Association between depressive symptoms and all- cause mortality in Chilean adult population: prospective results from the National Health Survey




Depression is a prevalent disorder with effects beyond mental health. An association with mortality has been reported, however, evidence is mixed and limited to a few high-income countries. This study aims to assess the association between depressive symptoms and all-cause mortality in the Chilean population.


This prospective study used data from the Chilean National Health Survey (ENS). Data from 3,151 and 3,749 participants from the 2003 and 2010 ENS respectively were linked to mortality register data. Cox survival analysis was performed using days from interview as time-to-event. The main exposure was depressive symptoms, measured with CIDI-SF (cut-off ≥ 5), and the outcome was all-cause mortality. The study period was limited to 8.5 years to allow for the same length of follow-up in both cohorts.


10% and 8.5% of participants from the 2003 and 2010 cohort died during the follow-up. Adjusting for age and sex, those with depressive symptoms were 1.50 (95% CI 1.11–2.02) and 1.51 (95% CI 1.11–2.04) times more likely to die than those without symptoms in the 2003 and 2010 cohort, respectively. In fully adjusted models, participants with depressive symptoms were 1.38 (95% CI 1.02–1.86) and 1.38 (95% CI 1.02–1.88) times more likely to die compared to those without symptoms in the 2003 and 2010 cohort.


Addressing mental health in the Chilean population could contribute not only to healthier but longer lives as well. Future research should collect data on more points in time to assess the effect of trajectories of depressive symptoms on mortality.


Depression is a common mental health condition with a range of effects on quality of life, disability, healthcare costs, mortality, among others [1]. This paper focuses on the association with mortality. A substantial number of studies have suggested those with depression tend to live shorter lives than those without depression [2]. Despite the relatively high number of studies assessing the association between depression and mortality, there is a large gap as evidence mostly comes from a few high-income Western countries [27]. Reviews focusing on populations from low-and-middle-income countries (LMIC) are limited to about 10 studies [8], while reviews encompassing studies mostly from high-income Western countries include more than 290 studies [2]. Moreover, there are methodological limitations in studies reported from countries with a similar income or from the same region as Chile [8].

A number of mechanisms have been posited to explain the association between depressive symptoms and mortality. These mechanisms can be divided into biological dysregulation and unhealthy behaviours [3]. It is relevant to note that these potential mechanisms are not mutually exclusive. Moreover, assessing the association between depression and mortality in different populations can contribute to elucidating if the posited mechanisms for this association are disease-specific —depression has been more often linked with cardiovascular disease [9]— or more generic. Similar results to studies conducted elsewhere would suggest that more generic mechanisms drive the association rather than disease-specific mechanisms. The estimated effect size of the association between depression and mortality varies widely between studies. Reports of positive associations ranged between relative risk (RR) 1.1 and 3.6. While many studies have reported a positive association; closer examination reveals most do not address relevant limitations. A meta-analysis pooling 238 studies reported an estimated RR of 1.64 (95% CI 1.56–1.76) [2], while a meta-analysis focusing exclusively on studies assessing the effect of clinically defined major depression reported an RR of 1.92 (95% CI 1.65–2.23) [3], suggesting severity may influence the association with all-cause mortality. This is because studies measuring depression through screening instruments tend to capture subclinical, less severe forms of depression [3].

Emerging evidence has suggested that this widely reported positive association is heavily determined by the quality of the evidence. In this context, quality of evidence is based on elements such as number of deaths, influence of small studies, excess of significance and heterogeneity. In the umbrella review by Machado and colleagues, this positive association was found to be highly influenced by reviews of lower quality [5]. Estimates considering the quality of the evidence consistently differed. There is evidence of smaller effect sizes in high-quality studies compared to studies with low quality. This indirect association between quality of evidence and size of the effect has been outlined by several reviews [2, 4, 5, 8]. Pooled estimates from high-quality studies are up to 36% smaller than pooled estimates from low-quality studies [2].

