Data are shown for first 4 years of follow-up, the period during which all deaths in the methadone group occurred. The cumulative incidence for sustained-release (SR) morphine, the reference group, was unadjusted; the cumulative incidence for methadone was adjusted by multiplying the unadjusted incidence by the ratio of the adjusted (calculated for the entire period of study follow-up) to unadjusted hazard ratios (HRs) (adjusted HR, 1.46; 95% CI, 1.17-1.83).
Adjusted methadone vs sustained-release (SR) morphine hazard ratio (HR). Limit lines indicate 95% CIs.
eMethods. Additional Details
eTable 1. Cohort Inclusion/Exclusion Criteria
eTable 2. Cause of Death Codes Consistent With Opioid Overdose Death
eTable 3. Distribution of Variables in the Propensity Score According to Study Opioid Use
eTable 4. Distribution of Propensity Score According to Study Opioid
eTable 5. Propensity Score–Matched Cohort Characteristics
eTable 6. Risk of Opioid Overdose or Sudden Cardiac Death During Current Use of Morphine Sustained Release and Methadone, Using Mutually Exclusive Definitions to Classify Deaths
eFigure 1. Study follow-up for 4 Hypothetical Cohort Members
eFigure 2. Study Death Adjudication
eFigure 3. Annual Incidence of Deaths According to the Global Mortality Risk Score Quantile
eFigure 4. Study Death Adjudication With Hazard Ratios
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Ray WA, Chung CP, Murray KT, Cooper WO, Hall K, Stein CM. Out-of-Hospital Mortality Among Patients Receiving Methadone for Noncancer Pain. JAMA Intern Med. 2015;175(3):420–427. doi:10.1001/jamainternmed.2014.6294
Copyright 2015 American Medical Association. All Rights Reserved. Applicable FARS/DFARS Restrictions Apply to Government Use.
Growing methadone use in pain management has raised concerns regarding its safety relative to other long-acting opioids. Methadone hydrochloride may increase the risk for lethal respiratory depression related to accidental overdose and life-threatening ventricular arrhythmias.
To compare the risk of out-of-hospital death in patients receiving methadone for noncancer pain with that in comparable patients receiving sustained-release (SR) morphine sulfate.
Design, Setting, and Participants
A retrospective cohort study was conducted using Tennessee Medicaid records from 1997 through 2009. The cohort included patients receiving morphine SR or methadone who were aged 30 to 74 years, did not have cancer or another life-threatening illness, and were not in a hospital or nursing home. At cohort entry, 32 742 and 6014 patients had filled a prescription for morphine SR or methadone, respectively. The patients’ median age was 48 years, 57.9% were female, and comparable proportions had received cardiovascular, psychotropic, and other musculoskeletal medications. Nearly 90% of the patients received the opioid for back pain or other musculoskeletal pain. The median doses prescribed for morphine SR and methadone were 90 mg/d and 40 mg/d, respectively.
Main Outcomes and Measures
The primary study end point was out-of-hospital mortality, given that opioid-related deaths typically occur outside the hospital.
There were 477 deaths during 28 699 person-years of follow-up (ie, 166 deaths per 10 000 person-years). After control for study covariates, patients receiving methadone had a 46% increased risk of death during the follow-up period, with an adjusted hazard ratio (HR) of 1.46 (95% CI, 1.17-1.83; P < .001), resulting in 72 (95% CI, 27-130) excess deaths per 10 000 person-years of follow-up. Methadone doses of 20 mg/d or less, the lowest dose quartile, were associated with an increased risk of death (HR, 1.59; 95% CI, 1.01-2.51, P = .046) relative to a comparable dose of morphine SR (<60 mg/d).
Conclusions and Relevance
The increased risk of death observed for patients receiving methadone in this retrospective cohort study, even for low doses, supports recommendations that it should not be a drug of first choice for noncancer pain.
