COUPLES AND FAMILIES

Frequency, severity, and evolution of split family alliances: Howobservable are they?

CRISTINA MUNIZ DE LA PENA1, MYRNA FRIEDLANDER1, & VALENTIN ESCUDERO2

1Department of Educational and Counseling Psychology, University at Albany, State University of New York, Albany, NY,

USA & 2Department of Psychology, Universidad de La Coruna, La Coruna, Spain.

(Received 7 April 2008; revised 3 September 2008; accepted 6 September 2008)

AbstractSplit alliances (within-family differences in the emotional bond with the therapist) were studied in 19 U.S. and 21 Spanishfamilies using the System for Observing Family Therapy Alliances (SOFTA; Friedlander, Escudero, & Heatherington,2006). Examining individual family members’ scores on the corresponding self-report and observational EmotionalConnection to the Therapist SOFTA scales, the authors identified mild, moderate, and severe split alliances. In bothsamples, self-reported splits occurred frequently and with almost all of the therapists. Although clients’ observedinteractions with the therapist often mirrored their self-reports, family members’ perceptions of the therapeutic bond weregenerally more discrepant than their behavior suggested. The majority of families that dropped out had a moderately orseverely split alliance in at least one session.

Keywords: couples and family systems therapy; psychotherapist training/supervision/development

It is not uncommon for clients to feel like ‘‘therapy

hostages’’ (Friedlander, Escudero, & Heatherington,

2006, p. 201) when family members make threats,

implicit or explicit, about their participation in

therapy: ‘‘If you don’t go to therapy, I’ll leave you/

divorce you/send you to a group home/bring charges

against you/expel you from school/cut off all finan-

cial support’’ and so on. Not surprisingly, such

behind-the-scenes negotiations do not bode well

for forming a strong working alliance. The conflict

over the need for treatment can easily turn into an

argument about the therapist’s style, personality, or

competence.

For this reason, it is challenging to create strong

bonds with families in conflict when one or more

people attend the sessions under duress. Little is

known, however, about whether differences in family

members’ emotional investments can reliably be

observed and whether these differences are mean-

ingful indicators of a therapy in trouble.

Compared with the abundant literature in indivi-

dual therapy, there has been relatively little research

on the working alliance in conjoint couple and family

therapy. As Pinsof and Catherall (1986) first noted,

there is no ‘‘single’’ family alliance. Each person has a

unique relationship with the therapist and also

observes every other family member interacting

with the therapist. To complicate matters further,

there are also two kinds of within-system alliances

(Pinsof, 1994, 1995): (a) the alliance between and

among family members with regard to the therapy and

(b) alliances within the institution (e.g., between

cotherapists, between therapist and supervisor, within

therapy teams). As Pinsof (1995) pointed out, ‘‘Each

alliance level influences, through mutual causality,

every other level and locus . . . [and] differential

causality asserts that some levels of the alliance may

play a more critical role than others’’ (p. 67).

To study the strength of multiple alliances, Pinsof

and Catherall (1986) developed the Couple and

Correspondence concerning this article should be addressed to Myrna Friedlander, Department of Educational and Counseling

Psychology, ED220, University at Albany, State University of New York, Albany, NY 12203, USA. E-mail: mfriedlander@uamail.

albany.edu

Psychotherapy Research, March 2009; 19(2): 133�142

ISSN 1050-3307 print/ISSN 1468-4381 online # 2009 Society for Psychotherapy Research

DOI: 10.1080/10503300802460050

Family Therapy Alliance Scales (CTAS and FTAS),

which ask clients to evaluate their own alliance with

the therapist in terms of goals, tasks, and bonds as

well as the alliances of other family members and the

family as a group in relation to the therapist. The

revised versions of the measures (CTAS-R and

FTAS-R; Pinsof, 1999; Pinsof, Zinbarg, & Kno-

bloch-Fedders, 2008) include a fourth dimension,

the within-system alliance, which asks clients to

report on the alliance among family members (i.e.,

apart from the therapist).