Most authors have identified inconsistencies in the association between depression and mortality [2, 4, 6, 8, 10] with several factors contributing to this, such as measurement of exposure, sample size, type of population, follow-up length, publication bias, outliers and adjustments [5, 11]. For type of population, there is evidence of a smaller effect for community samples compared to population with a specific disease [2]. As for follow-up, studies with a longer follow-up tend to yield smaller effect sizes than studies with a short follow-up [5]. There is also evidence of smaller effect sizes when removing outliers [2, 8]. A meta-analysis reported an attenuation of the estimated RR from 1.64 (95% CI 1.56–1.72) to 1.58 (95% CI 1.51–1.65) after removing outliers [2]. Most crucially, lack of adjustment for health conditions and health behaviours has been consistently identified as a relevant limitation [5].

Examination of studies from Latin America shows that none are nationally representative, most studies focused on older populations [1216] or in-patient samples [1721], and none were conducted in Chile. A vast majority of studies have sample sizes of less than 500 people [13, 1722]. A Brazilian study of older adults with 10 years of follow-up is among the few without some of these limitations [14]. Here, those with depressive symptomatology were 1.24 (95% CI 1.00-1.55) times more likely to die during the follow-up than those without symptomatology after adjustment for demographic, socioeconomic, functional limitations, cognitive features, lifestyle factors and chronic diseases. Other Latin-American research has reported both positive and no association between depression and mortality [14, 15, 19, 20]. Studies reporting no association were from in-patient samples [17, 18, 21, 22] rather than community samples. Based on the hypothesis of a higher risk of mortality among those with depressive symptoms compared to those without, the purpose of this paper is to extend the existing literature base and contribute to addressing the paucity of evidence from Latin America, using high-quality data from two Chilean nationally representative samples, with a specific measure of depressive symptoms, using a time-to-event analysis of all-cause mortality, and adjusting for a range of confounders.


Study population

The study sample came from the 2003 and 2010 Chilean National Health Surveys (ENS). These are random, geographically stratified, multi-stage cross-sectional surveys representative of Chilean adult population. The detailed study protocol of each ENS is described elsewhere [23, 24]. After exclusions due to missing exposure, outcome and covariates, the original survey samples —n = 3,619 and n = 5,293—, were reduced to n = 3,151 and n = 3,749 for the 2003 and 2010 ENS, respectively.

Exposure: depressive symptoms

Depressive symptoms were measured by trained interviewers at baseline using the Latin American version of the Composite International Disease Instrument Short Form (CIDI-SF) [25, 26], a screening instrument with a score ranging from 0–7, with a traditional cut-off indicating probable cases of depression of 5 [27]. Therefore, a binary exposure variable was created using this cut-off to indicate those with depressive symptoms.

Outcome: time to death

Mortality data were provided by the Chilean Ministry of Health, who linked the participant’s ENS identifier with administrative mortality data. This linkage data contained a variable for mortality, date and cause of death according to ICD-10 classification [28]. The time-to-event variable was estimated using the number of days from the interview date to the endpoint for the study (either the date of death or the end of follow-up). To have comparable results, follow-up was censored to 8.5 years in both ENS. As participants were followed over time, the surveys will be referred from now on as 2003 and 2010 cohorts.


The demographic covariates were age group (18–44, 45–64, 65 + years old), sex and marital status (living with partner, widowed/divorced, no partner). Socioeconomic status (SES) covariates were years of education (< 8 years, 8–12 years, 13 + years) and working status (employed/student, homemaker, retired, unemployed). Lifestyle variables included were physical activity (3 + times weekly, 1–2 times weekly, < 4 times per month, no physical activity) and smoking (smoker and non-smoker). Statistical models were also adjusted for the chronic diseases of type II diabetes and high-blood pressure (HBP), ascertained through physiological measurements and the use of prescribed medicine for these conditions. The reference group for the statistical models was youngest age group, female, living with a partner, 13 + years of education, employed/student, engaging in physical activity 3 + times weekly, non-smoker, and without HBP, diabetes and depressive symptoms.

Statistical analyses

Descriptive analysis

The sample characterization was displayed by cohort. Kaplan Meier plots representing the association between each category of exposures and mortality over time were analyzed. For the mortality data, the number and proportion of people who died and the survival function by cohort was inspected. Lastly, the survival function by cohort was examined. This function allows examining the probability of survival for participants over time.

