Phenylpropanolamine & Risk of Hemorrhagic Stroke

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Phenylpropanolamine & Risk of Hemorrhagic Stroke: JoAnn Guest Mar 06, 2005

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Final Report of The Hemorrhagic Stroke Project

Yale University School of Medicine 10may00

May 10, 2000

http://www.mindfully.org/Health/Phenylpropanolamine-Hemorrhagic-Stroke.htm

 

 

Prepared by:

 

Ralph I. Horwitz, M.D.

Harold H. Hines Jr. Professor of Medicine and Epidemiology

Yale University School of Medicine

 

Lawrence M. Brass, M.D.

Professor of Neurology and Epidemiology and Public Health

Yale University School of Medicine

 

Walter N. Kernan, M.D.

Associate Professor of Medicine

Yale University School of Medicine

 

Catherine M. Viscoli, Ph.D.

Associate Research Scientist

Yale University School of Medicine

 

On Behalf of the HSP Investigators

 

EXECUTIVE SUMMARY

Case reports have linked exposure to phenylpropanolamine (PPA) to the

occurrence of hemorrhagic stroke. Many of the affected patients have

been young women using PPA as an appetite suppressant, often after a

first dose. To further examine the association between PPA and

hemorrhagic stroke, we designed a case-control study involving men and

women ages 18 to 49 years who were hospitalized with a subarachnoid

hemorrhage (SAH) or intracerebral hemorrhage (ICH). Eligible case

subjects had no prior history of stroke and were able to participate in

an interview within 30 days of their event. Case subjects were recruited

from hospitals in four geographic regions of the United States. For each

case subject, random digit dialing was used to identify two control

subjects who were matched on age, gender, race, and telephone exchange.

Cases and control subjects were interviewed to ascertain their medical

history, health behaviors, and medication usage. A subject was

classified as exposed to PPA if they reported use within 3 days of the

stroke event for case subjects or a corresponding date for control

subjects, and the exposure was verified.

 

The final study cohort comprised 702 case subjects and 1376 control

subjects. All control subjects were matched to their case subjects on

gender and telephone exchange. Age matching was successful for 1367

controls (99%) and ethnicity matching was achieved for 1321 controls

(96%). For the association between hemorrhagic stroke and any use of PPA

within three days, the adjusted odds ratio was 1.49 (lower limit of the

one-sided 95% confidence interval (LCL)=0.93, p=0.084). For the

association between hemorrhagic stroke and PPA use in cough-cold

remedies within the three-day exposure window, the adjusted odds ratio

was 1.23 (LCL=0.75, p=0.245). For the association between hemorrhagic

stroke and PPA use in appetite suppressants within the three-day

exposure window, the adjusted odds ratio was 15.92 (LCL=2.04, p=0.013).

For the association between PPA in appetite suppressants and risk for

hemorrhagic stroke among women, the adjusted odds ratio was 16.58

(LCL=2.22, p=0.011). For first dose PPA uses among women, the adjusted

odds ratio was 3.13 (LCL= 1.05, p = 0.042). All first dose PPA use

involved cough-cold remedies.

 

In conclusion, the results of the HSP suggest that PPA increases the

risk for hemorrhagic stroke. For both individuals considering use of PPA

and for policy makers, the HSP provides important data for a

contemporary assessment of risks associated with the use of PPA.

 

INTRODUCTION

Phenylpropanolamine (PPA) is a synthetic sympathomimetic amine

structurally similar to pressor amines (i.e., epinephrine,

phenylephrine, and ephedrine) and central nervous system stimulants

(i.e., ephedrine, amphetamine). It is a common ingredient in cough-cold

remedies and appetite suppressants. Each year, billions of doses are

consumed in the United States, making PPA one of the most commonly used

non-prescription medications(1).

 

Since 1979, over 30 published case reports have described the occurrence

of intracranial hemorrhage after PPA ingestion(1-3). Early reports

involved diet pills including both PPA and caffeine(4-7), a combination

that was removed from the market in 1983 because of abuse potential(.

Later reports involved use of PPA alone(9, 10), often as a reported

first-ever dose (7, 11-14). Like the earlier ones, however, the later

reports primarily involved PPA in diet pills. Affected patients were

most commonly young persons, particularly women, ages 17 to 45 years. At

least five reports, however, involved PPA in cough cold

preparations(15-19).

 

Other than case reports, there has been only one epidemiologic study of

PPA and stroke, published in 1984(20). Investigators at a large health

maintenance organization examined the occurrence of cerebral hemorrhage

among patients ages less than 65 years who filled a prescription for PPA

during 1977-1981. The relative risk for hemorrhage for PPA users

compared to non-users was 0.59 (95% confidence interval 0.03-2.9). Since

the relative risk was less than one and the upper bound of the

confidence interval was about three, the authors concluded that any

hemorrhage risk related to PPA, if present at all, is very small.

