Judith L. Kinman

Defining the clinically important difference in pain outcome measures.

&NA; The purpose of this study was to determine the levels of change on standard pain scales that represent clinically important differences to patients. Data from analgesic studies are often difficult to interpret because the clinical importance of the results is not obvious. Differences between groups, as summarized by a change in mean values over time, can be difficult to apply to clinical care. Baseline scores vary widely and group mean differences could reflect large changes in a few patients, small changes in many patients, or any combination of these outcomes. Determination of the proportion of patients who have a clinically important improvement in their pain would provide a more interpretable result with direct clinical implications. However, determining a clinically important outcome requires information about the degree of change over time that is clinically important. Data from the titration phase of a multiple cross‐over randomized clinical trial of oral transmucosal fentanyl citrate (OTFC) for the treatment of cancer‐related breakthrough pain were re‐analyzed to examine the differences in pain scores between treatment episodes that did and did not yield adequate pain relief. The scales evaluated were absolute pain intensity difference (PID, 0–10 scale), percentage pain intensity difference (PID%, 0–100% scale), pain relief (PR, 0 (none), 1 (slight), 2 (moderate), 3 (lots), 4 (complete)), sum of the pain intensity difference (SPID over 60 min), percentage of maximum total pain relief (% Max TOTPAR over 60 min), and global medication performance (0 (poor), 1 (fair), 2 (good), 3 (very good), 4 (excellent)). Adequate relief was defined by the patients decision not to use another dose of opioid medication as a rescue, in addition to the study medication, to treat each painful episode. One hundred thirty OTFC naive patients contributed data on 1268 episodes of breakthrough pain. The scales that were converted to a percentage change yielded the best accuracy in predicting adequate relief, with balanced sensitivity and specificity. The best cut‐off point for both the % Max TOTPAR and the PID% was 33%. The best cut‐off points for the absolute scales were absolute pain intensity difference of 2, pain relief of 2 (moderate), and SPID of 2. The global medication performance of 2 (good) had excellent values as well. This study presents data‐derived cut‐off points for the changes in several pain scales, each reflecting the clinically important improvement for patients treating breakthrough cancer pain episodes with OTFC. Confirmation in other patient populations and different pain syndromes will be needed. The use of consistent clinically important cut‐off points as the primary outcome in future pain therapy clinical trials will enhance their validity, comparability, and clinical applicability.

Dental and Cardiac Risk Factors for Infective Endocarditis: A Population-Based, Case-Control Study