Survival analysis

To assess a potential association between depressive symptoms and all-cause mortality, four models were fitted using Cox proportional hazard models [29], with participants with CIDI-SF < 5 as the reference group. The aim of the study and the nature of the outcome —time-to-event— make Cox survival analysis an appropriate method to use [30]. Firstly, a model adjusted by age and sex (model 1) was built, then it adjusted for marital status and SES variables (model 2), after that, health behavioural variables were added (model 3) and, lastly, adjustments for chronic diseases were made (model 4). Interactions between the exposure, age and gender were examined but no evidence of such interactions was found.

One of the key aspects of Cox proportional hazard model is the proportional hazard assumption. This was assessed by examining log-log plots, including a time interaction in the model, performing the non-zero slope of the Schoenfeld residuals test [31] and graphically examining scaled Schoenfeld residuals. The assumption was tested in the fully adjusted model.

Sensitivity analyses

Sensitivity analyses were carried out to assess potential biases in the study. To assess representativeness of the analytical sample, a comparison with excluded participants by survey was made using chi-square or chi-square for trend. The Cox model built implicitly assumes that depressive symptoms deteriorate participants’ health, which in turn increased the risk of mortality. To examine a potential reverse causality, we excluded those who died in the first 6 months. These models were adjusted only by age and sex to avoid statistical power issues. All analyses were done using RStudio version 1.4


Sample characterization

For the 2003 (n = 3,151) and 2010 (n = 3,749) cohorts, there were 25,505 and 30,578 person-years of observation respectively. About 16% of the sample in each cohort had depressive symptoms. There were some slight differences in the characteristics of the analytical samples for the two cohorts (Table 1). The main difference related to SES: the 2010 cohort had a lower proportion of unemployed people and a higher proportion of people with 13 + years of education compared to the 2003 cohort. The bivariate association between the different variables and mortality was assessed through Kaplan Meier plots. Supplementary Fig. 1 shows the Kaplan Meier plots for each variable. Results of these plots suggest that groups with poorer health or lower SES tend to have shorter survival over time.

Table 1

Sample characterization of the analytical sample by cohort.



2003 Cohort

n = 3,151

2010 Cohort

n = 3,749

n (%)


Deaths per 1,000 person-years

n (%)


Deaths per 1,000 person-years

Depressive symptoms

(CIDI-SF ≥ 5)

Non depressed

2,638 (83.7)



3,136 (83.7)




513 (16.3)



613 (16.4)





1,721 (54.6)



2,227 (59.4)




1,430 (45.4)



1,522 (40.6)



Age groups (years)


1,344 (42.7)



1,742 (46.5)




1,024 (32.5)



1,282 (34.2)




783 (24.9)



725 (19.3)



Marital status

With Partner

1,866 (59.2)



2,203 (58.8)




529 (16.8)



712 (19.0)



No Partner

756 (24.0)



834 (22.2)



Years of education


389 (12.4)



738 (19.7)




1,144 (36.3)



2,012 (53.7)



Less than 8

1,618 (51.4)



999 (26.7)



Working status

Employed + Student

1,396 (44.3)



2,046 (54.6)




935 (29.7)



912 (24.3)




433 (13.7)



578 (15.4)




387 (12.3)



213 (5.7)



Physical activity

3 + times weekly

262 (8.3)



274 (7.3)



1–2 times weekly

305 (9.7)



280 (7.5)



< 4 times per month

131 (4.2)



172 (4.6)



No sport

2,453 (77.9)



3,023 (80.6)





2,052 (65.1)



2,395 (63.9)




1,099 (34.9)



1,354 (36.1)



High-blood pressure (HBP)


2,068 (65.6)



2,694 (71.9)




1,083 (34.4)



1,055 (28.14)



Type II Diabetes


2,832 (89.9)



3,392 (90.48)




319 (10.1)



357 (9.52)



Mortality characterization

In total, 633 participants (315 and 318 in the 2003 and 2010 cohort, respectively) died during the 8.5-year follow-up. The all-cause data suggested a gradual divergence in mortality between cohorts in absolute terms. The proportion of people who died was very similar after a 2-year follow-up, but this difference increased over time. People from the 2003 cohort had higher mortality —of about 1.5%— compared to the 2010 cohort. (Table 2) Details of the proportion of deaths by covariates can be found in Supplementary Table 1.

Table 2

Number and percentage of all-cause mortality by follow-up time and cohort.