 

Responding to on-going concern about PPA and risk for hemorrhagic

stroke, in 1992 the United States Food and Drug Administration joined

with manufacturers of products containing phenylpropanolamine to

recommend the conduct of an epidemiological study of the association.

Because several case reports involved young women using PPA as an

appetite suppressant, often after a first dose, the FDA and

manufacturers identified women as having high priority in research

planning. In response, the investigators of this research designed and

implemented the Hemorrhagic Stroke Project (HSP), with three co-equal

specific aims: Among men and women ages 18-49 years, to estimate the

association between PPA and hemorrhagic stroke; Among the same target

group, to estimate the association between PPA and hemorrhagic stroke by

type of PPA exposure (cough-cold remedy or appetite suppression); And

among women ages 18-49 years, to estimate: a) the association between

first use of PPA and hemorrhagic stroke and b) the association between

PPA in appetite suppressants and hemorrhagic stroke.

 

METHODS

Recruitment and Classification of Patients with Hemorrhagic Stroke

Between December 1994 and July 1999, we identified potential case

subjects from two hospital networks located in Connecticut/Southern

Massachusetts and Southern Ohio/Northern Kentucky and two tertiary care

hospitals in Providence, Rhode Island and Houston, Texas through active

surveillance of all admissions (See Appendix A for list of hospitals).

Surveillance involved review of admission rosters and direct monitoring

of admissions by one or more designated individuals, such as a discharge

planner or stroke nurse. As a check on the completeness of case

ascertainment, discharge rosters were reviewed from each participating

hospital.

 

Case subject eligibility criteria included admission to a participating

hospital, age between 18 and 49 years (inclusive) and symptomatic

primary subarachnoid hemorrhage (SAH) or primary intracerebral

hemorrhage (ICH). Subdural hematomas and hemorrhages related to ischemic

infarctions, trauma, thrombolytic therapy, or cerebral vein thrombosis

were not considered primary events and were not eligible for this study.

A SAH was diagnosed based on clinical symptoms and specific diagnostic

information. Required symptoms included sudden, severe headache, stiff

neck or change in level of consciousness. Required diagnostic

information included the presence of a high intensity signal in the

subarachnoid space on computed tomography, or xanthochromasia on lumbar

puncture not explained by other etiologies (e.g. liver disease,

increased CSF protein, hypervitaminosis A). An ICH was diagnosed by

symptoms (sudden headache, focal neurological symptoms, or change in

consciousness) accompanied by a computed tomographic (CT) scan showing a

hyperintense signal within brain parenchyma. Magnetic resonance imaging

was accepted for the diagnosis of SAH or ICH only if other studies were

not diagnostic. Subjects were ineligible for enrollment if they died

within 30 days, were not able to communicate within 30 days, had a

previously diagnosed brain lesion predisposing to hemorrhage risk (e.g.

arteriovenous malformation, vascular aneurysm, or tumor), or a prior

stroke. We also excluded patients who first experienced stroke symptoms

after being in the hospital for 72 hours (e.g., for an unrelated

matter).

 

Permission to contact each potential case subject was sought from the

treating physician. If permission were received, a researcher met with

the patient and reviewed pertinent data to confirm eligibility.

 

Once subject interviews were completed and medical records were

acquired, a second and final check on eligibility was completed at the

central office in New Haven by a researcher who was kept unaware of

medication exposures. This procedure was designed to ensure uniform

standards for documentation and eligibility across all research sites.

The researcher reviewed medical records and all study forms except the

interview booklet. (For subjects enrolled in the Ohio/Kentucky site,

medical records were reviewed on site. Study forms, however, were also

checked in New Haven). Missing radiology or laboratory data was

obtained. Patients with uncertain eligibility were reviewed with the

local investigator and the New Haven investigators before being

disqualified.

 

Recruitment of Controls

We attempted to identify two matched controls for each case subject.

Matching criteria included: 1) gender; 2) ethnic group (black versus

non-black); and 3) age (within 3 years for case subjects less than 30

years and within 5 years for cases 30 years or over). In addition, all

control subject interviews had to be completed within 30 days of the

case's stroke event to minimize seasonal differences in the likelihood

of exposure to cough-cold remedies. A computer-generated list of random

telephone numbers (matching the first three digits of the case subject

telephone number) was used to identify potential control subjects.

Eligibility criteria for control subjects were the same as for case

subjects except for those criteria related to the stroke event.

 

Subject Interviews

Eligible patients were invited to participate and give verbal informed

consent. During the consent procedure, all subjects (cases and controls)

were told that the study was designed to examine causes of hemorrhagic

stroke in young persons without specific mention of PPA or other

potential risk factors. Case subjects who did not speak English (n = 35)

were interviewed using a translator who, with five exceptions, was not a

relative or acquaintance. Most case-subject interviews were conducted in

the hospital, but some were completed at home and three were completed

by telephone. Most control-subject interviews were conducted in person

in the control subject's home, a doctor's office, or other convenient

location (44 were conducted by telephone). Control subjects were offered

twenty dollars to defray their expenses.

