Ransom, Jeanine E.

Background: Tennis elbow is commonly encountered by physicians, yet little is known about the cost of treating this condition. Hypothesis: The largest cost associated with treating tennis elbow is procedural intervention. Study Design: Descriptive epidemiology study. Level of Evidence: Level 4. Methods: This retrospective population-based study reviewed patients who were treated for new-onset tennis elbow between January 1, 2003 and December 31, 2012. All patients were followed up through their medical and administrative records to identify health care encounters and interventions for tennis elbow. Unit costs for each health service/procedure were adjusted to nationally representative unit costs in 2013 inflation-adjusted dollars. Results: In a cohort of 931 patients who had 2 or more clinical encounters for new-onset lateral epicondylosis during a 12-month period after initial diagnosis, 62% received a median of 3 physical therapy sessions (cost, $100/session) and 40% received a median of 1 steroid injection (cost, $82/injection). Only 4% of patients received surgical intervention with mean costs of $4000. The mean (median) total direct medical cost of services related to lateral epicondylosis for the entire cohort was $660 ($402) per patient over the 1-year period after diagnosis. Patients who continued to be treated conservatively between 6 and 12 months after diagnosis incurred relatively low median costs of $168 per patient. Conclusion: In this cohort, a second encounter with a physician for tennis elbow was a strong predictor of increased treatment cost due to a higher likelihood of specialist referral, use of physical therapy, or treatment with steroid injection. Clinical Relevance: The majority of direct medical spending on tennis elbow occurs within the first 6 months of treatment, and relatively little expense occurs between 6 and 12 months after diagnosis unless a patient undergoes surgical intervention.

Abstract Background Objective cost estimates and source of cost differences are needed across the spectrum of cognition, including cognitively normal (CN), mild cognitive impairment (MCI), newly discovered dementia, and prevalent dementia. Methods Subjects were a subset of the Mayo Clinic Study of Aging stratified‐random sampling of Olmsted County, MN, residents aged 70 to 89 years. A neurologist reviewed provider‐linked medical records to identify prevalent dementia (review date = index). Remaining subjects were invited to participate in prospective clinical/neuropsychological assessments; participants were categorized as CN, MCI, or newly discovered dementia (assessment date = index). Costs for medical services/procedures 1‐year pre‐index (excluding indirect and long‐term care costs) were estimated using line‐item provider‐linked administrative data. We estimated contributions of care‐delivery site and comorbid conditions (including and excluding neuropsychiatric diagnoses) to between‐category cost differences. Results Annual mean medical costs for CN, MCI, newly discovered dementia, and prevalent dementia were $6042, $6784, $9431, $11,678, respectively. Hospital inpatient costs contributed 70% of total costs for prevalent dementia and accounted for differences between CN and both prevalent and newly discovered dementia. Ambulatory costs accounted for differences between CN and MCI. Age‐, sex‐, education‐adjusted differences reached significance for CN versus newly discovered and prevalent dementia and for MCI versus prevalent dementia. After considering all comorbid diagnoses, between‐category differences were reduced (e.g., prevalent dementia minus MCI (from $4842 to $3575); newly discovered dementia minus CN (from $3578 to $711)). Following the exclusion of neuropsychiatric diagnoses from comorbidity adjustment, between‐category differences tended to revert to greater differences. Conclusions Cost estimates did not differ significantly between CN and MCI. Substantial differences between MCI and prevalent dementia reflected high inpatient costs for dementia and appear partly related to co‐occurring mental disorders. Such comparisons can help inform models aimed at identifying where, when, and for which individuals proposed interventions might be cost‐effective.