Methadone hydrochloride, a μ-opioid agonist long used as evidence-based treatment for opioid dependence,1 has been increasingly prescribed for chronic pain. In 2009, 4.4 million methadone prescriptions in the United States were for treatment of pain, accounting for 9% of prescribed opioid analgesics on a dose-adjusted basis.1 Methadone's efficacy is comparable to that of other long-acting opioids2; however, its primary advantages as an analgesic are a long elimination half-life3 and low cost.1
There are major concerns regarding methadone's relative safety. The risk for accidental overdose and lethal respiratory depression may be greater than that for other long-acting opioids. Because the duration of methadone's respiratory depressant effects is longer than that of its analgesic effects,4,5 inadvertent intoxication can occur as the dose is increased to provide greater pain relief. This risk may be exacerbated by methadone's highly variable pharmacokinetics.4,5 In 2006, the US Food and Drug Administration issued an advisory and the label was modified to warn of the potential for unintentional overdose.5-7 This concern was reinforced by autopsy series of opioid overdose deaths with overrepresentation of methadone-related cases8,9 and a US study demonstrating a disproportionate number of prescription opioid–related overdose deaths with methadone involvement.1
Methadone also has adverse cardiac effects. It prolongs the QT interval10 and has been implicated in numerous case reports10-13 of life-threatening ventricular arrhythmias. Cases of sudden cardiac death, most of which are due to ventricular arrhythmias,14,15 have been reported in patients receiving methadone.16
Data on these life-threatening adverse effects have led to questions regarding the appropriateness of the widespread use of methadone for the treatment of chronic pain, particularly given the availability of equally effective alternatives.1,2,17 However, the one cohort study18 comparing methadone with sustained-release (SR) morphine sulfate unexpectedly found that adjusted overall mortality was 44% lower for patients receiving methadone. Given this controversy, we conducted a cohort study of patients receiving either methadone or morphine SR for noncancer pain. Given the multiple mechanisms by which an opioid can increase mortality, the primary end point of the present study was total mortality during study follow-up.
We conducted a retrospective cohort study of Tennessee Medicaid enrollees aged 30 to 74 years with a filled prescription for methadone or morphine SR between January 1, 1997, through December 31, 2009. The Vanderbilt University institutional review board approved the study, waiving the need for informed consent. The Medicaid files provided an efficient source of data for identifying the cohort, determining periods of probable exposure to medications, and ascertaining deaths.19,20 The study Medicaid database included enrollment, pharmacy, hospital, outpatient, and nursing home files and was augmented with linkage to death certificates19,21 and a statewide hospital discharge database.
To improve the study’s capacity to identify deaths related to opioids and thus reduce the potential for confounding, we focused on deaths outside the hospital in patients for whom such deaths should otherwise be relatively infrequent. Thus, to decrease the likelihood of deaths related to terminal illness, the cohort excluded persons aged 75 years or older, patients with cancer or other life-threatening diseases, and nursing home residents (eMethods and eTable 1 in the Supplement). Patients in the hospital could not enter the cohort until 30 days after discharge, because deaths during this period may be related to the reasons for the hospitalization. We excluded persons with recorded evidence of drug abuse, given the increased risk for opioid overdose unrelated to therapeutic use.
Patients entered the cohort on the date of filling the first prescription for methadone or morphine SR when they met the study inclusion or exclusion criteria (eTable 1 in the Supplement). They remained in the cohort until the end of the study, death, failure to meet inclusion criteria, or the cessation of study opioid use. Patients who left the cohort could reenter if they subsequently became eligible.
Given that both respiratory depression and cardiac arrhythmias are acute drug effects,22,23 study follow-up consisted of current use of study opioids (eFigure 1 in the Supplement). Each study opioid prescription contributed a period of current drug use (length equal to prescription days of supply) to the study follow-up. Persons given prescriptions for both study opioids during the follow-up period contributed person-time to both the morphine SR and methadone categories, although overlapping use was not permitted (eFigure 1 in the Supplement).
