Changes in Depressive Symptoms and Subsequent Risk of Stroke in the Cardiovascular Health Study
Background and Purpose—Depression is associated with stroke, but the effects of changes in depressive symptoms on stroke risk are not well understood. This study examined whether depressive symptom changes across 2 successive annual assessments were associated with incident stroke the following year.
Methods—We used visit data from 4319 participants of the Cardiovascular Health Study who were stroke free at baseline to examine whether changes in depressive symptoms classified across 2 consecutive annual assessments predicted incident first stroke during the subsequent year. Depressive symptoms were assessed using the 10-item Center for Epidemiologic Studies Depression scale (high versus low at ≥10). Survival models were inverse probability weighted to adjust for demographics, health behaviors, medical conditions, past depressive symptoms, censoring, and survival.
Results—During follow-up, 334 strokes occurred. Relative to stable low scores of depressive symptoms, improved depression symptoms were associated with almost no excess risk of stroke (adjusted hazards ratio, 1.02; 95% confidence interval, 0.66–1.58). New-onset symptoms were nonsignificantly associated with elevated stroke risk (adjusted hazards ratio, 1.44; 95% confidence interval, 0.97–2.14), whereas persistently high depressive symptoms were associated with elevated adjusted hazard of all-cause stroke (adjusted hazards ratio, 1.65; 95% confidence interval, 1.06–2.56). No evidence for effect modification by race, age, or sex was found.
Conclusions—Persistently high symptoms of depression predicted elevated hazard of stroke. Participants with improved depressive symptoms had no elevation in stroke risk. Such findings suggest that strategies to reduce depressive symptoms may ameliorate stroke risk.
Major depressive disorder and depressive symptoms have been associated with an elevated risk of incident all-cause stroke.1,2 Debate continues on whether this association reflects confounding by undiagnosed vascular disease or causal effects of depression mediated by behavioral or physiological pathways.3,4 Many studies documenting the depression–stroke association have evaluated only baseline measures of depression or depressive symptoms and confounders.1 Such studies cannot assess whether effects are because of cumulative damage from years of exposure or shorter-term alterations imposed by transient depressive symptoms. Further, previous studies have rarely attempted to estimate the effects of changes in depressive symptom burden on stroke risk, partially because of methodological challenges related to assessment of time-varying confounders that may partially mediate the depression–stroke relationship.5,6 Estimating associations of stroke risk with a nuanced assessment of depressive symptoms can help elucidate timing of the effect of depression on stroke risk. This may help suggest potentially modifiable mechanisms and provide evidence on whether successful treatment of depressive symptoms is likely to reduce stroke risk. For example, if depression has acute effects on stroke risk, we might expect stroke risk to be heightened directly after onset of symptoms. Conversely, if depression alters long-term biological processes and behaviors that influence stroke cumulatively, we might expect impacts on stroke risk to emerge well after the occurrence of depressive symptoms and to remain elevated after remission.
We use data from the Cardiovascular Health Study, a large population-based cohort with annual assessment of depressive symptoms during 10 years and stroke ascertainment during follow-up, to examine whether changes in depressive symptoms over 2 successive points in time are associated with elevated stroke risk. We adjusted for time-varying confounders, survival, and study attrition by implementing stabilized inverse probability weights (SIPWs).
For this study, we defined the exposure window as 2 successive assessments of depressive symptoms that were 1 year apart. We tested the hypothesis that the risk of stroke would be greatest for those with elevated depressive symptoms at both time points. We hypothesized that stroke risk would be moderately increased for individuals with elevated depressive symptoms at only 1 time point.
Data were from the Cardiovascular Health Study, a large population-based cohort of men and women 65 and older from a sample of Medicare-eligible individuals in 4 Field Centers: Sacramento County, CA; Washington County, MD; Forsyth County, NC; and Pittsburgh, PA.7,8 In 1989 to 1990, the Field Centers recruited 5201 participants. In 1992 to 1993, an additional 687 blacks were recruited using similar methods. Participants had annual clinical evaluations until 1998 to 1999 with semiannual follow-up phone interviews continuing since enrollment. The overall refusal rate for annual contact was 5.1%.9 Institutional review committees at each study site approved the study, and all participants provided informed consent.9 Current analyses were deemed exempt by the Harvard School of Public Health Human Subjects Committee.