2003 (N = 3,151)

77 (2.44%)

174 (5.52%)

315 (10.00%)

2010 (N = 3,749)

68 (1.81%)

169 (4.51%)

318 (8.48%)


Survival function examination

Mortality by cohort was assessed with the survival function. Figure 1 compares the survival function of the 2003 and 2010 cohort within the 8.5-years follow-up period. The survival function seems to be lower for the 2003 cohort than for the 2010 cohort. After the follow-up period, there was a 90% and a 91.5% of survival probability, respectively. Nevertheless, the overlapping of confidence intervals between cohorts suggests that the survival functions are statistically indistinguishable. Based on the observed trend,it is feasible that with a longer follow-up, the divergence of the survival functions between cohorts could large enough for the confidence intervals to not overlap.

Cox Models for the association between depressive symptoms and all-cause mortality with an 8.5-year follow-up.

Table 3 shows the estimated hazard ratio (HR) for the association between depressive symptoms and all-cause mortality for all models. Model 1, adjusted by age and sex, suggests a positive association between depressive symptoms and all-cause mortality in both cohorts. Subsequent adjustments for marital status and SES variables (model 2) substantially attenuated the estimated effect size. Those with depressive symptoms had a 40% and 45% increased risk of mortality compared to those without depressive symptoms in the 2003 and 2010 cohorts, respectively. Further adjustment for physical activity and smoking (model 3) had little effect on the estimated HR. Lastly, adjustments for the presence of HBP and diabetes slightly attenuated the estimated effect (model 4). For the 2003 cohort, those with depressive symptoms were 1.38 (95% CI 1.02–1.86) times more likely to die at any point during the 8.5-year follow-up period than those without depressive symptoms. As for the 2010 cohort, the estimated effect was very consistent, with slight differences in the 95% CI. Participants with depressive symptoms from this cohort also had 38% (95% CI 1.02–1.88) increased risk of mortality over the analyzed period. In the fully adjusted model, the data suggest that the effect of depressive symptoms on mortality was very similar between the older (2003) and newer (2010) cohorts. This suggest no secular trends for the association between depressive symptoms and all-cause mortality among Chilean adult population.

Table 3

Association between depressive symptoms and all-cause mortality with an 8.5-year follow-up by cohort.


2003 cohort

(n = 3,151)

2010 cohort

(n = 3,749)

HR (95% CI)


HR (95% CI)


Model 1

1.50 (1.11, 2.02)


1.51 (1.11, 2.04)


Model 2

1.40 (1.04, 1.89)


1.45 (1.07, 1.96)


Model 3

1.40 (1.04, 1.90)


1.43 (1.06, 1.95)


Model 4

1.38 (1.02, 1.86)


1.38 (1.02, 1.88)


*Model 1: Model adjusted for sex and age. Model 2: Model adjusted for demographic variables, education and working status. Model 3: Model adjusted for demographic, SES variables, physical activity and smoking.

Model 4: Model adjusted for demographic, SES, lifestyle variables, high-blood pressure and diabetes.


Supplementary Table 2 contains the estimated HR for all covariates in the fully adjusted model for the 2003 and 2010 cohorts. In these models, being older, male, unemployed, retired, not engaging in physical activity, and having diabetes were associated with a higher risk of mortality compared to participants who were younger, female, employed, physically active or did not have diabetes in fully adjusted models. The variables with the largest effect on mortality were age groups, employment status and sex. Those in the oldest age group had about 25 times the risk of mortality compared to the youngest group, unemployed participants had up to 4.1 (95% CI 2.63–6.36) times the risk of mortality compared to employed participants and males had about a 50% increased risk of mortality compared to females; after adjusting for all other variables. Lastly, there was no evidence of a violation of the proportional hazard assumption of Cox models in the fully adjusted model.

Sensitivity analyses

The comparison of the sample excluded due to missing data and the analytical sample of the two cohorts showed good agreement between them. There is no evidence that excluded participants had systematically poorer health than the analytical sample. The sensitivity analysis excluding those who died in the first 6-months of the follow-up —to assess reverse causality— supported our findings. The estimated HR for the effect of depressive symptoms on all-cause mortality was somewhat attenuated in this analysis, yet it was robust to the aforementioned exclusion of participants.