Research with the CTAS and FTAS, as well as

with other measures and methods (Beck, Friedlan-

der, & Escudero, 2006; Bennun, 1989; Bourgeois,

Sabourin, & Wright, 1990; Knobloch-Fedders, Pin-

sof, & Mann, 2004; Robbins, Turner, Alexander, &

Perez, 2003; Robbins et al., 2006, 2008; Symonds &

Horvath, 2004), has consistently found that the

relationship between self-reported alliances and

therapeutic outcome is not straightforward. Family

members often perceive the therapy process quite

differently, and differences in perceptions have con-

sequences for treatment completion. Results have

shown, for example, that discrepancies in parents’

and adolescents’ alliances put families at greater risk

for dropout than either family member’s alliance

with the therapist considered in isolation (Robbins

et al., 2003). In one study of couples (Quinn,

Dotson, & Jordan, 1997), therapy was most success-

ful when wives’ perceptions of the alliance with the

therapist exceeded those of their husbands, whereas

another couples study (Symonds & Horvath, 2004)

found the opposite pattern. To complicate matters

further, in the latter study similarity in the partners’

alliances with the therapist, whether favorable or

unfavorable, was more predictive of treatment

success than either partner’s rating of the alliance

alone.

How, then, can we best capture the alliance

empirically? Arguably, adding or averaging family

members’ alliance scores to arrive at a ‘‘total’’ family

score is misleading because doing so obscures in-

dividual clients’ scores. Yet how much of a discre-

pancy in family members’ scores is meaningful?

Addressing this question, Heatherington and Fried-

lander (1990) estimated the frequency of what Pinsof

and Catherall (1986) called split alliances by defining

a split in terms of standard deviation units on one

subscale (Self with Therapist) of the CTAS or FTAS.

When the split was defined as family members having

scores that were 1 SD apart, 43% of couples and 75%

of families seen in a U.S. hospital clinic had a split

alliance. Based on a more conservative decision rule

(i.e., a 2 SD difference in scores), 14% of the couples

and 42% of the families had split alliances. More

recently, Mamodhoussen, Wright, Tremblay, and

Poitras-Wright (2005) reported that 32% of couples

in a French-speaking university clinic in Quebec had

a 1 SD split alliance and 13.3% had a 2 SD split.

Similarly, Knobloch-Fedders et al. (2004) found that

roughly 40% of couples in a large outpatient agency

had a split alliance in either Session 1 or Session 8.

Despite differences in the three samples and settings,

it is reasonable to conclude that split alliances are

common occurrences.

With few exceptions, the assessment of split

alliances has relied solely on clients’ self-reported

perceptions. Robbins et al. used selected items from

the observational Vanderbilt Therapeutic Alliance

Scale-Revised (VTAS-R; Diamond, Liddle, Hogan,

& Dakof, 1999) to estimate the therapeutic alliance

in functional family therapy (FFT) for adolescent

behavior problems (Robbins et al., 2003) and in

multidimensional family therapy (MDFT; Robbins

et al., 2006) and brief strategic family therapy

(BSFT; Robbins et al., 2008) for adolescent drug

use. Compared with completer families, families

that dropped out of FFT and BSFT demonstrated

more ‘‘unbalanced’’ alliances (Robbins et al., 2003,

2008). In the MDFT study (Robbins et al., 2006),

mothers exhibited significantly better working rela-

tionships with the therapist than did their sons, and

when both clients’ alliance scores decreased from

Session 1 to Session 2, the family was likely to

terminate prematurely.

Although significant correlations have been re-

ported between family members’ perceptions of the

alliance and their in-session behavior (Escudero,

Friedlander, Varela, & Abascal, 2008; Friedlander,

Escudero, Horvath, et al., 2006), these group studies

have little to say about whether, within a given

family, split alliances can be detected behaviorally.

To date, only one study has considered the degree to

which discrepancies in family members’ self-re-

ported alliances are reflected in observable behavior.

Beck et al. (2006) analyzed four family therapy cases

seen at a university clinic in Spain. Behavioral

manifestations of clients’ emotional bonds with

therapists were rated using the Spanish version of

the System for Observing Family Therapy Alliances

(SOFTA; Escudero & Friedlander, 2003). Results

showed that two of the four families had mildly split

alliances in their postsession perceptions (i.e.,

FTAS-R Self-with Therapist scores), but the splits

were minimally reflected in the behavioral ratings.

These findings suggest that mildly split alliances may

not be readily observable.

The correspondence, or lack thereof, of observers’

and clients’ scores has long been of interest to

psychotherapy researchers (e.g., Tichenor & Hill,

1989), and it is generally agreed that clients’ scores are

more predictive of treatment success (e.g., Horvath &

134 C. M. de la Pena et al.

Bedi, 2002). More so than in individual therapy,

clients in conjoint therapy may be unwilling to show

their private feelings in session, when other family

members are closely observing their behavior. Some

clients hide their reactions because of the power

differential in the family or because they cannot risk

showing their true feelings in a therapy context. A

reluctant adolescent, for example, may simply pay lip

service to the therapist in order to appease his parents.