Infective endocarditis is uncommon but potentially fatal. Administration of antibiotic prophylaxis is conventional [1], but data supporting its effectiveness derive solely from anecdotal reports, studies of bacteremia after dental and other procedures, and animal models. The low incidence of disease [2] has made randomized human trials of antibiotic effectiveness impractical. Even if effective, antibiotic prophylaxis should be reserved for patients at increased risk, such as those with cardiac abnormalities who are undergoing dental procedures. However, controlled human studies of risk factors are lacking. Previous case series indicate that approximately 15% of patients with infective endocarditis caused by mouth organisms had undergone a recent dental procedure [3], but the comparable percentage from a general population is unknown. The single hospital-based casecontrol study did not find an elevated risk associated with dental therapy, except for a borderline increase with dental scaling [4]. We are unaware of controlled human studies that quantify the risk for infective endocarditis associated with cardiac valve abnormalities other than mitral valve prolapse. We therefore conducted a population-based casecontrol study to evaluate and quantify risk factors for infective endocarditis, especially those considered by the American Heart Association (AHA) to be indications for antibiotic prophylaxis [1]. Methods Participants From August 1988 to November 1990, we maintained surveillance for infective endocarditis in 54 hospitals of the Delaware Valley Case-Control Network, a population-based network of hospitals in the eight counties that constitute the Philadelphia Metropolitan Statistical Area and the county of New Castle, Delaware. Patients with a putative diagnosis of infective endocarditis were identified by hospital personnel and were reported to study nurses, who also actively sought cases. To assess the completeness of ascertainment, five high-yield hospitals and three low-yield hospitals were asked to list all patients discharged with a diagnosis of endocarditis over 3 months. These lists were compared with those obtained from our surveillance; charts were reviewed when differences were identified. We obtained informed consent from physicians and case-patients, then used structured forms to abstract medical records, including echocardiographic reports and hospital laboratory information on the infecting organism. We deleted information on purported risk factors for infective endocarditis and submitted these records for review by three of the authors, who are consultants in infectious diseases recognized for their expertise in infective endocarditis [5, 6]. These experts used their own global clinical judgment to classify potential cases as definite, probable, or possible cases or probable noncases. Agreement of two of the three reviewers was required to make the determination of a case or a noncase [6]. One control from the community was selected for each case-patient by using a modification of the Waksberg random-digit dialing method [7]. Controls and case-patients were matched for age (in 5-year strata), sex, and neighborhood of residence (by using area code, telephone exchange, and the first subsequent digit of the case-patients telephone number). We excluded from these analyses patients with infective endocarditis who were younger than 18 years of age, intravenous drug users, and patients who developed endocarditis in the hospital. This study received separate institutional review board approval at the University of Pennsylvania and all 54 participating hospitals. Data Collection Information was obtained from case-patients by conducting a structured telephone interview after hospital discharge. The date of hospital admission served as the study date for case-patients; for controls, the date of the telephone interview was used. Telephone interviewers collected information on demographic characteristics; diagnostic and therapeutic medical and dental procedures in the year before the study date; potential host risk factors, including preexisting cardiac lesions, preexisting local infection, risk factors for oral or dental disease, diabetes mellitus, immune deficiencies, family history of endocarditis, alcoholism, malignant conditions, and autoimmune disease; previous antibiotic use; and other recent illnesses. For each host risk factor, we requested the date of diagnosis, diagnostic method (for example, echocardiography for mitral valve prolapse), and the name of the practitioner who made the diagnosis. For each medical and dental procedure, we sought information about the procedure, the month and year in which it was performed, and the practitioner. We requested medical and dental records describing procedures and validating individual diagnoses. Study Variables Case-patients were considered infected with dental flora if the organism found was viridans streptococci; nutritionally variant streptococci; Actinobacillus species; Cardiobacterium hominis; anaerobes; -hemolytic streptococci (not group D); unspecified streptococci; or Haemophilus, Eikenella, Kingella, or Neisseria species. Because this study focused on indications for antibiotic prophylaxis, we examined host characteristics reported by patients as the primary risk factor variables, reflecting the information that would be available to a practitioner about to perform a procedure for which prophylaxis might be indicated. A variable called any valvular heart abnormality was defined as the presence of any of the following self-reported, preexisting conditions: mitral valve prolapse, congenital heart disease, history of rheumatic fever with heart involvement, prosthetic heart valve, previous episode of endocarditis, or other valvular heart disease. Dental visit information was obtained solely from dental records. Dental hygiene care was defined as preventive oral health services and therapeutic services, including coronal scaling and polishing. Consistent with AHA guidelines, invasive dental procedures were defined to include dental hygiene care, extractions, periodontal treatment (including scaling and root planing), endodontic treatment, mouth or gingival surgery, and treatment of tooth abscess. Noninvasive dental procedures were simple restorations, prosthetic and restorative dentistry, fluoride treatment, and other procedures (prosthetic services, including adjustments and suture removal). Unless otherwise specified, dental treatment refers to all dental treatment and is not limited to invasive procedures. Initial analyses focused on dental procedures performed at any time in the 3 months before the study date. Analyses were then narrowed to 2 months and 1 month before the study date. Time frames are approximate because the onset of infective endocarditis is often difficult to determine with certainty. We therefore chose the date of hospital admission as the study date, collected procedural data based on month rather than on a specific date, and calculated time frames under the assumption that procedures were performed on the 15th of the month. Statistical Analysis Frequencies and cross-tabulations between casecontrol status and potential risk factors were obtained. Conditional logistic regression was used to determine the independent effects of the various potential risk factors and the possibility of any interactions among factors [8]. Variables were included in multiple regression models if they were significant (P < 0.2) in unadjusted (matched) analyses (such as kidney disease and diabetes), if their inclusion had a substantial effect (>15% change) on coefficients for variables already in the model (such as insurance status) [9], or if they were strongly suspected a priori of being confounders (such as ethnicity). For analyses specific to participants with known cardiac valvular abnormalities, odds ratios and CIs were calculated from a model that included main effects for cardiac valvular abnormalities and dental treatment and the interaction between those variables. The odds ratio for various dental therapy variables among participants with cardiac valvular abnormalities was estimated by combining coefficients for the dental therapy variable and the interaction term. The CI for this combination of coefficients was estimated by using the appropriate variance and covariance terms [8]. With the interaction terms, participants with and those without valvular abnormalities were included in these analyses. Exact odds ratios and CIs, stratified on the matching variables, were calculated when data were too sparse for conditional logistic regression [10]. We used SAS statistical software (SAS Institute Inc, Cary, North Carolina) for data management and to obtain frequencies and cross-tabulations. We used EGRET (Epidemiological Graphics, Estimation and Testing software, version 0.25.1, Cytel Software Corp., Cambridge, Massachusetts) for conditional logistic regressions and exact analyses. All CIs are 95%, and all P values are two-sided. The sample size for the study was chosen so that by assuming an level of 0.05 (two-sided) and a power of 80%, we would be able to detect associations with an odds ratio of 2.0 or more for risk factors with a prevalence between 0.1 and 0.8. Results Participants We identified and recruited 416 potential case-patients (Figure 1). Our assessment process confirmed that more than 90% of true cases of infective endocarditis had been identified. The expert panel judged 379 patients to have definite, probable, or possible infective endocarditis; 37 (9%) were judged to be probable noncases and were excluded. Agreement among judgments was high, ranging from 92% to 96% [6]. Figure 1. Enrollment of case-patients. Of these 379 patients, 287 had community-acquired infective endocarditis not associated with intravenous drug use (248 on native valves and 39 on prosthetic valves), 27 had nosocomial infective endocarditis (18 on nativ

