Obligaciones del Ente Territorial (Municipio)

GDS, short form, 6+ symptoms: 6-7 mild, 8-10 moderate, 11+ severe

MORTALITY

Cardiovascular, cancer, or non- cancer, non-cardiovascular

From death certificate & hospital records, if available. ICD-9 coding.

295 cancer events

Frequencies for all cancer, lung, breast, colon and other cancer

ECA, 2000 Episodes of Major Depression (MDE), and of Dysphoria (DE).

MDE 140 (4.5%), DE 669 (21.5%)

DIS, DSM-III diagnoses from interviews INCIDENCE

Primary cancer event, any

Self-report, and death certificates

203 events

Sites: lung, prostate, colon, skin, breast

Table 1.2iii Summary of cohort studies, part 3: analysis method, results, covariates and notes

study, year Analysis Methods Results (95% Cl) Adjustments Notes

WEHS, 1981 _t

Multiple risk logistic regression model Odds Ratio

Psychological depression associated with cancer death: 2.3, p < .001, adjusted

Age, smoking, alcohol, family history of cancer, occupational status

Good choice of age group; considered role of other risk factors for cancer

WEHS, 1987 ANOVA, ANCOVA, M-H t Cox’s regression Relative risk

Adjusted:

High D & cancer incidence 1.38 (1.00-1.89); & cancer mortality 1.96 (1.33-2.90)

Age, smoking, alcohol, family history of cancer, occupational status, body mass index, serum cholesterol

High D also associated with risk for non­ cancer causes of death

Alameda County, 1988 Age adjusted rates (direct to 1970 adult pop.)

Cox’s proportional hazards model

D V non-D

Mortality, M: 0.83, NS; W: 1.19, NS Incidence, M: 0.97, NS; W: 1.27, NS Also HR for sites, NS

Age

Notes in discussion HR not affected if adjust for smoking education income SAH alcohol & race; no measurement details

Found association between depression and all cause mortality (M: 1.43, p < 0.001; W: 1.43, p < 0.001) and noncancer mortality (M: 1.58, p < 0.001; W: 1.49, p <0.001)

WCCDS, 1988 T-test for differences in sub-groups, x* Cox’s regression, multivariate life table analyses

Relative risk

D V non-D, 1.4 (0.8-2.4) unadjusted

D V non-D, 1.5 (0.9-2.5) adjusted

Age, nulliparity, obesity, prior hysterectomy

Did not report oral contraceptive use

NHANES, 1989 Proportional hazards Relative risk

CES-D: Unadjusted, 1.0 (0.7-1.5) Adjusted, 0.9 (0.6-1.3) GWB-D: Unadjusted, 1.2 (1.0-1.5) Adjusted, 1.2 (0.9-1.5)

Age, sex, marital status, smoking, family history of cancer, hypertension, cholesterol

Disregarded other mortality findings

Washington County, 1990

Woolfs

Cox’s proportional hazards model Relative risk, age-adjusted

Overall, D: 1.09 (0.69-1.71) Interaction: if current smoker

& depressed: 2.6 (1.41-4.80) & non-depressed: 1.24 (0.79-1.95) with linear trend for 1) rates of smoking (p=.03)

Age, sex, length of follow-up (others not relevant at 10% level of significance)

Increased cancer in older participants; broad age distribution; smoking data for only1863 participants

MFHS, 1996 Contrasted mean depressiveness score for different levels of confounding variables (age, education, smoking, lung fn etc). Cox’s proportional hazards model Relative risk

Male lung cancer: Increased risk with psychosis 4.7 (2.02-10.94) age-adjusted

Increased risk with highest tertile depressiveness: 3.32 (1.53-7.2) age-adjusted; 2.89(1.18-7.08) adjusted NS between psychiatric diagnosis & all cancers combined

Age, smoking status, BMI, serum cholesterol, alcohol intake, antidepressant use, education, marital status, area, general health, leisure exercise, various lung functions

Checked prevalence of quitting (NSD between levels)

Interaction between depressiveness score and smoking status (most risk determinants for lung cancer were associated with depressiveness score)

Repeated analyses, taking out cases from 1®* 4 yrs, taking out those on antidepressants => results unaffected

EPESE, 1998 Contrasted CD and non-CD (x^, t-test, Mann-Whitney)

Proportional hazards model stratified by community

Hazard ratios

Higher crude rate of cancer in CD than Non-CD CD HR 1.88 (1.13-3.14)

D HR 1.02 (0.73-1.42)

Also HR for sites, NS

Age, sex, ethnicity, physical disability, number of hospital admissions in follow- up, smoking, alcohol intake