In other words, because what happens, or does not

happen, in couple and family therapy can have

immediate interpersonal consequences, many clients

feel compelled to disguise their true responses (posi-

tive as well as negative) to the therapist.

In the present study, we began by examining the

frequency and severity of split alliance sessions. Our

primary question was, can observers detect split

alliances? That is, how often are discrepancies in

family members’ perceptions of the therapist re-

flected in their observable, in-session behaviors?

Are only severe splits readily observable? A second-

ary question was how split alliances evolve over

time in terms of severity and whether families with

severe splits are likely to drop out of treatment

prematurely.

These research questions were studied in two

existing data sets with (a) low-income families in a

nonprofit community clinic in the United States

and (b) middle-income families in a Spanish uni-

versity training clinic. We anticipated that our

results, if consistent across these diverse samples,

would have implications not only for future family

process research but also for practice. Because

mildly split alliances seem to be commonplace

(Beck et al., 2006; Heatherington & Friedlander,

1990; Knobloch-Fedders et al., 2004; Mamodhous-

sen et al., 2005), it may be easy for a therapist to

miss the signs of a poorly developing relationship

with a quiet family member. If only severely split

alliances can be detected in client behavior, it may

be too late for the therapist, who needs to rely on

observable behavior, to repair the split before the

family drops out. To build theory about how split

alliances can effectively be corrected, we first need

to know whether they can be observed, that is,

whether discrepancies in family members’ self-

reported feelings toward the therapist match their

behavior and, if so, whether the observed within-

family discrepancies are more or less extreme than

the self-reported ones. We also need to know how

often splits remain unchanged, improve, or deterio-

rate over time in treatment and how often they

result in premature termination.

Method

Participants

U.S. sample. Data for this study were drawn from

an existing data set of 29 low-income families who

volunteered to participate in a study on the effec-

tiveness of family therapy (Friedlander, Lambert, &

Muniz de la Pena, 2008). Clients were seen for a

maximum of 10 free conjoint sessions (M�8.8,

SD�2.1). All families had at least one school-age

child at risk, and presenting concerns included

family conflict; children’s conduct problems, poor

school performance, or sexual abuse; and parental

death, separation/divorce, substance abuse, or in-

carceration. The majority of families were headed by

single parents, and most families had fairly young

children (mean age�10.2 years, SD�3.69). For the

present study, we used data from 19 of the 29

families (i.e., all those that were seen for at least

three sessions in which two or more family members

completed the self-report alliance measure). (Chil-

dren younger than 12 years did not provide self-

report data.) Participating clients ranged in age from

12 to 58 years (M�28.02, SD�12.80).

The 10 therapists in the full sample (seven White

women, one Latina, two White men) had a wide age

range (23�59 years, M�41.3, SD�11.39) and

range of clinical experience (9 months to 23 years;

M�6.8, SD�6.64). Nine of these 10 therapists saw

families in the present subsample; four therapists

were graduate students, and the remainder were

either master’s-level (n�4) or doctorate-level (n�1)

clinicians. Various theoretical approaches were re-

presented, including family systems (n�4), eclectic

(n�2), psychodynamic (n�1), client centered (n�1), and cognitive�behavioral (n�1).

Spanish sample. Data from a sample of 37

mostly middle-income couples (n�16) and families

(n�21), seen for an average of 7.6 sessions (SD�3.1), came from a recent study of the relationship of

alliance to therapeutic progress (Escudero et al.,

2008). Only the 21 families (i.e., none of the

couples) were included in the present study. Pre-

senting problems included family conflict, problems

with communication, parenting concerns, separa-

tion/divorce, and parental death. All of the partici-

pating clients were White Spaniards, aged 12 to 72

years (M�35.3, SD�17.53).

The families were seen by six White therapists

(three women, three men) ranging in age from 30 to

50 years (M�40, SD�7.57) and in experience

ranging 4 to 20 years (M�11.2, SD�7.26). Two

therapists were doctorate-level and four were mas-

ter’s-level psychologists. The predominant theoreti-

cal orientation was an integrative model of brief

Split alliances 135

systemic therapy that included structural, strategic,

and solution-focused elements.