Seizure control and mortality in epilepsy.

Mortality rates are increased among people with epilepsy, and may be highest in those with uncontrolled seizures. Because epilepsy surgery eliminates seizures in some people, we used an epilepsy surgery population to examine how seizure control influences mortality. We tested the hypothesis that patients with complete seizure relief after surgery would have a lower mortality rate than those who had persistent seizures. Three hundred ninety‐three patients who had epilepsy surgery between January 1986 and January 1996 were followed after surgery to assess long‐term survival; 347 had focal resection or transection, and 46 had anterior or complete corpus callosotomy. A multivariate survival analysis was performed, contrasting survival in those who had seizure recurrence with survival of those who remained seizure free. Standardized mortality ratios and 95% confidence intervals were calculated. Overall, seizure‐free patients had a lower mortality rate than those with persistent seizures. This was true for the subset of patients with localized resection or multiple subpial transection. No patients died among 199 with no seizure recurrence, whereas of 194 patients with seizure recurrence, 11 died. Six of the deaths were sudden and unexplained. Most patients who died had a substantial reduction in postoperative seizure frequency. The standardized mortality ratio for patients with recurrent seizures was 4.69, and the risk of death in these patients was 1.37 in 100 person‐years, whereas among patients who became seizure free, there was no difference in mortality rate compared with the age‐ and sex‐matched population of the United States. Elimination of seizures after surgery reduces mortality rates in people with epilepsy to a level indistinguishable from that of the general population, whereas patients with recurrent seizures continue to suffer from high mortality rates. This suggests that uncontrolled seizures are a major risk factor for excess mortality in epilepsy. Achieving complete seizure control with epilepsy surgery in refractory patients reduces the risk of death, so the long‐term risk of continuing medical treatment appears to be higher than the risk of epilepsy surgery in suitable candidates. Ann Neurol 1999;46:45–50

Risk factors for gallbladder cancer. An international collaborative case–control study

Background. Gallbladder cancer has an unusual geographic and demographic distribution, suggesting many possible etiologies.

Risk Factors for Infective Endocarditis Oral Hygiene and Nondental Exposures

Background—The risks of infective endocarditis (IE) associated with various conditions and procedures are poorly defined. Methods and Results—This was a population-based case-control study conducted in 54 Philadelphia, Pa–area hospitals from 1988 to 1990. Community-acquired IE cases unassociated with intravenous drug use were compared with matched community residents. Subjects were interviewed for risk factors. Diagnoses were confirmed by expert review of medical record abstracts with risk factor data removed. Cases were more likely than controls to suffer from prior severe kidney disease (adjusted OR [95% CI]=16.9 [1.5 to 193], P =0.02) and diabetes mellitus (adjusted OR [95% CI]=2.7 [1.4 to 5.2], P =0.004). Cases infected with skin flora had received intravenous fluids more often (adjusted OR [95% CI]=6.7 [1.1 to 41], P =0.04) and had more often had a previous skin infection (adjusted OR [95% CI]=3.5 [0.7 to 17], P =0.11). No association was seen with pulmonary, gastrointestinal, cardiac, or genitourinary procedures or with surgery. Edentulous patients had a lower risk of IE from dental flora than patients who had teeth but did not floss. Daily flossing was associated with a borderline decreased IE risk. Conclusions—Within the limits of the available sample size, the data showed that IE patients differ from people without IE with regard to certain important risk factors but not regarding recent procedures.