No significant interaction between CD & smoking (quite the reverse)

Results stand after adjusting for competing causes

OFPC, 1998 Contrasted with/without depression for baseline characteristics (x^, t-test) Proportional hazards models Hazard ratios

Depressive symptoms not associated with cancer 1.0 (0.6-1.7, p = .93)

but associated with cardiovascular mortality, 1.8 (1.2- 2.5, p = .003), and non-cancer, non-cardiovascular mortality 1.8 (1.2-2.7, p = .01)

Age, history of Ml, stroke, COPD, hypertension, diabetes, smoking, perceived health and cognitive function

ECA, 2000 Logistic regression (no time measurement) Relative risk

BASIC MODEL: MDE 1.0 (0.5-2.1) DE 1.3 (0.9-1.9) ADJUSTED; MDE 1.3 (0.6-2.8)

DE 1.3 (0.9-1.9)

HIGHLIGHTS: Women with MDE, elevated risk of breast cancer, 3.8 (1.0-14.2)

Age, gender, smoking, alcohol use (DSM dependence or abuse)

SES & ethnicity NS in unadjusted model, dropped from further analyses

Very skewed age population, broad and unbalanced; loss to follow-up, query under­ ascertainment of depression

1.2.3.2 M ethodological issues: the d ep en d en t variable

Breslow and Day (1987) point out that most cancers are rare diseases, and if looking at rare ones, 'cohort studies are unlikely to be of m uch value, unless the relative risk associated w ith the exposure under study is very large' (p. 21). One m ight argue that since m any of these cohort studies have considered all cancers together as the outcome, rather than considering cancers of specific sites, that this p oint is fairly addressed. However, although the neoplastic disease process is itself essentially similar, different sites of cancer have different aetiologies, and it is inadvisable to regard them together as a homogenous outcome. Repeatedly, commentators on this literature have

recommended that attention be paid to different sites and stages of disease (Bieliauskas & Garron 1982; Fox 1978; Perrin & Pierce 1959). But for m any cohorts there may be insufficient events of any one site for reliable analysis. This limitation is illustrated by the w ide confidence intervals for findings in tw o of the studies previously m entioned which analysed cancer incidence by site. Elevated breast cancer risk in depressed w om en in the ECA study was based on 25 cases, 3 of whom were categorised as having had a major depressive episode (Gallo et al. 2000). Sub-group analysis of smoking status w ith respect to lung cancer in the Mini-Finland H ealth Survey was based on 1 case in 143 non-smokers, com pared w ith 13 cases of cancer am ong 137 smokers (Knekt et al. 1996).

The next issue concerns the type of cancer outcome. Some studies have focused on cancer mortality exclusively (e.g. the first follow up of the WEHS, and the OFPC), bu t m ost have considered both incidence and m ortality (although it is not always clear that the two have been acceptably differentiated, as for instance in the NHANES study). Typically there is some order of delay between an individual being diagnosed with cancer and that event being officially recorded as a registration. The delay is often shorter w ith recording death and cause of death, so in the absence of the registration itself, it makes sense to count the death event along w ith the incidence to increase ascertainment of cancer events and therefore the num bers available for analysis. But there are m any uncertainties w ith respect to timing and cancer: w hen the disease actually begins; betw een onset and discovery; betw een discovery and diagnosis, and thus registration (Fox 1978). More importantly there are other variables at work influencing the course of the disease and the risk of death after diagnosis, such as

diagnostic skill, efficacy of and response to treatment, psychosocial factors and so forth. Whüe there is a well-developed literature devoted to psychological variables and their influence on prognosis or survival, w hen examining the relationship to

developm ent of cancer, incidence of the first cancer event w ould be the preferred dependent variable or outcome.

As in any properly conducted study, the most valid and reliable m easures should be used, but these cohort studies vary in the quality of their outcome data. Although medical records or inspection of death certificates corroborated deaths from cancer, incidence was not always similarly assessed. The WEHS and ECA studies relied on the self-reporting of cancer by participants, a strategy w hich introduces elements of recall bias, while ascertainment in the WCCDS depended up o n its participants rem aining in the medical care program m e in which they were enrolled (Hahn & Petitti 1988). Other studies relied upon local registers or hospital discharges, although hospital care m ay not be accessible to all persons in some societies. In fairness, not all health-care systems have speedy or reliable cancer registries w ith a high percentage of coverage (unlike that serving the MFHS, which had almost 100% coverage), and several studies strove to compensate by setting their findings in the context of national data, i.e. the Alameda County, ECA, EPESE and W ashington County studies.