System for Observing Family Therapy

Alliances

SOFTA-o. The transtheoretical SOFTA-o was

created simultaneously in English (Friedlander

et al., 2004; Friedlander, Escudero, & Heathering-

ton, 2006; Friedlander, Escudero, Horvath, et al.,

2006) and Spanish (Sistema de Observacıon de

la Alianza Terapeutica en Intervencıon Familiar;

Escudero & Friedlander, 2003) to assess four

behavioral aspects of the therapeutic alliance in the

context of couple and family therapy. Two dimen-

sions reflect Bordin’s (1979) goals, tasks, and bonds:

Engagement in the Therapeutic Process and Emo-

tional Connection to the Therapist (EC). The other

two dimensions, Safety Within the Therapeutic

System and Shared Sense of Purpose Within the

Family, reflect aspects of the alliance that are unique

to conjoint treatment.

In the present study, only the clients’ EC dimen-

sion was of interest because, according to authors of

the SOFTA (Friedlander, Escudero, & Heathering-

ton, 2006), this dimension best captures ruptures in

clients’ individual relationships with the therapist.

Moreover, the EC dimension is the one most similar

to the FTAS self with therapist alliance, which was

used in previous research on split alliances (e.g.,

Beck et al., 2006; Heatherington & Friedlander,

1990). In the SOFTA, the emotional connection is

operationally defined as

Viewing the therapist as an important person in

the client’s life, almost like a family member; a

sense that the relationship is based on affiliation,

trust, caring, and concern; that the therapist

genuinely cares and ‘‘is there’’ for the client, that

he or she is on the same wavelength with the

therapist (e.g., similar life perspectives, values)

and that the therapist’s wisdom and expertise are

valuable. (Friedlander, Escudero, & Heathering-

ton, 2006, p. 88)

Using the SOFTA-o, trained observers make

inferences about clients’ perceptions of the alliance

from their observed interactions in either videotaped

or live couple and family therapy sessions. To use

this rating system, trained judges observe a session

and tally the frequencies of specific, operationally

defined positive and negative alliance-related beha-

viors (see later examples) in each of the four

dimensions as they occur. Based on these tallies,

each dimension is then rated from �3 (extremely

problematic) to 3 (extremely strong; with 0�unremark-

able or neutral). To go from the tallies to the global

ratings, judges consult the operational definitions

and rating guidelines in the training manual (Fried-

lander et al., 2004) and consider the valence

(positive or negative), frequency, intensity, and

clinical meaningfulness of the observed behaviors.

According to the guidelines, the rating must be

between 1 and 3 when only positive behaviors are

exhibited and between �1 and �3 when only

negative behaviors are exhibited; ratings must be

between �1 and 1 when a client exhibits both

positive and negative behaviors. In the EC dimen-

sion, 10 verbal and nonverbal behaviors are repre-

sented, including ‘‘Client expresses interest in the

therapist’s personal life,’’ ‘‘Client shares a joke or

lighthearted moment with the therapist,’’ and ‘‘Cli-

ent mirrors the therapist’s body posture.’’ The

negative behaviors include ‘‘Client avoids eye con-

tact with the therapist,’’ ‘‘Client is reluctant or

refuses to answer the therapist,’’ and ‘‘Client has

hostile or sarcastic interactions with the therapist.’’

The authors of the SOFTA-o (Friedlander, Escu-

dero, & Heatherington, 2006) reported intraclass

reliabilities ranging from .72 to .95 across several

studies.

SOFTA-s. The self-report SOFTA-s (Friedlander,

Escudero, & Heatherington, 2006) has four subscales

that correspond closely to the operationally defined

SOFTA-o dimensions. The 16 items, both positively

and negatively worded, are rated on 5-point Likert

scales (1�not at all and 5�very much). The four

items on the EC scale are ‘‘The therapist understands

me,’’ ‘‘The therapist is doing everything possible to

help me,’’ ‘‘The therapist has become an important

person in my life,’’ and ‘‘The therapist lacks the

knowledge and skills to help me.’’ With the last item

reverse scored, high EC scores indicate a strong

personal bond with the therapist (range�4�20).

In previous research, the SOFTA-s has been reli-

able (Cronbach alphas�.62�.87) and significantly

associated with SOFTA-o behavioral ratings (Fried-

lander, Escudero, & Heatherington, 2006). In the

present samples, the EC internal consistency reliabil-

ities were .76 (United States) and .67 (Spanish).