The present study is the first to assess the association between depressive symptoms and all-cause mortality in the Chilean population. The mortality risk remained elevated for those with depressive symptoms after adjustment for demographic, SES, behavioural and chronic conditions in the 8.5-year period analysed. There was consistent evidence of a 38% increased risk of all-cause mortality for those with depressive symptoms compared to those without such symptoms in both cohorts.

Relationship with literature

The results from our work agree with the hypothesis of a higher risk of all-cause mortality for depressed individuals compared to those without. This hypothesis has been previously tested with both positive and null findings [2, 5]. Our fully adjusted estimates are lower than Cuijpers’ review estimate [2] but the latter was attenuated after removing outliers and adjusting for publication bias (RR 1.58 95% CI 1.51–1.65). Compared to literature from LMIC, the estimates from our work also tended to be smaller. After adjusting for outliers and publication bias, the review of the topic from LMIC reported a pooled RR of 1.60 (95% CI 1.37–1.86). Nevertheless, when only studies of high-quality are considered, the estimate decreased to 1.48 (95% CI 1.32–1.67).

Although pooled estimates from the literature tend to be larger in size than that estimated from our work, when critically assessed and adjusted for relevant factors, there is a much better agreement with our results. Most estimates from the literature are only adjusted by age and sex [4] and do not address the limitations described in the background, such as measurement of exposure, sample size, type of population, follow-up length, publication bias, outliers and adjustment for confounders [5, 11]. When only age and sex adjustments are considered, model 1, our estimate shows good agreement with the estimate from Cuijper’s review after adjustments for publication bias and outliers. The estimates of the association between depression and mortality in the literature are relatively similar when elements such as quality of studies and variables adjusted for are considered [2, 8]. This strengthens the generalisability of our results and suggests that, rather than specific mechanisms that influence all-cause mortality, more generic mechanisms drive the association between depression and mortality [2].

The hypothesized biological dysregulation as mechanism posits that depression causes changes at the biological level, such as changes in inflammatory responses [32], in the hypothalamic-pituitary-adrenal (HPA) axis [33, 34], cortisol levels [35] and noradrenaline levels [36]. These biological changes caused by depression increase the risk of mortality. On the other hand, unhealthy behaviours as mechanism range from lack of physical activity [37, 38], smoking [39, 40] to alcohol and drug abuse [41, 42]. In turn, these behaviours associated with depression, increase the risk of mortality. In this paper, the effect of certain unhealthy behaviours —smoking and physical activity— and chronic diseases strongly associated with unhealthy behaviours —diabetes and HBP— was accounted for. Thus, the estimated effect of depression on all-cause mortality in our model was independent of these factors. However, if unhealthy behaviours, such as lack of exercise, mediate the association between depression and mortality, adjusting for chronic conditions also associated with those unhealthy behaviours could potentially lead to overadjustment of our results [43]. The comparison of models with and without adjustments for unhealthy behaviours shows an inconsequential attenuation of the estimated effect, suggesting that, after accounting for depressive symptoms, demographic and SES variables, these unhealthy behaviours have little effect on all-cause mortality in our data. On the other hand, adjustment for chronic diseases strongly linked to unhealthy behaviours showed a small attenuation of the estimated effect between depression and mortality. Therefore, a potential overadjustment due to consideration of potential mediators in the model was considered to be negligible.

Despite some consistency in the positive association between depression and mortality reported in the literature [4, 44], Machado et al. emphasized that quality of studies could influence the size of this association [5]. This is true to some extent, as evidence from Machado’s and other reviews suggest a larger effect of studies of low-quality compared to high-quality studies. Nevertheless, in sensitivity analysis adjusted for comorbidities in community samples, the evidence from the umbrella review [5] still provides evidence for a positive association between depression and mortality, with a reported effect size very similar to our work (RR 1.38, 95% CI 1.29–1.47). It has also been posited that the increased mortality risk among depressed individuals is caused by other physical disorders that in turn cause depression [7]. However, both our estimate for the effect of depressive symptoms on mortality was robust to adjustment for chronic diseases and our sensitivity analysis excluding those with poorer health at baseline to address reverse causality do not support this idea.