 

Specification of Focal Time

The first step of the interview process was to determine the focal time

for each case subject. This time refers to the calendar day (i.e. index

day) and time of day that marked the onset of symptoms plausibly related

to hemorrhage and that caused the case subject to seek medical

attention. The correct assignment of a focal time was critical because

exposures to PPA were defined in relation to this temporal anchor (i.e.,

only exposures that occurred prior to the focal time are relevant to the

analysis). To establish the focal time, we obtained from each case

subject a detailed account of his or her symptoms from onset to

diagnosis. Additional information was obtained from acquaintances or

witnesses to the case subject's illness.

 

Some patients with SAH or ICH may have a transient headache hours or

days before the event that actually causes them to seek medical

attention(21). These preceding symptoms have been termed " warning leaks "

or " sentinel headaches " (22). The cause of these sentinel headaches is

not known, although clinicians infer that some of them may be due to

minor bleeding(23). As expected, we encountered patients with sentinel

headache and recognized the possibility that the first onset of bleeding

may have coincided with that symptom. Since we defined the focal time by

the onset of the symptoms that actually brought them to medical

attention, we needed an additional research strategy to account for the

possibility of an earlier onset of bleeding. Accordingly, we defined an

alternate focal time as the time of onset of the sentinel headache. A

separate interview was conducted for this alternate focal time (for the

case and matched controls). How data on the alternate focal time were

analyzed is described below in the statistical analysis section.

 

The focal time for control subjects was established according to two

rules: first, it occurred on one of the 7 days prior to the control

interview; and second, it was matched to the case subject's focal time

according to day of week and time of day. For example, if the case

subject's focal time occurred on a Monday at 2:00pm and their control

subject was interviewed on a Thursday, then the focal time for the

control subject interview would be 2:00pm on the Monday preceding the

control interview. We maintained a short interval between control

subjects' focal times and interview dates to improve control subjects'

recall of pre-focal time exposures; the short interval was necessary to

balance case subjects' greater stimulation for recall of exposures

occurring before their stroke. The focal time matching was intended to

minimize differences between case subjects and control subjects for

differences in medication use and other exposures (e.g. alcohol,

cigarettes) that may occur by the day of week (e.g., weekday vs.

weekend) and time of day.

 

Ascertainment of Exposure Data and Other Subject Information

Case and control subjects were interviewed by a trained interviewer

using a structured questionnaire developed for this study. To assist

subjects in recalling details of their medication use, they were asked

to refer to a calendar for the period encompassing the index date

(before the focal time) and the preceding 14 days. The calendar was

marked with notable personal events, such as trips, birthdays and

doctor's visits that occurred during this period. As a memory aid,

subjects were asked to recall if specific symptoms of a cold or flu

(i.e., cough, runny nose, nasal congestion, or sore throat) were present

in this two-week period. If present, the subject was asked if they had

used any medication to treat the symptoms. The subject was then asked to

recall the names of any other medications (purchased over the counter or

prescribed by a doctor) taken during this time period. After all

volunteered medications were recorded, subjects were asked if they had

taken any of several specific classes of medications in the two-week

period before the focal time (i.e., aspirin, acetaminophen,

non-steroidal anti-inflammatory medications, anticoagulants, asthma

medications, and medications for depression). Subjects were specifically

asked about their lifetime use and last use of diet pills. For each

medication reported taken during the two-week period, details were

obtained regarding certainty of use (definite, probable, or uncertain)

and amount taken on the index day prior to focal time and on each of the

preceding three calendar days. For each medication, the timing and

amount of the last dose taken before the focal time was noted. Only uses

that were reported as probable or definite were counted in the analysis.

 

 

Verification of each reported medication was done after the interview.

Participants were asked to pick out reported brand-name medications from

a book containing photographs (Product Identification Book). They were

then asked to produce the actual container of each medication reported.

The exact name of the medication and the manufacturer's lot numbers were

recorded. If a container was not available, use of a brand name

medication was considered verified if the subject had identified it in

the Product Identification Book. Among 28 reported PPA uses in a three

day window for case subjects, 27 (96%) were verified. Among 35 reported

PPA uses for control subjects, 33 (94%) were verified.

 

To determine the active ingredients in each verified medication, we

relied on published sources(24, 25). For national brands and

prescription drugs that had possible formulation changes during the

study period and for generic or store brand medications, we verified

active ingredients directly with the manufacturer.

 

Definition of Exposure

The exposure window refers to the interval before the focal time when

subject's exposure status to PPA is assessed. For all analyses except

first-dose use, the exposure window was defined as the index day before

focal time and the preceding three calendar days. For first-dose use, a

subject was considered exposed if the PPA use occurred on the index day

before the focal time or on the preceding calendar day, with no other

PPA uses during the preceding two weeks. Subjects were only considered

exposed to PPA within the appropriate window if their reported exposure

was verified by the procedures described above. To maintain a consistent

reference group for all analyses, non-exposure was defined by no use of

PPA within two weeks of the index date.