Dropout. The U.S. families were offered a max-

imum of 10 conjoint family sessions, and the

Spanish families were offered eight sessions, with

the option to continue for an additional eight

sessions on the therapist’s recommendation. We

defined dropout in both samples as premature

termination (i.e., when a family unilaterally chose

to discontinue treatment for any reason, usually by

failing a scheduled appointment and not returning).

136 C. M. de la Pena et al.

Procedure

Adult clients and adolescents older than age 12

completed the SOFTA-s immediately after Sessions

3, 6, and 9 in the U.S. sample. In the Spanish

sample, the measure was administered only after

Sessions 3 and 6 because very few Spanish families

had nine or more sessions.

Four graduate students in the United States (two

doctoral and two master’s students in counseling

psychology), all White women, and three research

assistants in Spain were trained in the SOFTA-o for

more than 20 hr using practice videotapes until 90%

of their ratings varied by 1 point or less. The U.S.

judges were then split into two teams, each person

independently rating half the final sample from

videotapes. All three Spanish raters independently

rated a randomly selected subsample (30%) of

sessions, and the remaining sessions were rated by

a single judge. Both teams negotiated discrepancies

to consensus.

As is usual with the SOFTA-o, the judges rated

the behavior of all members of the family simulta-

neously. The judges were aware of the session

numbers (3, 6, or 9) but not the present research

questions, because the ratings were completed be-

fore this study was devised. In the larger U.S. sample

from which the present data were drawn, the judges

achieved a mean intraclass correlation (ICC) of .86

on the EC dimension. In the Spanish sample, the

judges achieved a mean ICC of .76.

Results

To begin, we calculated the EC means and standard

deviations for both the SOFTA-s and the SOFTA-o

separately for each sample and for each session (see

Table I). The distributions of self-reported EC scores

showed that clients in both samples perceived the

emotional bonds with their therapists quite favorably.

Notably, the Spanish sample scored significantly

higher and reported less variability than the U.S.

sample in both Session 3, t(97)�3.96, pB.0001,

and Session 6, t(68)�4.22, pB.0001.

The observational EC ratings were similar in the

two samples (see Table I), with means close to 0,

which indicates that, on average, the clients exhib-

ited unremarkable to moderately positive emotional

connections with their therapists. However, compar-

isons showed that, in Session 6, the Spanish clients

demonstrated relatively more positive and less vari-

able emotional connection behavior (M�0.42,

SD�0.67) than did the U.S. clients (M��0.04,

SD�1.01), t(83)�2.33, pB.02.

Self-Reported Splits

Identification. Split alliances were defined as ses-

sions in which two or more family members had

notably discrepant SOFTA-s scores. Forty-nine ses-

sions from the U.S. sample and 31 sessions from the

Spanish sample were examined for splits. (Because

of missing self-report data and because a number of

the U.S. children were too young to complete the

postsession members, some sessions could not be

examined for split alliances.)

As was done in other split alliance studies, we

identified split alliances based on standard deviation

units. However, going beyond the prior research, we

also took into account the sample means, which

allowed us to assess the degree of severity in the

families’ split EC scores. Mild splits were operation-

ally defined as occurring when two or more family

members’ self-reported EC scores differed by at least

1 SD and all clients scored at or above the respective

sample mean. Moderate splits occurred when two

family members’ EC scores differed by 1 SD and at

least one client scored below the sample mean.

Severe splits occurred when there was a difference

in standard deviation of 2 or more between or among

family members, with at least one client below the

sample mean.

Table I. Means and Standard Deviations of Self-Reported and Observed Emotional Connection Scores

U.S. sample Spanish sample

Variable M SD M SD p

SOFTA-s

Session 3 14.94 3.13 17.16 2.40 .0001

Session 6 14.82 4.20 18.27 1.72 .0001

Session 9 15.09 3.39 * * *

SOFTA-o

Session 3 0.27 1.03 0.22 0.71 �.05

Session 6 �0.04 1.01 0.42 0.66 .022

Session 9 0.35 1.07 * * *

Note. None of the Spanish families had nine sessions. SOFTA�System for Observing Family Therapy Alliances; s�self-report; o�observed. SOFTA-s range: 4�20; SOFTA-o range��3 to 3.