Background: Lateral elbow tendinosis (epicondylitis) is a common condition both in primary care and specialty clinics. Purpose: To evaluate the natural history (ie, incidence, recurrence, and progression to surgery) of lateral elbow tendinosis in a large population. Study Design: Descriptive epidemiology study. Methods: The study population comprised a population-based incidence cohort of patients with new-onset lateral elbow tendinosis between January 1, 2000, and December 31, 2012. The medical records of a 10% random sample (n = 576) were reviewed to ascertain information on patient and disease characteristics, treatment modalities, recurrence, and progression to surgery. Age- and sex-specific incidence rates were calculated and adjusted to the 2010 US population. Results: The age- and sex-adjusted annual incidence of lateral elbow tendinosis decreased significantly over time from 4.5 per 1000 people in 2000 to 2.4 per 1000 in 2012 ( P < .001). The recurrence rate within 2 years was 8.5% and remained constant over time. The proportion of surgically treated cases within 2 years of diagnosis tripled over time, from 1.1% during the 2000-2002 time period to 3.2% after 2009 ( P < .00001). About 1 in 10 patients with persistent symptoms at 6 months required surgery. Conclusion: The decrease in incidence of lateral elbow tendinosis may represent changes in diagnosis patterns or a true decrease in disease incidence. Natural history data can be used to help guide patients and providers in determining the most appropriate course at a given time in the disease process. The study data suggest that patients without resolution after 6 months of onset may have a prolonged disease course and may need surgical intervention.

Background and Purpose —There are few population-based data available regarding nursing home use after stroke. This study clarifies the use of a nursing home after stroke, as well as its dependence on stroke severity, in a defined population. Methods —All first stroke events among residents of Rochester, Minn, during 1987–1989 were ascertained, subtyped, and assigned Rankin disability scores (RS) before the event, at maximal deficit, and at specified intervals after stroke. Persons were followed from the date of stroke event to death, emigration from Rochester, or December 31, 1994, in complete community-based medical records and Minnesota Case Mix Review Program data tapes to determine nursing home residency before stroke and at 90 days and 1 year after stroke, proportion of survival days in a nursing home, and cumulative risk of admission to a nursing home. Results —There were 251 cases of first cerebral infarction, 24 intracerebral hemorrhages, and 15 subarachnoid hemorrhages among residents of Rochester during 1987–1989. The maximal deficit RS was 1 or 2 for 62 (25%), RS 3 for 72 (29%), and RS 4 or 5 for 117 (47%) of the cerebral infarct patients. Among patients surviving to 90 days or 1 year after cerebral infarction, 25% were in nursing home at 90 days and 22% at 1 year, respectively. Within these maximal deficit RS categories, the percentages of follow-up time spent in a nursing home during the first post–cerebral infarction year are as follows: RS 1 to 2, 4%; RS 3, 10%; and RS 4 to 5, 54%. Multivariate logistic regression revealed that increasing age and RS 4 to 5 at maximal deficit were independent predictors ( P <0.0001) of nursing home residency at 90 days and 1 year after stroke, whereas stroke type was not an independent predictor. At 1 year after cerebral infarction, the Kaplan-Meier estimates of proportion of people with at least 1 nursing home admission were 11% for RS 1 to 2, 22% for RS 3, and 68% for RS 4 to 5. Conclusions —This study provides unique population-based data regarding the short- and long-term use of a nursing home after stroke and its dependence on stroke severity. More than 50% of people with a severe cerebral infarction are in a nursing home 90 days and 1 year after the stroke, and by 1 year, nearly 70% will have required some nursing home stay. Age and stroke severity are independent predictors of nursing home residency after stroke.