 

Sample Size

The sample size was based on the need to determine if PPA, taken as a

first use, increases the risk of hemorrhagic stroke within 24 hours

among women ages 18-49. Based on available market data, we estimated

that 0.502% of controls would report an exposure to PPA within 24 hours

of the index date. For a one-tailed test of significance at the 0.05

level and 80% power to detect an odds ratio of 5.0 for first use

exposure in women, 324 female case subjects and 648 control subjects

were required. Because our research interest extended to men and to

exposures other than PPA, we doubled our sample size to include 350 men

and 350 women for a total of 700 case subjects and 1400 controls.

 

Statistical Analysis

In the first phase of the analysis, we compared case and control

subjects on a variety of clinical and demographic features, including

those used in matching. Statistical comparisons were made using

chi-square tests and the Fisher's exact test, where appropriate, for

categorical variables and the Student t-test for continuous variables

(SAS Institute Inc, Cary, North Carolina). In the second phase of the

analysis, conditional logistic models for matched sets (with a variable

number of controls per case) were used to estimate odds ratios, lower

limits of the one-sided 95% confidence intervals and p-values for the

risk factors under investigation. We report one-tailed statistical

results because the focus of this research was restricted to the effect

of PPA in increasing the risk of stroke. Each logistic model was

estimated with two mutually exclusive, binary exposure terms: the first

term captured the subject's primary exposure status as defined by the

specific aim (e.g., PPA use in 3-day window; yes/no); the second term

captured users of PPA who were not exposed according to the primary

definition (e.g., no PPA use in the three day window, but some PPA

exposure within two weeks of the focal time). This modeling technique

was employed in order to retain all matched subject-sets in the

estimation of the primary exposure of interest while maintaining a

consistent reference group of no-PPA use in two weeks for all analyses.

Unadjusted estimates were calculated using exact methods (LogXact

Program, version 2.1, Cytel Software Corporation, Cambridge, MA).

 

In multivariate conditional logistic models (using asymptotic methods),

we adjusted for race (black compared with non-black) because not all

cases and controls were successfully matched on this factor, and for

history of hypertension (yes/no) and current cigarette smoking (current

compared with never or ex-) because these are the major risk factors for

SAH or ICH(26, 27). We also examined other subject features to determine

if any of them, when added to this basic adjusted model, altered the

matched odds ratio by at least 10%. Clinical features examined included

body mass index, diabetes, polycystic kidney disease, congestive heart

failure, sickle cell anemia, and clotting disorders. Other subject

features included education (less than high school/high school or more),

family history of hemorrhagic stroke in a first degree relative,

consumption of more than two alcoholic beverages per day, cocaine use

within one day of the index day, current oral contraceptive use, and

several other pharmacological exposures. The final model included race,

current cigarette use, hypertension, and any features that changed the

adjusted matched odds ratio by at least 10%.

 

The study design incorporated two interim examinations of the data. As

specified in the protocol, each interim analysis examined the

association between stroke and any PPA use in the three-day window for

the entire cohort and first PPA use in women. To preserve the specified

alpha level for statistical inference (type I error rate), the

O'Brien-Fleming multiple testing procedures were used for testing the

significance of effect estimates(2. For three looks at the data (two

interim and one final look) and a final alpha level of 0.05 for a

one-tailed test of each objective, the O'Brien-Fleming chi-square

statistics corresponded to one-tailed p-values of 0.008 (first look),

0.023 (second look), and 0.044 (final look).

 

As described above, we encountered patients with sentinel symptoms for

whom we defined an alternate focal time. If the actual disease onset is

defined by the alternate focal time rather than the primary focal time,

an analysis based on the latter could lead to a biased estimate of the

odds ratio by two mechanisms. First, case subject exposure to PPA

between the alternate (earlier) and primary (later) focal time may

actually be in response to the disease (and not a potential cause of the

disease). If such exposures occur within the appropriate time window,

however, they would be counted in the analysis using the primary focal

time and would lead to an overestimate of the odds ratio. This bias is

sometimes referred to as temporal precedence bias(29, 30). Second, case

subject exposure to PPA before the alternate focal time may actually be

causally associated with the disease. If these earlier exposures were

outside the appropriate time window for the primary focal time analysis,

however, they would not be counted and the error would lead to an

underestimate of the odds ratio. To examine the potential effect of

these two sources of bias, we performed two ancillary analyses, one

excluding all subject sets where the case subject had a possible

sentinel symptom and one using the alternate focal time.

 

Oversight

An external Scientific Advisory Group (SAG) operated autonomously from

the investigators and study sponsors and provided general oversight for

the conduct of the HSP. The SAG reviewed the research protocol and

suggested revisions, reviewed research progress, developed criteria for

early termination, and evaluated interim and final analyses.

Nevertheless, the investigators accept all responsibility for the

conduct of the study, the analysis of the data, and the interpretation

of the results. A list of advisory group membership is included at the

end of this report.

