twice
as large (RR=10.33, 95% CI 5.56-19.20, P<0.001) as the effect in
neighbourhoods with values above the mean (RR=4.22, 95% CI 1.92-9.30,
P<0.001). There was no interaction between the proportion of
divorced persons and individual-level divorced marital status
(P=0.81).
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DISCUSSION |
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Methodological issues
Schizophrenia is a rare disease, and although inspection of the rates in
Table 2 suggests substantial
neighbourhood-related variation, the statistical power was relatively low with
regard to the random neighbourhood effect. We can nevertheless be 94.5%
confident that the level-2 variance representing the random neighbourhood
effect was not merely a chance finding. The random neighbourhood effect
suggests that neighbourhoods are different, and therefore provides strong
support for the use of multi-level modelling in the analysis of data clustered
at the neighbourhood level.
We were not able to match all neighbourhood-level variables with their individual-level equivalents. This would have required very detailed population data, which are not available in European countries. However, age, gender and marital status are all very relevant with regard to the incidence of schizophrenia, and, as far as we are aware, this is the first study combining their individual-level and neighbourhood-level effects. Similarly, this is the first study to examine the effect of neighbourhood deprivation variables after adjustment for individual-level age, gender and marital status in the appropriate multi-level model.
Single individuals and individuals living in more deprived neighbourhoods may have a poorer prognosis, remain in the system of mental health services for a longer period of time, and therefore have a greater likelihood of eventually receiving a lifetime diagnosis of schizophrenia later in the course of their illness. However, in a previous study using these case-register data, we showed that associations with neighbourhood deprivation and signle marital status were similar in cases aged 16-93 years who received an early diagnosis of schizophrenia, as compared with those who received the diagnosis later in the course of their illness (Driessen et al, 1998a).
Some individuals may have moved from one neighbourhood to another in the prodromal stages. However, this could only have biased our findings if one assumes that: (a) single individuals in the prodromal stages would have drifted selectively to neighbourhoods with fewer single individuals; and (b) a sufficient amount of such prodromal drift had taken place to cause an interaction between individual-level marital status and its neighbourhood-level equivalent.
In general, individual marital status is not quite the individual-level equivalent of the population-level proportion of single individuals, as population-level single individuals include some single individuals who share the same household. This type of misclassification, however, would in fact have made our neighbourhood-level exposure more diluted, making it more difficult to find an effect, rather than leading to a spurious one.
Not all cases of schizophrenia are treated within the mental health system. However, the reported incidence of 22.3 per 100 000 is within the normal range and does not suggest that many cases were missed. The mean age at first contact was 35.5 years. In Camberwell between 1965 and 1991, the mean age at first contact in incident schizophrenia patients younger than 65 years was 31.3 years, and in Dumfries and Galloway it was 35.3 years between 1979 and 1998 (R. McCreadie, personal communication, 1999).
The effects of neighbourhood-level variables appeared small because effects were expressed as the risk associated with a 1% increase in the exposure. For example, the relative risk associated with a 1% increase in the proportion of single persons was 1.02. Given the fact that the Maastricht mean was 38.4%, with a standard deviation of 11.2% a difference of one standard deviation between neighbourhoods would mean a relative risk of 1.0211.2=1.25, or a 25% excess risk for schizophrenia; a difference of two standard deviations would result in an excess risk of 56%. Similarly, a difference of one standard deviation in the proportion of divorced persons would result in an excess risk of 27% (1.122.1).
Neighbourhood-level effects over and above individual-level
equivalents
The effects of individual-level younger age, male gender, being single,
being divorced and their interactions were all in the expected direction
(Riecher Rossler et al,
1992; Tien & Eaton,
1992; Jablensky & Cole,
1997). The risk-increasing effect of single marital status may be
an indicator of premorbid social impairment in individuals at risk of
developing schizophrenia (Van Os et
al, 1995), or a reflection of social isolation in the
interpersonal sense, as a risk factor for psychosis
(Wilkinson, 1975). We found
that the neighbourhood-level proportions of single and of divorced persons
also increased the risk, even after adjustment for their individual-level
equivalents. This suggests that there may be a true environmental
neighbourhood effect associated with the proportions of single and of divorced
persons. There are two possible caveats with regard to such an interpretation.