Split alliances 137

Because most clients completed the SOFTA-s at

least twice, we avoided redundancy in the data by

using separate sample means and standard devia-

tions for each session. Thus, by rounding to the

nearest whole numbers, we used M�15 as a

decision rule for all three sessions with the U.S.

sample, but we used SD�3 for Sessions 3 and 9 and

SD�4 for Session 6 (see Table I). For the Spanish

sample, we used SD�2 for both sessions but M�17

for Session 3 and M�18 for Session 6.

Frequency and severity. As operationalized by the

self-report data, eight of the nine (89%) U.S.

therapists and five of the six (83%) Spanish thera-

pists had one or more families with a split alliance

session. At least one self-reported split session was

reported by 14 of the 19 (73.6%) U.S. families and

by 14 of the 21 (66.67%) Spanish families. Adoles-

cents’ scores were lower than their parents’ in 50%

of the U.S. families and 35.7% of the Spanish

families. In both samples, a parent occasionally had

the lower score; in several cases, the family member

reporting the lower score varied from one session to

another.

The self-reported splits occurred most often in

Session 3 (in 52.6% of U.S. families and 57.1% of

Spanish families). In the U.S. sample, four families

(21%) had one split session, eight (42%) had two

split sessions, and two (10.5%) had three split

sessions. In the Spanish sample, eight families

(38.1%) had one split session and six (28.57%)

had two split sessions.

Table II shows the frequencies of split alliances by

level of severity. For the U.S. sample, 26 sessions had

self-reported splits; six (23%) were mild, eight

(31%) were moderate, and 12 (46%) were severe.

Twenty (45%) of the Spanish family sessions had

self-reported splits, of which five (25%) were mild,

eight (40%) were moderate, and seven (35%) were

severe.

Consistency and Congruence of Observer

Ratings

Consistency. After identifying sessions in which

either a mild, moderate, or severe split was present

in the self-reported EC scores, we examined the

observers’ EC ratings of those same sessions. First,

we examined the consistency of the SOFTA-o

ratings of the family members who had self-reported

splits. To consider the observer ratings to be a

consistent reflection of the split reported by the

clients, the same two (or more) family members

needed to have at least mildly discrepant observer

ratings (as defined later) in the same direction. For

example, if the mother’s self-reported SOFTA-s

score were 2 SDs higher than her daughter’s, for

the observed split to be considered consistent the

mother’s SOFTA-o behavioral rating needed to be at

least 1 SD higher than the daughter’s. If the mother’s

and daughter’s SOFTA-o ratings were identical, or if

the daughter’s behavior was rated higher than the

mother’s, the session was not counted as a consistent

observational split. In other words, we were only

interested in observational splits that matched the

self-reported splits.

As shown in Table II, most of the self-reported

splits in the U.S. sample (62%) were consistently

detected by the SOFTA-o, particularly those that

were moderate or severe. For the Spanish families,

40% of the self-reported splits were consistent with

behavior, predominantly those that were moderate

or severe.

Severity congruence. Corresponding to the opera-

tionalization of severity in SOFTA-s scores, we used

the SOFTA-o sample mean (0) and standard devia-

tion (1) to identify the degree of severity in the

observed splits. Thus, observed splits were consid-

ered mild when there was at least a 1-point difference

in two or more family members’ SOFTA-o ratings,

with no ratings below 0. Moderate splits were defined

Table II. Consistency and Congruence of Self-Reported and Observed Split Alliances

U.S. families Spanish families

Variable Self-report

Consistent

w/self-reporta

Congruent

in severityb Self-report

Consistent

w/self-reporta

Congruent

in severityb

Mild 6 .33 .17 5 .20 .20

Moderate 8 .63 .13 8 .50 .13

Severe 12 .75 .42 7 .43 .29

Total 26 .62 .27 20 .40 .20

Note. There were 49 total sessions (from 19 families) in the U.S. sample, 31 total sessions (from 21 families) in the Spanish sample, in which

at least two family members provided self-reported alliance ratings on the System for Observing Family Therapy Alliances�self-report

(SOFTA-s). Self-reported splits were defined based on clients’ SOFTA-s Emotional Connection scores.aProportions of self-reported split sessions that matched the observational (SOFTA-o) ratings in terms of which family members had higher

or lower Emotional Connection scores. bProportions of self-reported split sessions that were not only consistent but were also at the same

level of severity (i.e., mild, moderate, or severe) on the SOFTA-o.

138 C. M. de la Pena et al.

as a 1-point difference in SOFTA-o ratings, with at

least one family member below 0. Severe splits

required a difference of at least 2 points in family

members’ ratings, with at least one family member

below 0.