The majority of studies at least describe the distribution of cancer cases by site, w ith the possible sole exception of the reporting of the NHANES follow up. The authors of the W ashington C ounty study grouped cancer events of sites related to smoking and those not, and analysed accordingly. Interestingly, they found that the effect of depressed mood on risk of smoking-related cancers was increased in the presence of smoking, though only significantly so in the heaviest smokers (RR = 18.47, 95% Cl 4.58-74.41; Linkins & Comstock 1990).

1.2.3.3 M ethodological issues: the in d ep en d en t variable

The cohort studies have by no means been united in their definition and m easurem ent of depression. It is readily apparent from Table 1.2ii that the nature of the independent variable, or categorisation of exposure, differs in cohort studies from the preceding research. Moreover, a wide variety of m easures have been employed, further

constraining comparison. These scales and inventories have included: the Center for Epidemiologic Studies Depression scale (CES-D; Radloff 1977); the Diagnostic

Interview Schedule (DIS; Robins et al. 1981); the Cheerful v Depressed sub-scale of the General Well-Being Schedule (GWB-D; D upuy 1977); the short form Geriatric

Depression Scale (GDS; Yesavage 1988); the H um an Populations Laboratory

Depression Inventory (HPLDI; Berkman & Breslow 1983); the M innesota Multiphasic Personality Inventory, in tw o different versions (MMPI; Dahlstrom, Welsh, &

Dahlstrom 1972); the 36-item General Health Q uestionnaire (GHQ; Goldberg 1972); and the Present State Examination (PSE; Wing, Cooper, & Sartorius 1974).

The expense and logistical dem ands of cohort studies em phasises use of self­ adm inistered questionnaires to measure psychosocial data rather than interviews, which are m ore time-consuming and costly. A lthough this step facilitated the use of standardised assessment instruments, it also tended to limit the conceptual basis underlying the work, and the application of instrum ents was by no m eans consistent. Some studies using the same key instrum ent w ould employ certain sub-scales in preference to others, reflecting an ongoing diversity of opinion regarding the salient characteristics of psychological exposure. For example, the WCCDS used a shorter version of the MMPI than the WEHS studies and u sed a standard cut-off of 70 on the depression sub-scale (D) to identify those w ith 'severe' depression (Hahn & Petitti 1988). The W estern Electric studies used absolute m agnitude of score on this sub-scale, as well as a dichotomous m easure of depression for those w ho scored higher on the D sub-scale than on all other sub-scales (high D). Furthermore, the later study also took repression (Welsh's R scale) and Cattell's 16pf scale into account (Persky et al. 1987). Similarly, tw o studies used the cut-off of 16 on the CES-D (NHANES and W ashington County studies), while the EPESE used a cut-off of 20, on the grounds that it w ould be more stringent (Penninx et al. 1998).

Not all of the m easures employed in these cohort studies were examining the same object and this m ay contribute to differences in findings by shifting the denom inator at risk, artificially altering the exposure. There are two possible approaches: to consider psychological wellbeing as existing along a continuum, or as a dichotomous variable, this latter approach being preferred in psychiatry. The DIS is designed to classify

psychiatrie diagnoses according to DSM-III criteria and the PSE is similarly oriented; either the participant has major depression, or not. These m ethods obtain m ore conservative estimates for prevalence of major depression than self-rating scales or inventories, typically less than 5% (see Table 1.2ii) and those studies that employed these m ethods did not find an association w ith cancer risk. Nevertheless, these studies also undertook to identify participants w ho m ight fall outside these strict criteria, by assessing dysphoric episodes (ECA) or by also using a screening instrum ent, which in the MFHS did dem onstrate a significant association w ith cancer incidence.

O ther studies used self-rating scales of depressive sym ptom s for the m ost part, although these m easures vary in their assessment of state or trait characteristics and m ost tend to require a clinician examination to confirm diagnostic status. By these means, the participant has psychological disturbance or depression to a lesser or greater degree. Some scales were derived from a personality approach to this aspect of m ental health, others were more influenced by the stress literature; certainly the

variety of m easures used does not engender an untroubled com parison of like w ith like (Temoshok & Heller 1984). The various questionnaire or inventory m easures obtain a range of 9 - 21% for prevalence of depressive or dysphoric sym ptom s (see Table 1.2ii). Interestingly, the chronically depressed m ade u p around 3% of the EPESE population, while those depressed only at baseline constituted 13%. This m ay reflect the improved sensitivity and specificity for major depression the authors cite for the higher CES-D cut-off (Penninx et al. 1998), as well as the advantage of increased sensitivity to chronic depression from the use of repeated measures.

