In elderly general population sub-syndromal clinically significant levels of depressive symptoms are highly prevalent and associated with high co-morbidity and increased mortality risk. However changes in depressive symptoms over time and etiologic factors have been difficult to characterise notably due to methodological shortcomings. Our objective was to differentiate trajectories of depressive symptoms over 10 years in community-dwelling elderly men and women using statistical modelling methods which take into account intra-subject correlation and individual differences as well as to examine current and life-time risk factors associated with different trajectories.

Participants aged 65 and over were administered standardised questionnaires and underwent clinical examinations at baseline and after 2, 4, 7 and 10 years. Trajectories over time of the Center for Epidemiologic Studies Depression scores were modelled in 517 men and 736 women separately with latent class mixed models which include both a linear mixed model to describe latent classes of trajectories and a multinomial logistic model to characterise the latent trajectories according to baseline covariates (socio-demographic, lifestyle, clinical, genetic characteristics and stressful life events).

In both genders two different profiles of symptom changes were observed over the 10-year follow-up. For 9.1% of men and 25% of women a high depressive symptom trajectory was found with a trend toward worsening in men. The majority of the remaining men and women showed decreasing symptomatology over time, falling from clinically significant to very low levels of depressive symptoms. In large multivariate class membership models, mobility limitations [odds ratio (OR) = 4.5, 95% confidence interval (CI) 1.6–12.9 and OR = 4.9, 95% CI 2.3–10.7, in men and women respectively], ischemic pathologies (OR = 2.9, 95% CI 1.0–8.3 and OR = 3.1, 95% CI 1.0–9.9), and recent stressful events (OR = 4.5, 95% CI 1.1–18.5, OR = 3.2, 95% CI 1.6–6.2) were associated with a poor symptom course in both gender as well as diabetes in men (OR = 3.5, 95% CI 1.1–10.9) and childhood traumatic experiences in women (OR = 3.1, 95% CI 1.6–5.8).

This prospective study was able to differentiate patterns of chronic and remitting depressive symptoms in elderly people with distinct symptom courses and risk factors for men and women. These findings may inform prevention programmes designed to reduce the chronic course of depressive symptomatology.

N-3 polyunsaturated fatty acids (PUFAs) have been hypothesised to be protective for depression during pregnancy. However, there are few data and no consensus regarding this association. In this line, we aim to evaluate if the concentration of n-3 and n-6 PUFAs, and their ratio, are associated with depressive symptoms throughout pregnancy.

A prospective cohort of 172 Brazilian women was followed at 5–13th, 20–26th and 30–36th weeks of gestation. The presence of depressive symptoms was evaluated using the Edinburgh Postnatal Depression Scale (EPDS) at each pregnancy trimester. Depression was defined as an EPDS score ≥11. The concentrations of n-3 [α-linolenic acid; eicosapentaenoic acid (EPA); docosapentaenoic acid (DPA) and docosahexaenoic acid (DHA)] and n-6 PUFAs [linoleic acid; γ linolenic acid; eicosadienoic acid; eicosatrienoic acid; arachidonic acid; docosatetraenoic acid and docosapentaenoic acid] were expressed as absolute (μg/ml) values. The total n-6/n-3 ratio was calculated. Statistical analyses were performed using univariate and adjusted random intercept logistic model for each fatty acid (FA) considering the longitudinal nature of data. Covariates were selected as potential confounders based on their biological plausibility of having an association with the concentration of FA and depressive symptoms during pregnancy.

The prevalence of depressive symptoms was high in all pregnancy trimesters (1st = 33.7%; 2nd = 18.9%; 3rd = 17.4%). We did not find differences in means FA concentrations by depressive symptom classification, for each follow-up visit. The women presented a 5% decrease in the odds of having depressive symptoms for each one-week increase in the gestational age. As individual women progressed through pregnancy, higher concentrations of EPA (odds ratio (OR) = 0.92; 95% CI: 0.86–0.99), DHA (OR = 0.96; 95% CI: 0.93–0.99), DPA (OR = 0.87; 95% CI: 0.77–0.99) and total n-3 (OR = 0.98; 95% CI: 0.96–0.99) were associated with a lower odds of depressive symptoms, while higher total n-6/n-3 ratio were associated with greater odds of depressive symptoms (OR = 1.40; 95% CI: 1.09–1.79). We detected a decrease in the probability of depressive symptoms as concentrations of total n-3 FA, α-linolenic acid, DPA, and DHA increased. We also observed a sharper decline for women with initial greater chance of depressive symptoms compared with those with lower chance of having these symptoms.