The first is that the proportion of single or of divorced persons in a
neighbourhood may in fact be proxies for some other relevant
neighbourhood-level indicator (Geronimus
& Bound, 1998). However, we were able to show that the effect
of the proportion of single and the proportion of divorced persons persisted
even after inclusion of a range of neighbourhood characteristics in the model
(unemployment, foreign-born, mutations, etc.), suggesting an effect truly
associated with these variables. For example, the proportion of single persons
could be a mere proxy for a greater proportion of young, mobile, and
unemployed persons. Had this been the case, however, we would have expected
the effect of the proportion of single persons to disappear after adjustment
for the proportion of young individuals, total mutations and the proportion of
unemployed, whereas the opposite occurred. The second caveat is that the
proportion of single or of divorced persons in a neighbourhood may be a proxy
for some individual-level variable other than marital status, for
which we failed to adjust (Morgenstern,
1998). Although this possibility cannot be discarded, recent work
has provided evidence for the existence of true neighbourhood-level effects on
a range of health-related outcomes (Lillie
Blanton et al, 1993;
Diez Roux et al, 1997;
Sampson et al, 1997;
Driessen et al,
1998b).
Interpretation
Faris & Dunham (1939)
suggested that a high proportion of persons living alone was indicative of
social isolation, which they in turn suggested had a causal
influence on the development of psychotic symptoms. Although the ecological
validity of the presumed relationship between the neighbourhood proportion of
single persons and neighbourhood social isolation remains to be established,
on the face of it, it has some validity. Hare
(1956b) found a
correlation of 0.63 between the age- and gender-standardised incidence of
schizophrenia and the proportion of single-person households in Bristol. He
suggested that neighbourhood differences in the incidence of schizophrenia
were not so much the result of differences in terms of population density or
material deprivation, but of differences in the proportion of people living
alone. The findings of the present investigation agree with this suggestion.
None of the neighbourhood-level indicators of deprivation had significant
effects after adjustment for individual-level age, gender and marital status,
or after adjustment for the neighbourhood-level proportion of single and
proportion of divorced persons. Hare
(1956b) concluded
that the risk associated with the proportion of single persons could have an
individual-level explanation (such as segregation of vulnerable (single)
individuals) or a macro-environmental explanation (such as high levels of
social isolation in areas with a high proportion of single households). In
addition, he stated that "these two hypotheses are by no means
incompatible and both factors may be operative". The current results
indeed suggest that the two factors may be interactive, and additionally
suggest that the proportion of divorced persons also independently contributes
to the increase in risk. The interaction was such that the individual-level
effect of single marital status was higher in area with fewer single
individuals. This parallels the findings reported for unemployment and
suicide, for example. Suicide is associated with unemployment at the
individual level. However, the risk for individual-level unemployment is
higher in areas where the proportion of unemployed is low
(Platt, 1986). It has been
suggested that the cognitive impact of unemployment is worse if most other
individuals in the environment are not unemployed, so that those without work
stand out as isolated exceptions
(Neeleman, 1997). Similarly,
single marital status may more easily give rise to perceived isolation if most
other individuals are living with a partner. Thus, premorbid vulnerability
resulting in single marital status may be more likely to progress to overt
disease in an environment with a higher perceived level of social isolation.
Uncovering such person-environment interactions remains essential for the
elucidation of causal mechanisms in schizophrenia.
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Clinical Implications and Limitations |
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LIMITATIONS
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Received for publication April 29, 1999. Revision received August 31, 1999. Accepted for publication September 1, 1999.