OBJECTIVE—Birth weight is a risk factor for both diabetes and mortality. Diabetes is a risk factor for mortality. Whether the excess mortality observed for diabetes varies with birth weight is unclear. RESEARCH DESIGN AND METHODS—Among all 2,508 Rochester, Minnesota, residents who first met research criteria for adult-onset diabetes in 1960–1995, 171 were born locally in-hospital after 1922 (i.e., birth weights available) as singleton, term infants. Each case subject and two age- and sex-matched nondiabetic control subjects (born locally, residing locally when the case subject met the criteria for diabetes) were followed through 31 December 2000 for vital status. RESULTS—Of the diabetic case subjects, 16% (27 of 171) died vs. 7% (25 of 342) of control subjects (P = 0.004). The difference was less for normal-birth-weight (NBW) (2,948–&lt;3,856 g) individuals (12% [12 of 102] vs. 8% [20 of 246], P = 0.31) than for abnormal-birth-weight individuals (low birth weight [LBW] 20% [8 of 39] vs. 2% [1 of 46], P = 0.01; high birth weight [HBW] 23% [7 of 30] vs. 8% [4 of 50], P = 0.16), as confirmed with age- and sex-adjusted Cox proportional hazards (diabetes-associated hazard ratio 1.4 [95% CI 0.69–2.90] for NBW vs. 4.8 [1.7–13.3] for abnormal birth weight, test for interaction P = 0.056). The observed diabetes deaths were greater than expected, based on mortality for the general population (27 vs. 13.3, P &lt; 0.001), with 70% of excess deaths occurring among LBW (8 vs. 2.2, P &lt; 0.001) and HBW (7 vs. 3.1, P = 0.03) individuals. CONCLUSIONS—The excess mortality observed for diabetes appears disproportionately concentrated among abnormal-birth-weight individuals, thus identifying a subset of at-risk diabetic individuals and reinforcing the importance of NBW deliveries.

OBJECTIVE—The aims of this study were to provide estimates of 1) the risk of mortality for individuals with both diabetes and peripheral arterial disease (PAD) relative to that for individuals with either condition alone and 2) the association between PAD progression and mortality for individuals with diabetes, PAD, and both conditions. RESEARCH DESIGN AND METHODS—This longitudinal cohort study was conducted in Rochester, Minnesota. Local residents age 50–70 years with a prior diagnosis of PAD and/or diabetes were identified from the Mayo Clinic diagnostic registry and invited to a baseline examination (1977–1978). Those who met inclusion criteria were assessed for PAD progression at 2 and 4 years and followed for vital status through 31 December 1999. RESULTS—The numbers who met criteria for PAD, diabetes, and both conditions at baseline were 149, 238, and 186, respectively. Within each group, observed survival was less than expected (P &lt; 0.001). The adjusted risk of death for both conditions was 2.2 times that for PAD alone. Among the 449 who returned at 4 years, the risk of subsequent death was greater for those whose PAD had progressed; among individuals with diabetes alone at baseline, 100% (17 of 17) who met criteria for PAD progression were dead by 31 December 1999 compared with 62% (111 of 178) of those who had not met criteria (adjusted relative hazard 2.29 [95% CI 1.30–4.02], P = 0.004). The increased mortality associated with PAD progression was significant only for individuals with diabetes (alone or with PAD). CONCLUSIONS—Diabetes is a risk factor for both PAD and PAD-associated mortality, emphasizing the critical need to detect and monitor PAD in diabetic patients.

OBJECTIVE—Increasing obesity within the general population has been accompanied by rising rates of diabetes. The extent to which obesity has increased among people with diabetes is unknown, as are the potential consequences for diabetes outcomes. RESEARCH DESIGN AND METHODS—Community medical records (hospital and ambulatory) of all Rochester, Minnesota, residents aged ≥30 years who first met standardized research criteria for diabetes from 1970 to 1989 (n = 1,306) were reviewed to obtain data on BMI and related characteristics as of the diabetes identification date (±3 months). Vital status as of 31 December 1999 and date of death for those who died were obtained from medical records, State of Minnesota death tapes, and active follow-up. RESULTS—As of the identification date, data on BMI were available for 1,290 cases. Of the 272 who first met diabetes criteria in 1970–1974, 33% were obese (BMI ≥30), including 5% who were extremely obese (BMI ≥40). These proportions increased to 49% (P &lt; 0.001) and 9% (P = 0.012), respectively, for the 426 residents who first met diabetes criteria in 1985–1989. BMI increased significantly with increasing calendar year of diabetes identification in multivariable regression analysis. Analysis of survival revealed an increased hazard of mortality for BMI ≥41, relative to BMI of 23–25 (hazard ratio 1.60, 95% CI 1.09–2.34, P = 0.016). CONCLUSIONS—The prevalence of obesity and extreme obesity among individuals at the time they first met criteria for diabetes has increased over time. This is disturbing in light of the finding that diabetic individuals who are extremely obese are at increased risk of mortality compared with their nonobese diabetic counterparts.