We found a high prevalence of depressive symptoms in low-income Brazilian pregnant women and no significant associations between n-6 FA and depressive symptoms. Lower serum concentrations of DHA, EPA and DPA and a higher n-6/n-3 ratio at each pregnancy trimester were associated with higher odds of depressive symptoms throughout pregnancy.

The purpose of this paper is to describe variation, over the months of the year, in major depressive episode (MDE) prevalence. This is an important aspect of the epidemiological description of MDE, and one that has received surprisingly little attention in the literature. Evidence of seasonal variation in MDE prevalence has been weak and contradictory. Most studies have sought to estimate the prevalence of seasonal affective disorder using cut-points applied to scales assessing mood seasonality rather than MDE. This approach does not align with modern classification in which seasonal depression is a diagnostic subtype of major depression rather than a distinct category. Also, some studies may have lacked power to detect seasonal differences. We addressed these limitations by examining the month-specific occurrence of conventionally defined MDE and by pooling data from large epidemiological surveys to enhance precision in the analysis.

Data from two national survey programmes (the National Population Health Survey and the Canadian Community Health Survey) were used, providing ten datasets collected between 1996 and 2013, together including over 500,000. These studies assessed MDE using a short form version of the Composite International Diagnostic Interview (CIDI) for major depression, with one exception being a 2012 survey that used a non-abbreviated version of the CIDI. The proportion of episodes occurring in each month was evaluated using items from the diagnostic modules and statistical methods addressing complex design features of these trials. Overall month-specific pooled estimates and associated confidence intervals were estimated using random effects meta-analysis and a gradient was assessed using a meta-regression model that included a quadratic term.

There was considerable sampling variability when the month-specific proportions were estimated from individual survey datasets. However, across the various datasets, there was sufficient homogeneity to justify the pooling of these estimated proportions, producing large gains in precision. Seasonal variation was clearly evident in the pooled data. The highest proportion of episodes occurred in December, January and February and the lowest proportions occurred in June, July and August. The proportion of respondents reporting MDE in January was 70% higher than August, suggesting an association with implications for health policy. The pattern persisted with stratification for age group, sex and latitude.

Seasonal effects in MDE may have been obscured by small sample sizes in prior studies. In Canada, MDE has clear seasonal variation, yet this is not addressed in the planning of services. These results suggest that availability of depression treatment should be higher in the winter than the summer months.

Timely recognition and treatment of mental disorders with an onset in childhood and adolescence is paramount, as these are characterized by greater severity and longer persistence than disorders with an onset in adulthood. Studies examining time-to-treatment, also referred to as treatment delay, duration of untreated illness or latency to treatment, and defined as the time between disorder onset and initial treatment contact, are sparse and all based on adult samples. The aim of this study was to describe time-to-treatment and its correlates for any health care professional (any care) and secondary mental health care (secondary care), for a broad range of mental disorders, in adolescents.

Data from the Dutch community-based cohort study TRacking Adolescents’ Individual Lives Survey (TRAILS; N = 2230) were used. The Composite International Diagnostic Interview (CIDI) was administered to assess DSM-IV disorders, the age of onset, and the age of initial treatment contact with any health care professional in 1584 adolescents of 18–20 years old. In total 43% of the adolescents (n = 675) were diagnosed with a lifetime DSM-IV disorder. The age of initial treatment contact with secondary care was based on administrative records from 321 adolescents without a disorder onset before the age of 10. Descriptive statistics, cumulative lifetime probability plots, and Cox regression analyses were used analyze time-to-treatment.

The proportion of adolescents who reported lifetime treatment contact with any care varied from 15% for alcohol dependence to 82% for dysthymia. Regarding secondary care, proportions of lifetime treatment contact were lower for mood disorders and higher for substance dependence. Time-to-treatment for any care varied considerably between and within diagnostic classes. The probability of lifetime treatment contact for mood disorders was above 90%, whereas for other mental disorders this was substantially lower. An earlier age of onset predicted a longer, and the presence of a co-morbid mood disorder predicted a shorter time-to-treatment in general. Disorder severity predicted a shorter time-to-treatment for any care, but not for secondary care. Time-to-treatment for secondary care was shorter for adolescents from low and middle socioeconomic background than for adolescents from a high socioeconomic background.