We excluded person-time during hospitalization and in the 30 days following hospitalization to decrease the likelihood of deaths unrelated to opioid toxic effects and thus reduce potential confounding and improve our capacity to detect an adverse effect of opioids. This method could introduce bias if the study groups differed with regard to the proportions of patients with a lethal opioid adverse effect who survived until hospital admission but died in the hospital. However, this scenario seems unlikely, given that opioid-related deaths typically occur outside the hospital. Opioid-related respiratory depression is infrequently fatal in patients administered an opioid antagonist.24 Torsades de pointes is rapidly lethal, most frequently leading to death before the patient can seek medical care.14,15,25
The primary end point was all deaths outside the hospital during study follow-up. To provide insight into the potential mechanisms for opioid toxic effects, deaths were classified into 3 subgroups: (1) sudden unexpected deaths consistent with either opioid overdose or life-threatening arrhythmias, (2) other respiratory or cardiovascular deaths for which opioid involvement was possible but less certain, and (3) other deaths, which were less likely to be related to opioid toxic effects. Classification was based on the death certificate–documented underlying cause of death, adjudication of terminal medical records (eFigure 2 in the Supplement), and computerized files with both the terminal medical encounters and death certificate information (eMethods in the Supplement).
Sudden unexpected deaths met previous definitions for either opioid overdose1,24 or sudden cardiac death (eMethods in the Supplement).10,23,26,27 These were consistent with either opioid-related respiratory depression or cardiac arrhythmias, typically occurred unexpectedly within a short interval in persons in a usual state of health, and had no evident cause other than opioid toxic effects. Opioid overdose deaths were identified from the death certificate underlying cause-of-death codes (eTable 2 in the Supplement). Prior studies28 suggest that these codes have a positive predictive value greater than 90%. Sudden cardiac deaths10 were identified from adjudication of terminal medical records23,26,27 or, when these records were unavailable, from a previously validated computer-based definition with a positive predictive value of 87% to 90% (eMethods in the Supplement).23,26
Given the US Food and Drug Administration’s focus on unintentional methadone overdose,5-7 we sought to identify opioid overdose deaths. However, the clinical circumstances of overdose and sudden cardiac death often are similar (eg, unexpected death during sleep) and it can be difficult to distinguish these mechanisms post mortem.11,13 Thus, when terminal medical records were available for adjudication, a death with a coded underlying cause of death of opioid overdose also could meet the definition for sudden cardiac death (eMethods in the Supplement). Sudden unexpected deaths were therefore further classified as meeting the definition for opioid overdose only, meeting the definition for sudden cardiac death only, and meeting both definitions. We also performed sensitivity analyses in which these definitions were mutually exclusive.
Other respiratory and/or cardiovascular deaths did not meet the definition for sudden unexpected death but potentially were related to adverse respiratory or cardiac drug effects (eMethods in the Supplement). These included deaths coded as due to respiratory causes, in which opioids may play a role.29 They also included deaths coded as due to cardiovascular causes in which sudden cardiac death had not been ruled out because medical records were unavailable (eMethods in the Supplement).
Other deaths during follow-up were considered as less likely to be related to opioid toxic effects. These were predominantly injury deaths, cardiovascular deaths for which sudden cardiac death was deemed unlikely following medical record adjudication, alcohol-related deaths, and deaths due to nonrespiratory infections.
The relative risk of death between groups defined by study opioid use status, adjusted for patient characteristics, was estimated with the hazard ratio (HR) from a proportional hazards regression model, with study opioid use as a time-dependent covariate. A single person could have both morphine SR and methadone current use person-time in the analysis (eFigure 1 in the Supplement). However, for each person, these time periods were nonoverlapping and the end point (death) occurred only once. Thus, statistical independence assumptions were not violated.30
The HRs were adjusted for potential differences between patients currently receiving methadone and morphine SR. Patient characteristics were described by 196 covariates (eTable 3 in the Supplement), which included calendar time, demographic factors, opioid indication (as previously defined,31 eMethods in the Supplement), use and dose of nonstudy opioids, cardiovascular medications and diagnoses, psychiatric medications and diagnoses, medications for musculoskeletal disorders, respiratory conditions, indicators of frailty, other proarrhythmic medications, other comorbidity, and recent medical care utilization. Given that patient comorbidity could vary markedly during follow-up, covariates were updated at the time of each prescription fill.
Given the large number of study covariates, we controlled for these by stratifying the regression analyses (eMethods in the Supplement) by deciles of either a propensity score32-34 or a mortality risk score,34-36 both of which were time dependent. The propensity score, used for comparisons of methadone vs morphine SR with no exposure subcategories, was the estimated probability that a study opioid prescription was for methadone given the covariate values at the time of the prescription fill. The distributions of the propensity score for methadone and morphine overlapped substantially (eTable 4 in the Supplement).