Stroke assessments examined at each study wave during 1991 to 1999 were linked to a depressive symptom exposure period consisting of the 2 consecutive annual assessments immediately preceding it. To ensure that measured confounders were temporally before the depression measures, we used measurements from the interview conducted 3 years before stroke assessment (Figure 1). From the 5888 individuals enrolled, individuals were excluded for the following reasons: having a history of stroke or missing an interview before the end of the first exposure period (277 individuals; 4.71%), missing depression scores during either of the first 2 exposure waves (779 individuals; 13.23%). and missing baseline covariates (513 individuals; 8.71%). The missing data patterns at enrollment is given in Table I in the online-only Data Supplement.
The remaining 4319 eligible participants contributed 24 940 observations with depressive symptom assessments. The overall median follow-up time from baseline was 8.96 years (range: 1.99–10.89), with a follow-up time of 9.69 years (range: 1.99–10.89) for participants enrolled in 1989 to 1990 and 6.92 years (range: 2.14–7.26) for participants enrolled in 1992 to 1993.
Strokes were identified by participant or proxy notifying Cardiovascular Health Study staff, by Cardiovascular Health Study staff asking about previous medical events during annual clinical visits and semiannual phone interviews, or during review of medical records.8 A committee comprising neurologists, a neuroradiologist, and a physician representative from the Coordinating Center adjudicated all potential stroke events and categorized them as ischemic, hemorrhagic, or uncertain based on medical records, brain magnetic resonance imaging, patient or proxy interviews, death certificates, and the Medicare hospitalization database as described previously.8,10 Participants classified with uncertain stroke type were excluded from analyses examining effects of depressive symptoms on specific stroke types. Transient ischemic attacks were not considered as outcome events in this analysis.
Depressive Symptom Exposure
Depressive symptoms were measured using a previously validated shortened version of the Center for Epidemiologic Studies Depression scale with 10 items focusing on mood, energy level, sleep, concentration, and irritability.11,12 Response options were on a 4-point scale with higher scores indicating greater frequency of symptoms during the past week (Table II in the online-only Data Supplement for item wording). A total depressive symptom score was calculated by summing scores across all items (reverse coding as appropriate) for a maximum score of 30. For each exposure year, a total score of at least 10 was classified as elevated depressive symptoms. This threshold has been previously validated in an older population against the 20-item Center for Epidemiologic Studies Depression scale (Cohen κ statistic=0.97).13
Two-year depressive symptom exposure was classified into 4 categories based on the 2 consecutive annual assessments immediately preceding the stroke assessment wave, which was shifted 1 year at a time to create a moving window for exposure categorization. These categories were (1) stable low: no elevated symptomatology at either the first or second exposure wave, (2) remitted: elevated depressive symptoms at the first exposure wave but not the second exposure wave, (3) new onset: elevated depressive symptoms at the second but not at the first exposure wave, and (4) stable high: persistently elevated depressive symptoms at both first and second exposure waves. The reference group for all analyses was participants with stable low depressive symptoms.
Cumulative previous exposure to depressive symptoms reflects an individual’s depressive symptom history before the 2-year exposure window. Specifically, it was calculated as the proportion of assessment waves an individual reported elevated depressive symptoms, including all waves from baseline up to 3 waves before stroke assessment (ie, through the wave before, but not including, the moving window described above). This cumulative exposure measure has a theoretical range from 0 (elevated depressive symptoms at none of the previous waves) to 1 (elevated depressive symptoms at all previous waves). The data structure for an individual enrolled in 1989 is illustrated in Figure 1.