 

RESULTS

Study Participants

Between December 1994 and July 1999, 1,714 potentially eligible patients

with hemorrhagic stroke were identified (Table 1). Among these, 784

subjects were ineligible for enrollment (389 died, 194 were unable to

communicate within 30 days of the event, 120 had a history of stroke, 48

had a known brain tumor or arteriovenous malformation, and 33 were in

the hospital over 72 hours before their event). Among the 930 eligible

subjects, 222 were not enrolled (182 were not contacted within 30 days,

37 declined to give informed consent, and for 3 their physician did not

give permission for contact). The total number of eligible case subjects

enrolled was 708. There were no significant differences (p<0.05) in age

or ethnicity between the 708 enrolled patients and the 222 eligible

patients who were not enrolled. Subjects in the enrolled case group,

however, were more likely to be female (55% compared with 45%, p = 0.02)

and to have a SAH (60% compared with 47%, p < 0.01). Six patients were

not included in the final analysis because no eligible controls were

identified (three patients), the interview date was over 30 days from

the stroke event (two patients), or the focal time could not be

determined (one patient). Thus, the final case group comprised 702

subjects, including 425 subjects (60%) with a SAH and 277 (40%) with an

ICH. Hemorrhage was associated with an aneurysm in 307 patients (44%),

an arteriovenous malformation in 50 patients (7%), and a tumor in one

patient (0.1%). (See Appendix B for additional details on case

enrollment)

 

 

No.

 

Hemorrhagic Strokes Identified (Dec. 1994 – Aug. 1999)

1,714

 

 

 

 

Ineligible Subjects

784

 

Died within 30 days of stroke

389

 

Not able to communicate within 30 days of stroke

194

 

Prior history of stroke

120

 

Prior history of brain tumor or AVM

48

 

In hospital > 72 hours before stroke

33

 

 

 

 

Eligible Subjects1

930

 

Not enrolled

222

 

Not contacted within 30 days

182

 

Refused participation

37

 

No MD approval to contact

3

 

Enrolled

708

 

 

 

 

Two control subjects were located for 674 case subjects (96%) and one

control subject for 28 case subjects (4%). All control subjects were

matched to their case subjects on gender and telephone exchange. Age

matching was successful for 1367 controls (99%) and ethnicity matching

was achieved for 1321 controls (96%). On average, for each case subject,

we called 151 telephone numbers (range 3-1119) and identified 2.8

eligible persons for each enrolled control (range 1-12).

 

The mean interval between the case index date and the interview date was

13 days (range 0 to 30) for case subjects and 19 days for their matched

control subjects (range 0 to 35). All control subjects except six were

interviewed within 30 days of the case subject's index date. We allowed

enrollment of six controls within an interval of 31-35 days since the

purpose of interviewing controls within 30 days of the case subject's

index date was simply to ensure seasonal similarity when collecting data

on pharmaceutical exposures.

 

Table 2 shows selected characteristics of case and control subjects.

Compared to control subjects, case subjects were significantly (p<0.05)

more likely to be black (21% compared with 17%). Case subjects were also

more likely to report lower educational achievement (20% did not

graduate from high school compared with 9% of control subjects), current

cigarette smoking (51% compared with 30%), a history of hypertension

(39% compared with 20%), family history of hemorrhagic stroke (9%

compared with 5%), heavy alcohol use (14% compared with 7%), and recent

cocaine use (2% compared with <1%). For all other clinical variables

examined, case and control subjects were not dissimilar.

 

Table 3 shows non-PPA pharmacological exposures of case and control

subjects. Case subjects were significantly (p<0.05) less likely to

report use of NSAIDS and significantly more likely to report use of

caffeine and nicotine in the three days before their focal time.

 

To identify variables for inclusion in subsequent multivariable models,

we sequentially tested each clinical feature (Table 2) and

pharmacological exposure (Table 3) in the basic conditional logistic

model that included race, hypertension, and current cigarette smoking.

Under any PPA exposure definition, only education changed the adjusted

odds ratio for the association between PPA and hemorrhagic stroke by

more than 10% and was included in all subsequent models.

 

Association Between PPA and Hemorrhagic Stroke

Our first co-equal specified aim was to determine whether PPA users,

compared to non-users have an increased risk of hemorrhagic stroke. In

Table 4, frequencies are shown for exposures to PPA in an unmatched

format for the case and control subjects, with unadjusted and adjusted

matched odds ratios provided. (See Appendix C for exposures presented in

matched format.) The odds ratios were calculated from conditional

logistic models for matched sets; the adjusted models included the same

four variables: race, history of hypertension, current cigarette

smoking, and education. For any use of PPA within three days (either for

cough-cold remedy or appetite suppression), the unadjusted odds ratio

was 1.67 (p=0.040) and the adjusted odds ratio was 1.49 (lower limit of

the one-sided 95% confidence interval (LCL)=0.93, p=0.084).