INCIDENCE OF INFECTIVE ENDOCARDITIS IN THE DELAWARE VALLEY, 1988-1990

This population-based study aimed to determine the incidence of native, prosthetic, and bioprosthetic valve nosocomial infective endocarditis (IE), and IE associated with the use of injected drugs. Patients with IE during 27 months over the years 1988 to 1990, and residing in any of 6 counties in the Philadelphia metropolitan area were identified. An expert panel reviewed all patients to verify the diagnosis. Incidence rates were estimated after adjustment for failure to recruit and underreporting. Of 853 potential patients, 670 (79%) met the inclusion criteria. The overall incidence rate of IE was 11.6 cases/100,000 person-years (95% confidence interval [CI] 10.8 to 12.4). The rates for specific types of IE were: 4.45 (95% CI 3.97 to 4.94) for community-acquired native valve, 0.94 (95% CI 0.72 to 1.12) for prosthetic valve, 0.94 (95% CI 0.71 to 1.16) for nosocomial, and 5.34 (95% CI 4.80 to 5.87) for IE associated with use of injected drugs. Previous population studies found overall incidence rates of 1.7 to 4 cases/100,000 person-years, similar to our rate for community-acquired native valve IE. Type-specific rates have not been previously reported. The higher overall rate in this study is partly related to the high prevalence of injection drug use in our area.

Parenteral ketorolac: the risk for acute renal failure.