We considered several potential confounders that remain stable over time and some confounders that change over time. Time-constant confounders included sex, age at enrollment (modeled as linear plus quadratic terms), self-reported race/ethnicity (non-Hispanic black, non-Hispanic white, Hispanic, or other), education (less than high school, high school, some college/vocational school, or college or more), enrollment phase (1989/1990 or 1992/1993), and clinic site. Time-updated confounders included income quartile and the following health behaviors and conditions: at least 1 alcoholic beverage a week (yes/no), current smoking (yes/no), blocks walked per week (continuous), total kilocalorie expended during physical activity (continuous), obesity (body mass index ≥30), heart diseases (yes/no), hypertension (yes/no), diabetes mellitus (yes/no), antidepressant use (yes/no), and cognitive function (continuous). Hypertension status was based on history of hypertension, medication use, and blood pressure measurements obtained during clinic visits. Diabetes mellitus status was based on the American Diabetes Association criteria.14 Total kilocalorie expenditure was estimated during the first, fourth, and eighth assessments waves using self-reported frequency and duration of 16 activities, including chores. Study personnel recorded antidepressant use during the 2 weeks before clinical visits by directly transcribing information from prescription bottles brought in by participants.7 Cognitive function was measured using 30-point Mini-Mental State Examinations the first assessment wave and the 100-point Modified Mini-Mental State Examination during the remainder of follow-up.15,16 Missing values on time-updated covariates were imputed by carrying forward the most recent previous report.
Methods of Analysis
We examined the distribution of depressive symptoms and time-updated confounders at each wave. In our main analysis, we used marginal structural models to concurrently estimate the hazard ratio for incident stroke associated with the depressive symptom categories in the most recent 2-wave exposure window (with stable low depressive symptoms as the reference group) and cumulative previous exposure to depressive symptoms. The model was estimated with discrete-time pooled logistic regressions to accommodate the fact that strokes were examined in 1-year intervals. Time-constant demographic variables were directly included in marginal structural models predicting hazard of stroke.
We controlled for time-updated confounders without conditioning on possible mediating pathways by applying SIPWs, which are calculated in such a way that after weighting depressive symptom patterns are unrelated to confounders. We estimated SIPWs to also adjust for differential study dropout and survival. The weights were trimmed to the 99th percentile to reduce the influence of outliers. Details on construction of time- and person-specific SIPWs are reported elsewhere.5,6
We evaluated whether effects of depressive symptom categories were similar for ischemic and hemorrhagic strokes. We assessed possible multiplicative effect modification by sex, baseline age (continuous), and race on all-cause stroke. Because of the small number of participants with self-reported Hispanic ethnicity or self-reported other race, these groups were excluded in analyses examining possible effect modification by race/ethnicity. We also conducted sensitivity analyses requiring a change of at least 6 points (20% of possible score) in depressive symptom levels for participants to be classified as having new-onset or remitted depressive symptoms. We examined the association of depressive symptom change with all-cause strokes, further adjusting for the Digit Symbol Substitution Test, a measure of cognitive function. We also compared models in which we dropped the adjustment for cumulative previous depressive symptom exposure. Last, we examined models controlled for only demographics and weighted for survival and participation (ie, not accounting for time-updated confounders). Participants were censored at first stroke, death, or the first time they did not respond to depressive symptom questions. Two-sided P values <0.05 and <0.10 were considered statistically significant and marginally significant, respectively. All analyses were conducted using SAS 9.3.
The 4319 participants comprising the analytic sample (Table 1) experienced 334 incident strokes (285 ischemic, 33 hemorrhagic, and 16 undetermined stroke type) during 1992 to 1999. Across each assessment wave, the most common exposure category was stable low depressive symptoms (Table III in the online-only Data Supplement).
Compared with those with stable low depressive symptoms, participants with stable high depressive symptoms had significantly elevated hazard of stroke (adjusted hazards ratio [aHR], 1.65; 95% confidence interval [CI], 1.06–2.56; Figure 2). New-onset symptoms were associated with elevated stroke risk (aHR, 1.44; 95% CI, 0.97–2.14), although the association was not statistically significant. Remitted symptoms were not associated with excess risk (aHR, 1.02; 95% CI, 0.66–1.58). Cumulative depressive symptom exposure was not associated with stroke incidence when accounting for depressive symptoms change at the most recent 2 waves (aHR, 1.15; 95% CI, 0.69–1.93). Results remained similar in models further adjusting for the Digit Symbol Substitution Test (Table IV in the online-only Data Supplement).
When considering stroke types separately (Figure 2), individuals with stable high depressive symptoms had a 60% excess hazard of ischemic stroke, but this effect estimate did not reach statistical significance (95% CI, 0.99–2.58). Depressive symptoms were not associated with hemorrhagic strokes, but effect estimates were imprecise because of the small number of hemorrhagic events (Figure I in the online-only Data Supplement).