Another m atter concerns the type of population for which a m easure was designed. It is not straightforw ard to compare a scale devised for geriatric populations such as the Geriatric Depression Scale (OFPC) w ith one designed for use in the general population; indeed this study estimated a prevalence of depression of 6.3% in their sample,

som ewhat less than the prevalence estim ated by the other questionnaire measures. Furtherm ore, clinical definitions of depressive disorder and psychological disturbance have changed over time in successive issues of the DSM criteria, as well as in the International Classification of Diseases, albeit in m inor ways (Horwitz & Scheid 1999).

1.2.3.4 M ethodological issues: study p o p u latio n & follow up

The next consideration that impacts on both rates of depression and cancer risk is the nature of the population from which the cohort members are draw n. It is m ore efficient to choose a study population which is likely to have experienced the risk factor or exposure to some degree, and will be at reasonable risk of the disease of interest, particularly w hen one does not expect the association betw een each to be of great magnitude. Age is a significant risk factor for cancer (lARC 1990) and an im portant factor in depression (Doris et al. 1999; Levi 1998; W ittchen et a l 1994). Perhaps m indful of this, the MFHS for example had a lower age limit of 30 years at baseline. The study sample should also be carefully screened for cancer to begin w ith and then diligently followed-up to ascertain cancer events as completely as possible.

If the age distribution of a cohort is skewed for example by including m any younger people at lesser risk of developing cancer over follow up, then this reduces the likelihood of any association being discovered. The target age range of parent studies restricted several cohorts in this regard. The w om en in the WCCDS were chosen as a function of their likely use of oral contraceptives, and so age in that cohort ranged from less than 25 to over 45 years (Hahn & Petitti 1988). A lthough follow u p of 12 to 14 years w ould have placed the older participants w ithin an appropriate age bracket in terms of risk, the younger participants w ould not have been at the same risk, therefore reducing the likely num ber of cancer events obtained. The 'nationally representative' samples of the NHANES ^ and Alameda County Study seem to cover the entire age range using a stratified probability sampling strategy. Nevertheless since the authors omit clear details of either age or gender in their description of the cohort (Kaplan & Reynolds 1988; Zonderm an et al. 1989), it is very difficult to tell w hat this m eans in practice.

Both the W ashington County and ECA cohort include members aged betw een 18 and 65 years, b u t the latter has a particularly skewed distribution, w ith the youngest group contributing m ost to the cohort size. On the other hand, some cohorts m ight be

considered to benefit from their target age bracket. Participants in the OFPC cohort

According to Kelsey, Thompson, & Evans (1986), the NHANES I study participants had an age range of 1 to 74 years, and NHANES II participants had an age range of 25 to 74 years.

were at risk of osteoporosis and were on average aged 72 years (Whooley & Browner 1998), and those in the EPESE had a m ean age of 79 years (Penninx et al. 1998). Cancer rates are typically higher in these age groups. However, these older samples may be considered likely to have lost individuals w ho w ould have developed or died from cancer or some other cause and so conclusions from these studies m ay be limited to an older population. In some respects the WEHS cohort was ideal for addressing the research question, w ith an age range of 40 to 55 years at baseline.

The majority of studies were sam pled from the general population, and included both men and women, w ith the exception of the WCCDS, OFPC and WEHS cohorts. The WCCDS enrolled women from a health care program m e in Califomia, which suggests the possibility of some selection bias and their findings w ould not necessarily

generalise to those who w ould have been unable to avail of such a programm e. Indeed the authors describe their study sample as mostly white, m arried and m oderately well educated (H ahn & Petitti 1988). The OFPC was also a female-only study sample. On the other hand, the WEHS comprised male workers from an electrical factory and only a small proportion were office workers or supervisors. Thus there is significant

contrast betw een the studies m terms of study base and sample characteristics. Further, it seems that for m ost studies, investigators seem to have assum ed that the effect of depression or depressive symptomatology on risk, if any, will be uniform irrespective of the gender, age, and socioeconomic make-up of the study sample, or indeed cancer site.

Participants in these cohort studies should be cancer free at the time of entry to the study. How ever, it can be very difficult to establish disease-free status at baseline. A lthough some studies relied upon self-reports of cancer rather than official records, neither the OFPC nor the NHANES authors indicated how they accounted for disease- free status and presum ably m ay not have excluded participants on this basis. The ECA investigators m ade an effort to exclude occult cancers by excluding those who rated

In document Contribución a la actualización del plan de gestión integral de residuos sólidos del Municipio de Santa Rosa de Cabal (página 99-103)