Ketorolac tromethamine is the first nonsteroidal anti-inflammatory drug (NSAID) to be approved in the United States for parenteral use as an analgesic. Clinical trials done before the drug was marketed showed that its efficacy was similar to that of moderate doses of parenteral opioids in patients having surgery [1]. Although ketorolac therapy has been discontinued less often than have meperidine hydrochloride and morphine therapy, ketorolac has been associated with the same adverse events that are seen with other NSAIDs; these adverse events include gastrointestinal events, rare allergic reactions, and liver dysfunction. Other NSAIDs have been associated with the renal syndromes of acute renal failure, interstitial nephritis, the nephrotic syndrome, hyponatremia, and hyperkalemia [2-14], but these were not reported in the premarketing clinical trials of ketorolac. Since ketorolac has been marketed, it has been widely used in clinical settings other than clinical trials. As have other NSAIDs [14-18], ketorolac has been associated with acute renal failure [19-27]. The appropriate role of ketorolac and all NSAIDs has consequently been questioned, especially for patients who are considered to be at high risk for acute renal failure [13, 14]. We did a large cohort study to evaluate the effects (including nephrotoxicity) of parenteral ketorolac in the postmarketing clinical setting. We previously reported on the risks for gastrointestinal bleeding and surgical-site bleeding associated with ketorolac [28]. We now compare the potential risk for acute renal failure after administration of ketorolac with the risk after administration of opioids among hospitalized patients. Methods Study Sample This retrospective cohort study was done using 35 community-based and tertiary care hospitals in the Philadelphia area. Data collection began on 18 November 1991 and ended on 31 August 1993. All patients who were identified from hospital pharmacy records as having received parenteral ketorolac during the data collection period were potentially eligible for inclusion in the group receiving ketorolac, regardless of whether they had concomitantly received opioids. The comparison group consisted of patients who received parenteral opioids (without ketorolac) and was matched to the ketorolac group by hospital, admitting service (any medical service compared with any surgical service), and date on which therapy was initiated. Use of ketorolac or opioids was verified by examining medication administration records. Patients who were receiving long-term dialysis were excluded. A course of ketorolac or opioids was defined as the time from administration of the first dose through the third day after administration of the final dose. If more than 3 days had elapsed between consecutive doses, a new course was defined as starting after the lapse. We collected data on all courses of ketorolac. Data on repeated courses of opioids were not abstracted because the purpose of the unexposed comparison group was to serve as a control group that had indications similar to those of the group receiving ketorolac and not to identify all adverse events that occurred in patients receiving opioids. If more than one course of opioids was available, we chose the course that had the initiation date closest to that of the matched patients course of ketorolac. Data were abstracted from the hospital charts of 9850 patients who had received 10 219 courses of ketorolac and of 10 145 patients who had received 10 145 courses of opioids. Only 326 of the 9850 patients receiving ketorolac (3.3%) received more than one course. Of these patients, 291 received two courses, 27 received three courses, and 8 received four courses. All analyses are presented by treatment course because each course represented a separate opportunity for an adverse outcome. However, separate analyses of each patients first course alone yielded nearly identical results. Data Collection Trained nurses used a computer-based data entry system to abstract data from hospital charts. The data collected included demographic characteristics, medical history, dose and duration of ketorolac or opioid therapy, occurrence of surgery, use of concomitant medication, laboratory data, and adverse events (regardless of whether the hospital staff or the abstracter thought that these events were caused by the drug). Definitions of Acute Renal Failure The principal definition of acute renal failure was a peak serum creatinine concentration that was 50% greater than the baseline value and 1) an absolute increase of at least 44.2 mol/L if the baseline concentration was less than 132.6 mol/L or 2) an absolute increase of at least 88.4 mol/L if the baseline concentration was 132.6 mol/L or greater. Patients for whom baseline serum creatinine values were not available did not meet the definition of acute renal failure even if their peak serum creatinine concentration was abnormally elevated. Our secondary definition required, in addition to laboratory evidence, a notation in the hospital chart that acute renal failure had occurred during the course of therapy with the analgesic drug. Unless otherwise stated, the results presented are those obtained using our principal definition. Statistical Analysis Data on demographic characteristics and medical history were compared between the two groups using the independent sample t-test for continuous variables and the chi-square statistic [29] for discrete variables. The proportion of patients in each group for whom data on serum creatinine concentration were included in the medical record was described. Both matched and unmatched analyses were done. Because point estimates and 95% CIs did not greatly differ between the two types of analysis, we report the results of the unmatched analyses [30]. We did a survival analysis using Cox proportional-hazards regression to explore the association of ketorolac administration with the rate of acute renal failure [30, 31]. Survival was defined as the interval from the initiation of analgesic drug therapy until either acute renal failure or the end of the course (3 days after the end of drug therapy), whichever occurred first. Unadjusted rate ratios comparing the rate of acute renal failure in patients receiving ketorolac with the rate in patients receiving only opioids were calculated using standard proportional hazards methods and are reported with 95% CIs [30, 31]. Multivariate proportional hazards models were fit, with simultaneous adjustment for the influence of potential confounding variables that were defined a priori. These variables included age; type of pain (acute or chronic) that served as the indication for analgesic administration; medical admission; admission to an intensive care or trauma unit; concomitant use of NSAIDs other than ketorolac, aminoglycoside antibiotics and other antibiotics, or angiotensin-converting enzyme inhibitors and other antihypertensive drugs; and a history of cancer, congestive heart failure, kidney disease, diabetes mellitus, hypertension, NSAID use, drug abuse, or cirrhosis. A time-dependent covariate that indicated the duration of analgesic therapy was incorporated into all models. We did a sensitivity analysis to assess the potential effect of the fact that a smaller proportion of patients in the ketorolac group had serum creatinine values measured during their treatment course. In this analysis, we recalculated the unadjusted relative risk for acute renal failure under the assumption that the risk for acute renal failure among study patients without measures of serum creatinine was the same as the risk among patients in the same group who did have measures recorded. This assumption is extreme because it assigns the risk for acute renal failure that was seen among patients who had laboratory data to patients who did not have laboratory data and thus probably had low morbidity. Because we were interested in the possible nephrotoxicity of ketorolac in patients who had a high risk for acute renal failure, we evaluated the interactions between the administration of ketorolac and coexisting conditions that may have predisposed patients to acute renal failure. These coexisting conditions include congestive heart failure; cirrhosis; a history of renal disease, diabetes mellitus, or hypertension; age older than 65 years; and heart failure, cirrhosis, or a history of renal disease [13]. We also explored potential interactions between ketorolac and the concomitant use of either aminoglycoside antibiotics or angiotensin-converting enzyme inhibitors. Finally, we explored the potential interaction of ketorolac with the presence of any condition known to predispose patients to acute renal failure or the concomitant administration of aminoglycoside antibiotics or angiotensin-converting enzyme inhibitors. Interactions were assessed on the basis of the statistical significance of the relevant product term in the multivariate models. Duration of analgesic therapy was defined as the number of days during which the analgesic drug was administered (in patients who did not have renal failure and in those whose event occurred after the last day of therapy) or the number of days from the onset of therapy until renal failure. The interaction between duration of therapy and choice of analgesic agent was analyzed in two ways. First, a set of proportional hazards models was fit; each model included patients who had received analgesic therapy for no longer than a specified duration (that is, those receiving therapy for as many as 2 days, as many as 3 days, and so forth). In any given model, patients who received an analgesic drug for longer than the specified duration for that model were included, but their follow-up was censored at that specified duration. For example, a patient who received analgesic therapy for 5 days was included in the model of as many as 4 days of analgesic therapy but was censored in that analysis a