In stratified analyses, stable high depressive symptoms were marginally associated with a 58% and 59% excess hazard of stroke among females and non-Hispanic whites, respectively (Table 2). Among these groups, elevated risk of stroke was also marginally associated with new onset of depressive symptoms. Although stable high depressive symptoms were associated with elevated effect estimates among the remaining subgroups, estimates were not statistically different from stable low depressive symptoms. Remitted depressive symptoms were not associated with elevated hazard of stroke among any subgroup. Cumulative previous exposure was marginally significantly associated with stroke risk among non-Hispanic blacks. The Wald χ2 P values for interaction terms examining possible differences in effect by sex (0.72), race (0.76), or baseline age (0.93) were not statistically significant.
In sensitivity analysis requiring at least a 6-point change in Center for Epidemiologic Studies Depression scale score for depressive symptoms to be classified as changed (either new onset or remitted), the effect estimate of stable high depressive symptoms was associated with excess elevated hazard of all-cause stroke at 77% and ischemic stroke at 69% (Table 3). New-onset depressive symptoms were associated with a 50% excess hazard of all-cause stroke. Neither remitted depressive symptoms nor cumulative previous exposure to depressive symptoms was associated with an elevated hazard of all-cause or ischemic stroke. The effect estimates of stable high depressive symptoms and new-onset depressive symptoms on hemorrhagic stroke were elevated but imprecise and not significant.
When we did not control for cumulative previous exposure of depressive symptoms in the models, the effect estimate for stable high depressive symptoms remained elevated (aHR, 1.77; 95% CI, 1.25–2.50), while the risk associated with new-onset depressive symptoms was elevated and marginally significant (aHR, 1.48; 95% CI, 1.00–2.18). Remitted depressive symptoms were not associated with stroke (aHR, 1.05, 95% CI, 0.69–1.59). Models with all depressive symptoms measures but controlling for only demographics and weighted for survival and participation had similar, slightly stronger effect estimates (Table V in the online-only Data Supplement).
In a large cohort of older Americans with adjudicated strokes, participants with persistently elevated depressive symptoms over 2 consecutive annual assessments had a 65% increased hazard of incident all-cause stroke during the following year compared with their counterparts with consistently low depressive symptoms after adjusting for known potential confounders, sample attrition, and depressive symptom history. Relative to those with consistently low depressive symptoms, participants with new onset of depressive symptoms during the exposure period had a 44% increase in hazard of incident all-cause stroke, though this was of marginal statistical significance. Remitted depressive symptoms and cumulative previous exposure to depressive symptoms were not significantly associated with an elevated hazard of stroke in any analysis. There was no evidence of effect modification by sex, race, or age.
Overall, these findings support the hypothesis that individuals with stable high depressive symptoms across 2 consecutive annual assessments have excess risk of stroke during the next year compared with individuals with stable low depressive symptoms. This is consistent with a previous study in the Health and Retirement Study using similar methodology, which found that individuals with persistent elevated depressive symptoms across 2 biennial interview waves had more than double the risk of stroke in the following 2-year window than individuals with stable low depressive symptoms.6 Contrary to our hypothesis and previous results in the Health and Retirement Study, we did not find evidence that individuals with remitted depressive symptoms remain at elevated stroke risk during the following year. Furthermore, unlike our results in the Health and Retirement Study, these findings suggest that onset of depressive symptoms may also be associated with elevated risk of stroke.
Pathways linking depressive symptoms and stroke remain unclear but may be categorized as operating in the shorter and longer term. Depression has been associated with stroke triggers such as infection17,18; some work has suggested that depression may downregulate the immune system thereby increasing risk of infection.19,20 The few mediation analyses that have been conducted in previous literature examined possible pathways that may operate over a longer time frame. One possibility is that depression leads to higher inflammation or greater metabolic dysregulation that increases risk of stroke.21–24 One study found evidence suggesting that a composite of major metabolic diseases (hypertension, hyperlipidemia, and diabetes mellitus) partially mediated the association between major depressive disorder and all-cause stroke.23 However, the various metabolic diseases were not examined separately, leaving the precise pathway unclear. Depressive symptoms or diagnosis have also been associated with physiological dysregulation that develops over a long period of time (eg, atherosclerosis,25 white matter hypertensities26) and may directly increase risk of stroke. However, depression and stroke could simply share risk factors with stroke. Future research should continue to examine possible mediators of the relationship between depressive symptoms and stroke functioning both in the short and long terms.