 

Our second co-equal specific aim was to estimate the association between

use of PPA and hemorrhagic stroke according to type of PPA exposure. The

results are shown in Table 4. (See Appendix C for exposures presented in

matched format.) For the association between PPA use in cough-cold

remedies within the three-day exposure window, the unadjusted odds ratio

was 1.38 (p=0.163). The adjusted odds ratio was 1.23 (LCL=0.75,

p=0.245). For the association between PPA use in appetite suppressants

within the three-day exposure window, the unadjusted odds ratio was

11.98 (p=0.007) and the adjusted odds ratio was 15.92 (LCL=2.04,

p=0.013).

 

To examine the relation between recency of PPA exposure and risk for

hemorrhagic stroke, we calculated odds ratios according to the timing of

most recent PPA use (Table 5). For this analysis, the pre-specified

definition for current use was use of any PPA-containing product on the

index day before focal time or the preceding calendar day. Prior use was

defined as use two or three calendar days before the focal time. As

shown in the table, the odds ratio was slightly higher for current use

(adjusted OR (AOR)=1.61, LCL=0.93, p=0.078) than for prior use

(AOR=1.16, LCL=0.47, p=0.393). Within current use, we then calculated

odds ratios according to first use or not-first use. First use was

defined as current use with no other use within the prior two weeks.

Not-first use included other uses within the two-week interval. The odds

ratio was higher for first use (AOR=3.14, LCL=1.16, p=0.029) than for

not-first use (AOR=1.20, LCL=0.61, p=0.329). All first uses of PPA

(n=13) reported in these data were in cough-cold remedies.

 

Our final co-equal specific aim was to estimate the association between

PPA and risk for hemorrhagic stroke among women for two exposure

definitions: appetite suppressant use within three days and first dose

use (See Appendix D). For the association between PPA in appetite

suppressants and risk for hemorrhagic stroke among women, the unadjusted

odds ratio was 12.19 (p=0.006) and the adjusted odds ratio was 16.58

(LCL=2.22, p=0.011). Among HSP subjects, all appetite suppressant use

within the three-day exposure window occurred among women. For first

dose PPA uses, eleven of the 13 exposures were among women (seven cases

compared with four controls). The unadjusted odds ratio was 3.50

(p=0.039) and the adjusted odds ratio was 3.13 (LCL= 1.05, p = 0.042).

Appendix C shows exposure data for cases and controls in a matched

format.

 

Additional Analyses

Based on the findings that risk for hemorrhagic stroke seemed to be

concentrated among current users, we examined the association between

current PPA dose and risk for hemorrhagic stroke. Among 21 exposed

control subjects, the median current dose of PPA (i.e., total amount

taken on the index day or preceding day) was 75mg. Analysis according to

dose shows that the odds ratio was higher for current doses above the

median (>75mg) (AOR=2.31, LCL=1.10, p=0.031) than for lower doses

(AOR=1.01, LCL=0.43, p=0.490). Among first dose users, four of eight

cases and two of five controls were exposed to >75mg of PPA.

 

To examine the potential effect of ambiguity in the correct focal time,

we recalculated the odds ratios after excluding all 154 case subjects

who were classified as having a definite (n=76) or uncertain (n=7

sentinel symptom. The magnitude of the adjusted odds ratios did not

change substantially (see Appendix E). We also recalculated the odds

ratios using the alternate focal time data for the 53 patients who, with

their matched control subjects, had a completed interview for this focal

time. Again, the adjusted odds ratios did not change substantially (see

Appendix F).

 

Description of Case and Control Subjects

Table 6 describes the 27 case subjects who reported an exposure to PPA

on their index day before focal time or during the preceding three days.

Among the 27 exposed case subjects, 21 (78%) were women and 6 (22%) were

men. Known risk factors for SAH or ICH (i.e., smoking, hypertension)

were present for 18 subjects (66%). About half of all exposures to PPA

occurred within 6 hours of the focal time, with a range of 6 minutes to

3.5 days and about half of all exposures (n = 17) involved slow release

preparations. The amount of PPA consumed at the last dose ranged from 7

mg to 150 mg (mean 82 mg). Eight of 27 last dose exposures (30%)

exceeded normal adult doses of 50 mg for immediate release preparations

and 75 mg for sustained release preparations(31).

 

Table 7 describes the 33 control subjects who reported an exposure to

PPA on their index day before focal time or during the preceding three

days. Among the 33 exposed case subjects, 20 (61%) were women and 13

(39%) were men. Known risk factors for SAH or ICH (i.e., smoking,

hypertension) were present for 18 control subjects (54%). One third of

all exposures to PPA occurred within 6 hours of the focal time, with a

range of 40 minutes to 3.2 days and about half of all exposures (n = 17)

involved slow release preparations. The amount of PPA consumed at the

last dose ranged from 12 mg to 150 mg (mean 61 mg). Five of 32 last

known dose exposures (16%) exceeded normal adult doses. In summary,

compared to PPA-exposed case subjects, exposed control subjects were

less likely to be female (61% compared with 78%), to have stroke risk

factors (54% compared with 66%), to consume PPA within 6 hours of the

focal time (33% compared with 52%), and to take excessive doses of PPA

(16% compared with 30%). Further details of PPA-exposed and non-exposed

subjects are presented in Appendix G.