The mortality risk score (eFigure 3 in the Supplement) was a disease risk score34-36 for total study mortality. It was the expected risk of death during follow-up as a function of the study covariates in the absence of methadone use. Disease risk scores facilitate multiple exposure category analyses (eg, dose-specific comparisons), given that propensity scores are less suited to nonbinary comparisons.34-36
We analyzed the risk of death according to quartiles of the dose of study opioid therapy at the time of each prescription fill. We identified equivalent doses according to the cohort dose percentile distribution, which guarded against possible nonlinear dose equivalence of methadone and morphine SR. Tests for dose response or tests of a methadone effect within categories defined by dose were performed with appropriate contrasts (eMethods in the Supplement).
We conducted several sensitivity analyses. We assessed the effect of nonproportional hazards by including a time by study opioid interaction in the model.37 We performed an analysis in which each patient receiving methadone was propensity score–matched to a patient receiving morphine SR in which the absolute difference in key covariate prevalence between the 2 groups was never greater than 1% (eTable 5 in the Supplement). We analyzed several important subgroups, including those defined by calendar year, restricting follow-up to the first year of study opioid use, censoring of person-time on switching of study opioids, being younger than 65 years, having a known opioid indication, and being a new user38 of study opioids. The latter group was defined as patients with no prescription for a study opioid filled in the 31 to 365 days preceding cohort entry. We permitted prescriptions in the 30 days preceding cohort entry because this commonly occurred for patients initiating long-acting opioid therapy following hospitalization who could not enter the study cohort until 30 days following discharge.
We estimated the absolute difference in the risk of death between methadone and morphine SR. The annual incidence of total study mortality for morphine SR was unadjusted; that for methadone was calculated by multiplying the incidence in the morphine group by the HR for the entire study population.
All analyses were done with SAS, version 9.3 (SAS Institute). All P values are 2 sided.
The cohort included 38 756 persons. At the time of cohort entry, 32 742 and 6014 individuals had filled a prescription for morphine SR or methadone, respectively, of which 75.7% and 71.9% were new users. The median age of the participants was 48 years, and 57.9% were female (Table 1). The indication for the study opioid was either back pain or other musculoskeletal pain for nearly 90% of the patients. The median doses prescribed for morphine SR and methadone were 90 mg/d and 40 mg/d, respectively. There was concurrent use of nonstudy opioids for 65.4% and 47.7% of the patients receiving morphine SR and methadone, respectively. Comparable proportions of the morphine SR and methadone groups had received cardiovascular, psychotropic, and other musculoskeletal medications.
There were 477 deaths during 28 699 person-years of cohort follow-up (166 deaths per 10 000 person-years). Of the study deaths, 346 (72.5%) were sudden unexpected deaths, 53 (11.1%) were other respiratory/cardiovascular deaths, and 78 (16.4%) were other deaths.
After adjustment for study covariates, patients receiving methadone had a 46% increased risk of death during the follow-up period (Figure 1 and Table 2) (HR, 1.46; 95% CI, 1.17-1.83, P < .001). Methadone users had 72 (95% CI, 27-130) excess deaths per 10 000 person-years of follow-up.
Patients currently receiving methadone had an increased risk for sudden unexpected death (Table 2) (HR, 1.47; 95% CI, 1.13-1.90, P = .004). For deaths meeting the definition of opioid overdose, patients receiving methadone had more than a 2-fold increased risk (Table 2). There was no significantly increased risk for deaths meeting the definition for sudden cardiac death but not for opioid overdose (HR, 1.12; 95% CI, 0.80-1.59, P = .51). A sensitivity analysis with mutually exclusive definitions of opioid overdose and sudden cardiac death had essentially similar findings, as did an analysis restricted to adjudicated deaths (eFigure 4 and eTable 6 in the Supplement).
Current methadone users also had an increased risk for other respiratory/cardiovascular deaths (HR, 1.78; 95% CI, 0.91-3.46), but this was of borderline statistical significance (P = .09). The HR for other deaths was 1.26 (95% CI, 0.70-2.26; P = .45).