There were several strengths in this study. This is the first Latin American study on the topic to be conducted on two nationally representative samples, enhancing the generalisability of its results. Moreover, both cohorts had a large sample size and there were enough deaths to ensure an appropriate power for this study. The use of two cohorts allowed to examine secular trends in the estimated effect. Our main exposure showed a remarkable stable effect on mortality. The decision to restrict the follow-up up to 8.5 years allowed us to make a reasonable comparison between cohorts, as there is some evidence of changes in the estimated effect according to the length of the follow-up period [2]. Attrition has been consistently identified as a limitation in prospective studies [45]. As the mortality data of ENS’ participants were linked using administrative data and not by contacting all participants at follow-up, attrition was not a limitation. Furthermore, this study used reliable data. Diabetes and HBP were ascertained through physiological measurements made by a nurse and not through self-reporting. One of the most consistent limitations highlighted in the literature was the lack of adjustment for relevant confounders. In this paper, a wide variety of confounders was included from different domains: demographic, socioeconomic, lifestyle variables and chronic conditions. The attenuation between model 1 and the fully adjusted model suggests that these adjustments —especially the SES variables— were relevant to assess the association between depression and mortality.


The results from this study need to be examined cautiously as there were some limitations. Some pertinent variables were unavailable or had high level of missingness and could not be used in the analysis. These variables relate to diet, other chronic conditions, and other mental health conditions. Therefore, residual confounding cannot be discarded. The instrument for measuring our exposure was not a structured psychiatric interview, instead, we relied on a screening instrument. This has the potential limitation of misclassification of subjects compared to the gold standard. However, there is evidence that the used instrument shows a high level of agreement with structured psychiatric interviews [27]. The measurement of depressive symptoms at baseline could reflect either an acute state at the moment of interview or the presence of a more chronic condition. The trajectories of depressive symptoms cannot be ascertained by our design, consequently, it was not possible to assess how much of the estimated effect of depressive symptoms on mortality is due to an acute or repeated depressive episodes. Public health policies could focus either on prevention —if the mere presence of depression increases risk of mortality— or on the management of depressive episodes if repeated episodes are more relevant for mortality. Overadjustment based on including potential mediators in the association between depressive symptoms and mortality could underestimate the true association. However, as previously discussed, the potential overadjustment due to the inclusion of potential mediators of unhealthy behaviours and chronic conditions in the models 3 and 4, respectively, was judged to be negligible based on the small attenuation of the estimates in our analyses The potential bias due to incompleteness of mortality records was addressed by examining the quality of the data in terms of cause and number of deaths. Overall, there was a relatively constant number of deaths over time and a similar proportion in causes of death by cohort. The small differences in these aspects were considered irrelevant. Lastly, there is some evidence of a difference in the effect of depression on mortality depending on the cause of mortality and gender [44]. Our analysis did not consider these aspects solely based on a lack of power to detect significant differences for a more specific outcome, such as cardiovascular mortality, or by gender.


This study showed a consistent effect of depressive symptoms on mortality among Chilean population over an 8.5-year period. The association was similar in size to what has been reported in high-income Western countries. Future research in this population could focus on assessing the effect of trajectories of depressive symptoms on mortality through a design that considers data collection at more points over time.


Competing interest:

The authors have no relevant financial or non-financial interests to disclose.

Supplementary information (SI)

Supplementary information is available online.

Conflicts of interest:

The authors declare no conflict of interest or relationships with industry associated with this article.

Ethics approval:

The investigation was conducted in compliance with all applicable ethical norms, including the principles of the Declaration of Helsinki and the ethical standards and its later amendments. The protocol for all three ENS received the approval of the ethical committee at Universidad Católica de Chile, the institution in charge of all three surveys. All participants gave their written consent for the study. Data on mortality and its linkage with ENS’ participant identifier was requested to the Ministry of Health through the Chilean Transparency Law platform. This research was based on publicly anonymized data.


EL is supported by the Chilean Comisión Nacional de Ciencia y Tecnología (CONICYT). The study was based on publicly available data, ENS, and administratively collected mortality data, therefore, no further funding was required.

Data transparency:

ENS’ data is publicly available on the website of the Chilean Department of Epidemiology, Data on mortality is not publicly available but it can be requested to the Ministry of Health through the Transparency Law platform.

Author contributions:

Substantial contributions to conception were made by EL, AP and HP. EL Contributed to the acquisition and cleaning of the data. All authors contributed to the analysis and interpretation of data. EL drafter the paper. All three authors critically revised the draft and approved the final version and agree to be accountable for all aspects of the work in ensuring that questions related to the accuracy or integrity of any part of the work are appropriately investigated and resolved.