 

 

 

DISCUSSION

In the Hemorrhagic Stroke Project of subjects between ages 18 and 49

years, PPA use was associated with an increased risk for hemorrhagic

stroke. The increased risk was evident for PPA used as an appetite

suppressant and for first use exposure. Because first uses of PPA in the

HSP always involved cough-cold remedies, the association of PPA with

risk for hemorrhagic stroke is present for both customary indications

for PPA (as a cough-cold remedy and an appetite suppressant). The

association of PPA with risk for hemorrhagic stroke was evident in the

overall group of case and control subjects and among women. Because so

few men were exposed to PPA in the HSP (n=19), we were unable to

determine if their risk for hemorrhagic stroke (with use of PPA) is

different from that of women.

 

Prior to the HSP, information concerning the association between PPA and

risk for hemorrhagic stroke in humans came from clinical trials(, case

reports(2), and one epidemiological study(20). Clinical trials on the

effectiveness of PPA are not informative on the risk of brain hemorrhage

because they did not enroll enough patients to observe occurrences of

this rare event. The one epidemiologic study found no association

between PPA and hemorrhagic stroke, but design limitations reduce its

contribution to the assessment of PPA safety. The only direct evidence

for an association between PPA and risk for hemorrhagic stroke in humans

has come from case reports. Although these reports effectively called

attention to the possible association between PPA and hemorrhagic

stroke, the absence of control subjects means that these studies cannot

provide evidence that meets the usual criteria for valid scientific

inference. The HSP, therefore, provides the most comprehensive and valid

evidence to date on the association between PPA and risk for hemorrhagic

stroke.

 

Other than a valid association between PPA and stroke, possible

alternative explanations for our findings include residual confounding,

bias in the design of the study, and chance findings in the analysis.

Residual confounding refers to incomplete adjustment for factors related

to both exposure and outcome. Although some residual confounding may be

present within the HSP, our data provide little support for the

presumption that it is significantly distorting the observed association

between PPA and hemorrhagic stroke. In particular, adjustment for known

potential confounders did not produce odds ratios that were markedly

different from the unadjusted figures (Tables 4 and 5).

 

Biases that most commonly affect the design of case-control studies

include temporal precedence bias, recall bias, and selection bias.

Temporal precedence bias refers to a systematic error in which an

exposure is counted although it occurs after the onset of the disease

under study, often in response to disease symptoms(30). We checked for

temporal precedence bias with specific analytic strategies for patients

with sentinel symptoms and found no evidence that it has an important

influence on the measured odds ratios.

 

Recall bias refers to the tendency of case subjects, compared with

control subjects, to have more or less complete recall of exposures or

to report exposures that may never have happened(32). Although it is

possible that our case subjects over-reported PPA exposure in an attempt

to explain their illness, in our data case subjects were no more likely

than control subjects to report exposures to aspirin or non-PPA oral

sympathomimetics. These negative findings suggest that recall bias was

not an important source of distortion in the HSP, especially considering

the well-publicized risk of hemorrhage with aspirin. It seems likely,

furthermore, that case-subjects in the HSP may have had impaired recall

of exposures due to brain injury. This impairment would have biased the

study toward a finding of no association between PPA and hemorrhagic

stroke. Our data provide some evidence for a bias in this direction

because case-subjects in the HSP more frequently reported " uncertain "

exposures to PPA than control subjects. When these " uncertain " exposures

are counted, the odds ratios increase by over 10% for all exposure

definitions except for first dose use. For first dose use the odds ratio

falls from 3.14 (p = 0.029) to 2.88 (p = 0.044). In addition, to address

differential recall stimulus we set the index day for control subjects

within one week of their interview (compared with up to 30 days for case

subjects). The close proximity of the index date and interview date for

control subjects was intended to overcome the imbalance in stimulus for

recall between controls (who experienced no major life event on the

index date) and cases (who did). Although recall bias is commonly

discussed in relation to case-control research, it rarely has a

meaningful impact on research findings(33, 34).

 

Selection bias refers to the preferential referral to a case-control

study of case or control subjects with (or without) the exposure under

study(35). Selection bias in the HSP might result from publicity about

PPA and stroke, leading physicians to preferentially seek a diagnosis of

hemorrhage in a PPA user or to preferentially refer a patient to the

study if there were a history of PPA use(36). We adopted several

strategies to reduce the possibility of selection bias, including active

case surveillance and enrollment of subjects who met our eligibility

criteria, regardless of the diagnostic impression of their physicians.