The risk of death during follow-up rose with increasing dose for both study opioids (Figure 2) (P = .02); the dose-response trends for the 2 study opioids did not differ significantly (P = .32). Patients receiving methadone with doses below the median had a significantly greater risk of death than did comparable patients receiving morphine SR (HR, 1.70; 95% CI, 1.23-2.34, P = .001); for doses above the median risk was increased for methadone users but was not significant (HR, 1.31; 95% CI, 0.98-1.74, P = .07). Patients in the lowest methadone dose quartile (≤20 mg/d) had a greater risk of death during follow-up than did users of comparable doses of morphine SR (<60 mg/d) (HR, 1.59; 95% CI, 1.01-2.51, P = .046).
We assessed the robustness of the primary study finding in an analysis that controlled for nonproportional hazards, in a propensity score–matched analysis, and in analyses restricted to important subgroups, including new users of the study opioids (Table 3). In each of these analyses the HR for current methadone use was increased and statistically significant. There was no evidence that the magnitude of the HRs differed from those of the primary analysis: in every case the HR for the primary analysis was included in the 95% CI for the HR from the sensitivity analysis.
We identified a cohort of Tennessee Medicaid patients receiving morphine SR and methadone for noncancer pain using inclusion and exclusion criteria designed to minimize the risk of deaths unrelated to opioid-associated adverse effects. The patients receiving methadone had a 46% increased risk for mortality during study follow-up, or an absolute excess risk of 72 deaths per 10 000 person-years of follow-up. The increased risk for methadone was present for doses of methadone as low as 20 mg/d. Findings from multiple sensitivity analyses were similar to those from the primary analysis.
Our findings differ from those of a cohort study in Veterans Affairs patients,18 which reported that adjusted overall mortality for patients receiving methadone was 44% lower than that for patients receiving morphine SR. This unexpected finding might be partially explained by the study design. Because the Veterans Affairs study included patients with life-threatening illnesses, the results could reflect imbalances between the 2 study opioid groups that were incompletely controlled for in the analysis. There was some evidence of imbalance; for example, 15% of the patients receiving methadone had a diagnosis of cancer vs 26% of those receiving morphine.
Confounding also is a potential explanation for the increased mortality that we observed in patients receiving methadone. Such bias would occur if the study methadone group had a greater risk of death related to either baseline comorbidity or to the indications for and characteristics of opioid use. However, there were several lines of evidence that our findings are not the result of confounding.
The cohort excluded patients receiving opioids who were most likely to introduce confounding: those with an elevated risk of death unrelated to therapeutic opioid use. Thus, we excluded patients with cancer and other serious illnesses. Consequently, 84% of the deaths in the study cohort were potentially related to opioid toxic effects. Patients in the morphine SR and methadone groups were comparable with regard to indications for the study opioids, their prescribed doses, and the use of cardiovascular, psychotropic, and other musculoskeletal medications. The HRs were adjusted for differences between the groups in a statistical analysis that controlled for 196 potential confounders. Finally, the results did not differ materially in a propensity score–matched analysis in which the absolute difference in covariate prevalence between the 2 groups was never greater than 1%.
In the absence of confounding or bias, the occurrence of deaths less likely to be related to opioid toxic effects should not differ according to study opioid use. The HR for these deaths was 1.26 (95% CI, 0.70-2.26), which was less than that for other study deaths and was not statistically significant. However, given the study design, only 16% of the deaths fell into this category; thus, an increased risk could not be ruled out.
Nearly three-fourths of the study deaths met accepted definitions for either opioid overdose death or sudden cardiac death. For these deaths, the risk was most pronounced for the opioid overdose deaths, regardless of whether they also met the definition for sudden cardiac death. Indeed, for such deaths, the risk for patients receiving methadone was more than twice that of those receiving morphine SR. In contrast, for deaths meeting only the definition for sudden cardiac death, the small increase in risk for patients receiving methadone was not statistically significant. This finding persisted in analyses with mutually exclusive definitions for opioid overdose and sudden cardiac death and restricted to adjudicated deaths.
The absence of a significantly increased risk for deaths meeting only the definition for sudden cardiac death may reflect the relatively low methadone doses received by the study cohort. The proarrhythmic effects of methadone are strongly dose dependent and have been most frequently reported10 in patients receiving the higher doses prescribed for treatment of opioid addiction. Most case reports11,12 describing arrhythmias involved doses greater than 100 mg/d, which is a threshold for increased electrocardiographic monitoring of patients receiving methadone.10 In this cohort of patients receiving treatment for noncancer pain, the median methadone dose was 40 mg/d; only 11% of the cohort patients had doses exceeding 100 mg/d.