  1. World Health Organization (2010) Mental disorders: equity and social determinants. In: Blas E, Sivasankara K (eds) Equity, social determinants and public health programmes. Geneva, p 303
  2. Cuijpers P, Vogelzangs N, Twisk J, et al (2014) Comprehensive meta-analysis of excess mortality in depression in the general community versus patients with specific illnesses. Am J Psychiatry 171:453–462.
  3. Baxter AJ, Page A, Whiteford HA (2011) Factors Influencing Risk of Premature Mortality in Community Cases of Depression: A Meta-Analytic Review. Epidemiol Res Int 2011:832945.
  4. Wulsin LR, Vaillant GE, Wells VE (1999) A systematic review of the mortality of depression. Psychosom Med 61:6–17.
  5. Machado MO, Veronese N, Sanches M, et al (2018) The association of depression and all-cause and cause-specific mortality: An umbrella review of systematic reviews and meta-analyses. BMC Med 16:.
  6. Schulz R, Drayer RA, Rollman BL (2002) Depression as a risk factor for non-suicide mortality in the elderly. In: Biological Psychiatry. Elsevier, pp 205–225
  7. Cuijpers P, Smit F (2002) Excess mortality in depression: A meta-analysis of community studies. J Affect Disord 72:227–236.
  8. Brandão DJ, Fontenelle LF, da Silva SA, et al (2019) Depression and excess mortality in the elderly living in low- and middle-income countries: Systematic review and meta-analysis. Int. J. Geriatr. Psychiatry 34:22–30
  9. Penninx BWJH (2017) Depression and cardiovascular disease: Epidemiological evidence on their linking mechanisms. Neurosci. Biobehav. Rev. 74:277–286
  10. Cuijpers P, Schoevers RA (2004) Increased mortality in depressive disorders: A review. Curr. Psychiatry Rep. 6:430–437
  11. Miloyan B, Fried E (2017) A reassessment of the relationship between depression and all-cause mortality in 3,604,005 participants from 293 studies. World Psychiatry 16:219–220
  12. Lima MTDR, Silva RDS, Ramos LR (2009) Depressive symptomatology and its associated factors in an urban cohort of elderly. J Bras Psiquiatr 58:1–7.
  13. Maciel ÁCC, Guerra RO (2008) Limitação funcional e sobrevida em idosos de comunidade. Rev Assoc Med Bras 54:347–352.
  14. Diniz BS, Reynolds CF, Butters MA, et al (2014) The effect of gender, age, and symptom severity in late-life depression on the risk of all-cause mortality: The Bambuí cohort study of aging. Depress Anxiety 31:787–795.
  15. Piña-Escudero SD, Navarrete-Reyes AP, Ávila-Funes JA (2011) Depressive symptoms increase the risk of mortality in older Mexican community-dwelling adults. J. Am. Geriatr. Soc. 59:2171–2172
  16. Ferreira TCBR, Coimbra AMV, Falsarella GR, et al (2016) Mortality in Brazilian community-dwelling older adults: 7 years of follow up in primary care. Geriatr Gerontol Int 16:804–809.
  17. Von Ammon Cavanaugh S, Furlanetto LM, Creech SD, Powell LH (2001) Medical illness, past depression, and present depression: A predictive triad for in-hospital mortality. Am J Psychiatry 158:43–48.
  18. Santos PR (2012) Evaluation of objective and subjective indicators of death in a period of one year in a sample of prevalent patients under regular hemodialysis. BMC Res Notes 5:24.
  19. De Guevara MSL, Schauffele SI, Nicola-Siri LC, et al (2004) Worsening of depressive symptoms 6 months after an acute coronary event in older adults is associated with impairment of cardiac autonomic function. J Affect Disord 80:257–262.
  20. Arrieta Ó, Angulo LP, Núñez-Valencia C, et al (2013) Association of Depression and Anxiety on Quality of Life, Treatment Adherence, and Prognosis in Patients with Advanced Non-small Cell Lung Cancer. Ann Surg Oncol 20:1941–1948.
  21. Zimmermann PR, Camey SA, Mari JDJ (2006) A cohort study to assess the impact of depression on patients with kidney disease. Int J Psychiatry Med 36:457–468.