To check for possible selection bias, we compared eligible, enrolled

case-subjects with eligible, non-enrolled patients and found no

differences in age or ethnicity. Non-enrolled patients, however, were

more likely to be female and to have an ICH rather than a SAH. We did

not assess the eligible, non-enrolled patients for their ability to

communicate within 30 days; if we had, it is likely that many would have

been reclassified as non-eligible. Because ICH is more likely than SAH

to impair communication, this reclassification would be expected to

narrow the difference between eligible, enrolled and eligible,

non-enrolled patients in stroke type. Another potential source of

selection bias in the HSP is non-participation by eligible control

subjects who did not answer their telephone or who refused

participation. We do not know if such persons were more or less likely

to use PPA than the controls who agreed to participate.

 

Chance seems an unlikely explanation for our findings. Odds ratios for

appetite suppressant use and first use (all involving cough-cold

products) reached the conventional threshold for statistical

significance (p<0.05). Although the odds ratio for the association

between any PPA use in three days did not reach this threshold, chance

remains an unlikely explanation for the observed value of 1.49

(p=0.084).

 

Several features of the HSP support a valid association between PPA use

and risk for hemorrhagic stroke. First, in addition to the finding of

elevated odds ratios that reached statistical significance, the

magnitude of the odds ratios for PPA use as an appetite suppressant

(15.92) and as a first use (3.14) were large in magnitude even after

adjustment for important clinical features. Second, these data show an

association with brain hemorrhage across the two major formulations of

PPA (i.e., cough-cold remedy and appetite suppression).

 

There are potential limitations to our research. First, because we

excluded dead and non-communicative patients we cannot know if PPA is

more or less associated with brain hemorrhage in these patients compared

with those studied. Epidemiologists have long recognized a form of

research bias termed prevalence-incidence bias that may occur if

survivors of a potentially fatal illness have a different clinical

profile from those who die(37). This bias may either increase or reduce

the odds ratio associated with an exposure under investigation. In the

HSP, for example, if PPA confers a survival advantage in hemorrhagic

stroke, the odds ratio among survivors will be spuriously increased. If

PPA increases mortality, on the other hand, the odds ratio among

survivors will be spuriously reduced. Although we cannot be fully

certain that prevalence-incidence bias has not affected the HSP, we are

unaware of a plausible mechanism by which PPA might increase or decrease

the likelihood of survival after a stroke.

 

A second possible limitation of our study is the number of case and

control subjects who were exposed to PPA (n=60). This number of exposed

subjects limits the opportunity for subgroup analysis. It is worth

noting, however, that the exposure rate observed in women for the first

dose use analysis (0.533%) achieved the rate anticipated in the research

design (0.502%). As a final limitation, the HSP interviewers were not

blinded to the case-control status of subjects and some were aware of

the study purpose. Blinding would have provided extra protection against

unequal ascertainment of PPA exposure in case subjects compared with

control subjects. Instead of blinding, we relied on a highly structured

and scripted interview instrument. Overall, none of the limitations are

sufficiently important to invalidate our findings.

 

In conclusion, the results of the HSP suggest that PPA increases the

risk for hemorrhagic stroke. For both individuals considering use of PPA

and for policy makers, the HSP provides important data for a

contemporary assessment of risks associated with the use of PPA.

 

HSP INVESTIGATORS

Ralph I. Horwitz, M.D.

Harold H. Hines Jr. Professor of Medicine and Epidemiology

Yale University School of Medicine

New Haven, Connecticut

 

Lawrence M. Brass, M.D.

Professor of Neurology and Epidemiology and Public Health

Yale University School of Medicine

New Haven, Connecticut

 

Joseph P. Broderick, M.D.

Professor of Neurology

University of Cincinnati College of Medicine

Cincinnati, Ohio

 

Thomas Brott, M.D.

Professor of Neurology

Mayo Medical School

Rochester, Minnesota

 

Edward Feldmann, M.D.

Professor of Neurosciences Brown University School of Medicine

Providence, Rhode Island

 

Walter N. Kernan, M.D.

Associate Professor of Medicine Yale University School of Medicine

New Haven, Connecticut

 

Lewis Morgenstern, M.D.

Assistant Professor of Neurology and Epidemiology

The University of Texas Medical School and School of Public Health

Houston, Texas

 

Catherine M. Viscoli, Ph.D.

Associate Research Scientist Yale University School of Medicine

New Haven, Connecticut

 

Janet Lee Wilterdink, M.D.

Associate Professor of Neurosciences Brown University School of Medicine

 

Providence, Rhode Island

 

SCIENTIFIC ADVISORY GROUP MEMBERS

Louis C. Lasagna, M.D. (Chair)

Dean, Sackler School of Graduate Biomedical Sciences Professor of

Pharmacology and Experimental Therapeutics

Tufts University

 

J. P. Mohr, M.D.

Sciarra Professor of Neurology

College of Physicians and Surgeons Columbia University

 

Samy Suissa, Ph.D.

Professor of Medicine and Epidemiology and Biostatistics

McGill University

 

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_________________

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