Risk of venous thromboembolism from oral contraceptives containing gestodene and desogestrel versus levonorgestrel: a meta-analysis and formal sensitivity analysis.

Controversy exists regarding whether oral contraceptives (OCs) containing desogestrel and gestodene are associated with an increased risk of venous thromboembolism (VTE) versus OCs containing levonorgestrel. We were interested in synthesizing the available data, exploring explanations for mixed results, and characterizing the degree of uncontrolled confounding that could have produced a spurious association. We performed a meta-analysis and formal sensitivity analysis of studies that examined the relative risk of VTE for desogestrel and gestodene versus levonorgestrel. Twelve studies, all observational, were included. The summary relative risk (95% CI) was 1.7 (1.3-2.1; heterogeneity p = 0.09). If real, the incremental risk of VTE would be about 11 per 100,000 women per year. An association was present when accounting for duration of use and when restricted to the first year of use in new users. However, in the sensitivity analysis, the association abated in many, but not all, scenarios in which an unmeasured confounding factor increased the risk of VTE three to fivefold and in nearly all examined scenarios in which the factor increased the risk 10-fold. The summary relative risk of 1.7 does not appear to be caused by depletion of susceptibles, but is sensitive to a modest degree of unmeasured confounding. Whether such confounding occurred is unknown. However, given this sensitivity, this issue probably cannot be settled unequivocally with observational data. In the absence of a definitive answer, this apparent increased risk, together with its uncertainty and small magnitude and its important consequences, should be considered when selecting an OC for a given woman.

An Assessment of the Usefulness of the Duke Criteria for Diagnosing Active Infective Endocarditis

We evaluated the usefulness of the Duke criteria for diagnosing cases of active infective endocarditis (IE). Patients were identified prospectively over a 3-year period at 54 hospitals in the Philadelphia metropolitan area. Three of us independently reviewed abstracted hospital records and classified 410 patients as definite, probable, or possible cases of IE or as probable noncases. We then applied the Duke criteria to this sample to assess the degree of agreement between our diagnoses and the diagnoses based on these new criteria. Agreement was good to excellent, ranging from 72% to 90%, depending on the case definition used. The sensitivity of the Duke criteria was also good to excellent, varying from 71% to 99%, again depending on case definition used. Specificity was lower (0-89%). We conclude that use of the Duke criteria will result in little underdiagnosis of IE but that it may result in overdiagnosis of IE; therefore, these criteria should be applied prospectively to determine their clinical usefulness.

The effect of selective serotonin re‐uptake inhibitors on the risk of myocardial infarction in a cohort of patients with depression

AIM To evaluate whether selective serotonin re-uptake inhibitor (SSRI) exposure influences the risk of myocardial infarction (MI) in patients with depression. METHODS This study included 693 patients with MI (cases) and 2772 controls. Conditional logistic regression was used to calculate the odds ratio (OR). RESULTS SSRI exposure may be associated with a reduced MI risk (OR = 0.77, 95% CI 0.57, 1.03). However, reduced risk was only observed with longer term use (OR = 0.73, 95% CI 0.53, 1.00) and not with shorter term use (OR = 1.15, 95% CI: 0.65, 2.05). CONCLUSIONS Only longer term use of SSRIs was associated with reduced MI risk, suggesting that other mechanisms, besides an acute anti-platelet effect, may reduce MI risk.