  22. Diefenthaeler EC, Wagner MB, Poli-de-Figueiredo CE, et al (2008) Is depression a risk factor for mortality in chronic hemodialysis patients? Rev Bras Psiquiatr 30:99–103.
  23. Departamento de Salud Pública PUC (2003) Resultados I Encuesta de Salud, Chile 2003. Santiago
  24. Pontificia Universidad Católica (2010) Informe Final ENS 2009–2010. Santiago
  25. Kessler RC, Andrews G, Mroczek D, et al (1998) The World Health Organization Composite International Diagnostic Interview short-form (CIDI-SF). Int J Methods Psychiatr Res 7:171–185.
  26. Alderete E, Armando Vega W, Kolody B, Aguilar-Gaxiola S (2000) Lifetime Prevalence of and Risk Factors for Psychiatric Disorders Among Mexican Migrant Farmworkers in California. Am J Public Health 90:
  27. Kessler RC, Andrews G, Mroczek D, et al (2001) Scoring the World Health Organization´s Composite International Diagnostic Interview Short Form
  28. World Health Organization (2016) International Statistical Classification of Diseases and Related Health Problems (ICD) 10th Revision. Accessed 21 May 2019
  29. Hosmer WD, Lemeshow S, May S (2014) Applied Survival Analysis: Regression Modeling of Time-To-Event Data, 2nd ed. John Wiley & Sons
  30. Clark TG, Bradburn MJ, Love SB, Altman DG (2003) Survival Analysis Part I: Basic concepts and first analyses. Br J Cancer 89:232.
  31. Grambsch PM, Therneau TM (1994) Proportional Hazards Tests and Diagnostics Based on Weighted Residuals. Biometrika 81:515.
  32. Kiecolt-Glaser JK, Glaser R (2002) Depression and immune function: Central pathways to morbidity and mortality. J Psychosom Res 53:873–876.
  33. Holsboer F (2000) The corticosteroid receptor hypothesis of depression. Neuropsychopharmacology 23:477–501.
  34. Pariante CM (2003) Depression, stress and the adrenal axis. J Neuroendocrinol 15:811–812.
  35. Musselman DL, Nemeroff CB (1996) Depression and Endocrine Disorders: Focus on the Thyroid and Adrenal System. Br J Psychiatry 168:123–128.
  36. Ridker PM, Buring JE, Shih J, et al (1998) Prospective study of C-reactive protein and the risk of future cardiovascular events among apparently healthy women. Circulation 98:731–733.
  37. de Wit L, van Straten A, Lamers F, et al (2011) Are sedentary television watching and computer use behaviors associated with anxiety and depressive disorders? Psychiatry Res 186:239–243.
  38. Penninx BWJH, Geerlings SW, Deeg DJH, et al (1999) Minor and major depression and the risk of death in older persons. Arch Gen Psychiatry 56:889–895.
  39. Dierker LC, Avenevoli S, Stolar M, Merikangas KR (2002) Smoking and depression: An examination of mechanisms of comorbidity. Am J Psychiatry 159:947–953.
  40. Breslau N, Peterson EL, Schultz LR, et al (1998) Major depression and stages of smoking. A longitudinal investigation. Arch Gen Psychiatry 55:161–166.
  41. Roeloffs CA, Fink A, Unützer J, et al (2001) Problematic substance use, depressive symptoms, and gender in primary care. Psychiatr Serv 52:1251–1253.
  42. Degenhardt L, Hall W, Lynskey M (2003) Exploring the association between cannabis use and depression. Addiction 98:1493–1504.
  43. Kozela M, Bobak M, Besala A, et al (2016) The association of depressive symptoms with cardiovascular and all-cause mortality in Central and Eastern Europe: Prospective results of the HAPIEE study. Eur J Prev Cardiol 23:1839–1847.
  44. Cuijpers P, Vogelzangs N, Twisk J, et al (2014) Is excess mortality higher in depressed men than in depressed women? A meta-analytic comparison. J. Affect. Disord. 161:47–54
  45. Gustavson K, von Soest T, Karevold E, Røysamb E (2012) Attrition and generalizability in longitudinal studies: findings from a 15-year population-based study and a Monte Carlo simulation study. BMC Public Health 12:918.