A strength of our study was the ability to consider dynamic aspects of depressive symptoms because of the annual assessment of depressive symptoms. Rigorous assessment of stroke incidence and type reduces the risk of measurement bias. With information on a wide range of possible confounders of the association between depressive symptoms and stroke, death, and study participation, we can also appropriately control for time-varying confounders and account for selective attrition through SIPWs.
About study limitations, our sample did not allow for separate consideration of individuals who identified as Hispanic or a race other than black or white. Unmeasured confounders or residual confounding by measured variables remains a concern; a particularly relevant set of unmeasured confounders relate to subclinical vascular diseases, which have been hypothesized to increase depression risk.3 This analysis included only 1 depressive symptom assessment per year and did not address how trajectories of symptoms over a longer period of time might impact stroke risk. This study also does not examine if effects on stroke risk differ by reason for depressive symptom remission (eg, psychotherapy, pharmacotherapy, or natural resolution).
In conclusion, our study provides valuable information to public health and medical professionals by examining a dynamic relationship between 2 prevalent conditions: elevated depressive symptoms and incidence of stroke. We contribute to the research by showing evidence that sustained elevated depressive symptoms are associated with increases in stroke risk and suggesting that risk of stroke may decrease if depressive symptoms remit. Further research should consider even more comprehensively the effects of depressive symptom trajectories, depressive symptom variability, the consequences of different forms of depression treatment, the association between depression and subtypes of ischemic stroke, the mechanisms that may underlie the associations between depression and stroke, and over what time frame interventions are most effective.
Sources of Funding
This research was supported by contracts HHSN268201200036C, HHSN268200800007C, N01HC55222, N01HC85079, N01HC85080, N01HC85081, N01HC85082, N01HC85083, N01HC85086, and grant U01HL080295 from the National Heart, Lung, and Blood Institute (NHLBI), with additional contribution from the National Institute of Neurological Disorders and Stroke (NINDS). Additional support was provided by R01AG023629 from the National Institute on Aging (NIA). A full list of principal Cardiovascular Health Study (CHS) investigators and institutions can be found at http://www.CHS-NHLBI.org. We gratefully acknowledge financial support from the following NIH institutes: NHLBI (F31HL112613 to Dr Gilsanz); the National Institute of Mental Health (RC4MH092707 to Dr Kubzansky and Dr Glymour); National Institute of Aging (R21AG03438502 to Dr Glymour); National Institute of Allergy and Infectious Diseases (AI113251 and AI104459 to Dr Tchetgen Tchetgen); and National Institute of Environmental Health Science (AI113251 to Dr Tchetgen Tchetgen) and financial support from the Initiative for Maximizing Student Development (R25GM055353 to Dr Gilsanz); the Yerby Postdoctoral Fellowship (to Dr Gilsanz); and the American Heart Association (10SDG2640243 to Dr Glymour and Dr Gilsanz). The content is solely the responsibility of the authors and does not represent the official views of the National Institutes of Health. The study funders had no role in the design, execution, or interpretation of these analyses.
The online-only Data Supplement is available with this article at http://stroke.ahajournals.org/lookup/suppl/doi:10.1161/STROKEAHA.116.013554/-/DC1.
- Received March 24, 2016.
- Revision received September 16, 2016.
- Accepted November 3, 2016.
- © 2016 American Heart Association, Inc.
- Dong JY,
- Zhang YH,
- Tong J,
- Qin LQ
- Brodtmann A
- Berkman LF,
- Kawachi I,
- Glymour MM
- Kubzansky LD,
- Winning A,
- Kawachi I
- Gilsanz P,
- Walter S,
- Tchetgen Tchetgen EJ,
- Patton KK,
- Moon JR,
- Capistrant BD,
- et al
- Steffens DC,
- Helms MJ,
- Krishnan KR,
- Burke GL
- Radloff LS
- Empana JP,
- Sykes DH,
- Luc G,
- Juhan-Vague I,
- Arveiler D,
- Ferrieres J,
- et al
- Mezuk B,
- Eaton WW,
- Albrecht S,
- Golden SH
- Herrmann LL,
- Le Masurier M,
- Ebmeier KP