Although the time-to-treatment was shorter for adolescents than for adults, it was still substantial, and the overall patterns were remarkably similar to those found in adults. Efforts to reduce time-to-treatment should therefore be aimed at children and adolescents. Future research should address mechanisms underlying time-to-treatment and its consequences for early-onset disorders in particular.

Studies on depression and mortality in nursing homes have shown inconclusive findings, and none has studied the role of detection. We sought to measure the association of depression with long-term all-cause mortality in institutionalised older people and evaluate a potential modification in the association by its detection status.

We selected a stratified cluster sample of 591 residents aged 75 years or older (mean age 84.5 years) living in residential and nursing homes of Madrid, Spain, who were free of severe cognitive impairment at the 1998–1999 baseline interview. Mortality was ascertained until age 105 years or September 2013 (median/maximum follow-up 4.8/15.2 years) through linkage to the Spanish National Death Index. Detected depression was defined at baseline as a physician's diagnosis or antidepressant use, undetected depression as significant depressive symptoms (score of 4 or higher on the ten-item version of the Geriatric Depression Scale) without documented diagnosis or treatment, and no depression as the absence of diagnosis, treatment, and symptoms. Constant and age-dependent hazard ratios for mortality comparing detected and undetected depression with no depression were estimated using Cox models, and absolute years of life gained and lost using Weibull models.

The baseline prevalences of detected and undetected depression were 25.9 and 18.8%, respectively. A total of 499 participants died during 3575 person-years of follow-up. In models adjusted for age, sex, type of facility, number of chronic conditions, and functional dependency, overall depression was not associated with long-term all-cause mortality (hazard ratio 0.87, 95% confidence interval (CI): 0.70–1.08). However, compared with no depression, detected depression showed lower mortality (hazard ratio 0.63, 95% CI: 0.46–0.86), while undetected depression registered higher, not statistically significant, mortality (hazard ratio 1.35, 95% CI: 0.98–1.86). The median life expectancy increased by 1.8 years (95% CI: −3.1 to 6.7 years) in residents with detected depression and decreased by 6.3 years (95% CI: 2.6–10.1 years) in those undetected. Results were more marked in women than men and they were robust to the exclusion of antidepressants from the definition of depression and also to the use of a stricter cut-off for the presence of depressive symptoms.

The long-term mortality risk associated with depression in nursing homes depends on its detection status, with better prognosis in residents with detected depression and worse in those undetected. The absolute impact of undetected depressive symptoms in terms of life expectancy can be prominent.

Provision and need for mental health services among military personnel are a major concern across nations. Two recent comparisons suggest higher rates of mental disorders in US and UK military personnel compared with civilians. However, these findings may not apply to other nations. Previous studies have focused on the overall effects of military service rather than the separate effects of military service and deployment. This study compared German military personnel with and without a history of deployment to sociodemographically matched civilians regarding prevalence and severity of 12-month DSM-IV mental disorders.

1439 deployed soldiers (DS), 779 never deployed soldiers (NS) and 1023 civilians were assessed with an adapted version of the Munich Composite International Diagnostic interview across the same timeframe. Data were weighted using propensity score methodology to assure comparability of the three samples.

Compared with adjusted civilians, the prevalence of any 12-month disorder was lower in NS (OR: 0.7, 95% CI: 0.5–0.99) and did not differ in DS. Significant differences between military personnel and civilians regarding prevalence and severity of individual diagnoses were only apparent for alcohol (DS: OR: 0.3, 95% CI: 0.1–0.6; NS: OR: 0.2, 95% CI: 0.1–0.6) and nicotine dependence (DS: OR: 0.5, 95% CI: 0.3–0.6; NS: OR: 0.5, 95% CI: 0.3–0.7) with lower values in both military samples. Elevated rates of panic/agoraphobia (OR: 2.7, 95% CI: 1.4–5.3) and posttraumatic stress disorder (OR: 3.2, 95% CI: 1.3–8.0) were observed in DS with high combat exposure compared with civilians.

Rates and severity of mental disorders in the German military are comparable with civilians for internalising and lower for substance use disorders. A higher risk of some disorders is reduced to DS with high combat exposure. This finding has implications for mental health service provision and the need for targeted interventions. Differences to previous US and UK studies that suggest an overall higher prevalence in military personnel might result from divergent study methods, deployment characteristics, military structures and occupational factors. Some of these factors might yield valuable targets to improve military mental health.