This finding of no significantly increased risk for sudden cardiac death also could be influenced by misclassification of arrhythmia-related deaths as being due to opioid overdose. The clinical circumstances of each type of death may be similar, for example, an unexpected death during sleep. Postmortem drug levels often cannot reliably identify unintentional overdose deaths, given the highly patient-specific opioid toxic effect threshold and the changes in drug levels after death.29,39 When confronted with a sudden unexpected death in a patient who received long-term opioid treatment, a medical examiner reasonably might classify it as overdose related. Even when elevated drug levels are present, the proximate cause of death could be a ventricular arrhythmia. Indeed, it is widely acknowledged11,13 that given the generally available postmortem data, it is difficult to distinguish unexpected deaths due to respiratory depression from those due to arrhythmias.
Patients receiving methadone had an increased risk of death during follow-up for doses as low as 20 mg/d, which is consistent with the complex pharmacologic characteristics of methadone. Because the duration of respiratory depression is greater than that for analgesia, repeated use of low doses to control pain may lead to drug accumulation and inadvertent overdose.6 The likelihood of drug accumulation may be increased by methadone's highly variable elimination half-life, ranging from 7 to 65 hours.4,5
Our study has several limitations. Approximately one-fourth of the cohort received methadone or morphine SR prior to their entry into the study, which could attenuate the risk due to loss of persons particularly susceptible to methadone's adverse effects (depletion of susceptibles). However, restricting the cohort to new users did not materially change the findings. More than one-half of the cohort had concurrent use of a nonstudy opioid, reflecting the common practice of prescribing short-acting opioids for breakthrough pain. The drug abuse exclusion was based on medical care encounters, which could lead to bias if methadone was prescribed for unrecorded substance abuse. However, the increased risk for the lowest doses of methadone suggests that this did not explain the study findings. The cohort consisted of Tennessee Medicaid enrollees; thus, both the program and population characteristics might affect the findings.
In patients with noncancer pain, the risk of out-of-hospital death in patients receiving methadone was 46% greater than that for those receiving morphine SR. The absolute excess risk was 72 deaths per 10 000 person-years of study opioid use. An increased risk was present for methadone doses as low as 20 mg/d. These findings support recommendations that methadone should not be considered a drug of first choice for noncancer pain.
Accepted for Publication: September 14, 2014.
Corresponding Author: Wayne A. Ray, PhD, Department of Health Policy, Vanderbilt University School of Medicine, Village at Vanderbilt, Ste 2600, 1501 21st Ave, Nashville, TN 37212 (email@example.com).
Published Online: January 19, 2015. doi:10.1001/jamainternmed.2014.6294.
Author Contributions: Dr Ray had full access to all the data in the study and takes responsibility for the integrity of the data and the accuracy of the data analysis.
Study concept and design: Ray, Chung, Murray, Cooper, Stein.
Acquisition, analysis, or interpretation of data: Chung, Murray, Cooper, Hall.
Drafting of the manuscript: Ray, Murray, Stein.
Critical revision of the manuscript for important intellectual content: All authors.
Statistical analysis: Ray.
Obtained funding: Ray, Murray, Stein.
Administrative, technical, or material support: Ray, Chung, Murray, Hall, Stein.
Study supervision: Ray, Chung, Cooper, Stein.
Conflict of Interest Disclosures: None reported.
Funding/Support: The study was supported by a grant from the National Heart, Lung, and Blood Institute (HL081707), the National Institute of Arthritis and Musculoskeletal and Skin Diseases (K23AR064768), and a Vanderbilt Physician Scientist Development award (Dr Chung).
Role of the Funder/Sponsor: The funding sources had no role in the design and conduct of the study; collection, management, analysis, and interpretation of the data; preparation, review, or approval of the manuscript; and decision to submit the manuscript for publication.
Additional Contributions: We gratefully acknowledge the Tennessee Bureau of TennCare, the Tennessee Department of Health, and Tennessee Donor Services, which provided study data.
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