For all of the ‘‘consistent’’ sessions (those in which

the same family members had at least a mildly split

alliance on both measures), we determined the

degree of severity congruence. In other words, a

congruent split was defined as occurring when the

observed split was at the same level of severity as the

self-reported split (i.e., a mild self-reported split was

also observed as mild, a moderate split was observed

as moderate, and a severe split was observed as

severe). Thus, for example, if a mother and daughter

had SOFTA-o behavioral ratings that were 2 points

apart (mother�1, daughter��1), with one rating

below 0, this severe observed split would be con-

gruent with the severe self-reported split (SOFTA-s

scores: mother�18, daughter�10). On the other

hand, the observed split would be considered non-

congruent if the discrepancy in SOFTA-o ratings

was either mild (mother�1 and daughter�0) or

moderate (mother�0, daughter��1) rather than

severe.

Table II shows the severity congruence of self-

reported and observed splits. Congruence was gen-

erally poor in both the Spanish (20% of sessions)

and the U.S. (27%) samples. However, in both

samples, congruence was most likely for the severe

splits (42% of the severe U.S. sessions and 29% of

the severe Spanish sessions).

We also examined the noncongruent splits (not

tabled) to determine in what way the observed splits

differed in severity from the self-reported splits.

Notably, 75% of the noncongruent observed splits

in both samples were less severe than the self-

reports. In other words, when the observer ratings

were not congruent with the self-reported alliance

scores in level of severity, family members’ self-

reported alliances were generally more discrepant

than their behavior indicated.

Evolution of Self-Reported Splits

Table III indicates the number of families with

unchanged, improved, and deteriorated patterns of

split alliances, determined by examining levels of

severity over time. We defined an unchanged pattern

as one in which the level of severity of the self-

reported split remained constant from Session 3 to 6

(or, in the U.S. families, from Session 6 to 9). An

improved pattern was one in which the split became

less severe over time (i.e., severe 0 moderate, mild,

or no split; moderate 0 mild or no split). A

deteriorated pattern was one in which the split

worsened (i.e., mild 0 moderate or severe; moder-

ate 0 severe) over time. We only considered families

in which the same clients completed the SOFTA-s in

two or more sessions (ns�14 U.S. families, six

Spanish families).

Results showed that the majority of U.S. families

had split alliance patterns that either improved or did

not change, with relatively few deteriorating splits.

That is, 7 (50%) U.S. families’ splits improved in

level of severity over time, four (28.5%) remained

unchanged, and three (21.4%) deteriorated. None of

the Spanish families remained unchanged; 50%

improved and 50% deteriorated.

Relation of Splits to Premature Dropout

In the U.S. sample, six of the 19 total families

(31.57%) dropped out, four of which had either a

moderate or severe split alliance in at least one

session. The remaining 10 families with split alli-

ances did not drop out, and none of the families with

only a mild split dropped out.

In the Spanish sample, four of the 21 total families

(19%) dropped out. All four of the noncompleter

families had self-reported splits; three (75%) were

moderate or severe, and one (25%) was mild. The

remaining 10 families with split alliances did not

terminate treatment prematurely.

Discussion

Discrepancies in family members’ privately held

feelings about the therapist were reported by the

majority of families in both of our samples. In terms

of the degree to which these splits could be detected

in observable behavior, 62% of the self-reported

U.S. splits were mirrored in client behavior (e.g., a

Table III. Patterns of Self-Reported Split Alliances Over Time

Direction of change U.S. Families Spanish families

Unchanged

Mild 1 (.07) 0 (.00)

Severe 3 (.21) 0 (.00)

Improved

Moderate � 2 (.14) 2 (.33)

Severe � 5 (.36) 1 (.17)

Deteriorated

Mild ¡ 1 (.07) 1 (.17)

Moderate ¡ 2 (.14) 2 (.33)

Note. Values represent numbers of families, with proportions in

parentheses. Ns�14 U.S. families, six Spanish families. Only

families with the same individuals completing the System for

Observing Family Therapy Alliances�self-report twice are in-

cluded. Families with only one split alliance session were not

counted unless their pattern improved (i.e., movement from

reporting a split to reporting a nonsplit alliance).

Split alliances 139

mother’s score at least 1 SD higher than her

daughter’s score on both the self-report and obser-

vational measures). However, only 27% of the

observed splits were congruent in terms of severity

with the self-reported splits. The majority of the

congruent splits were severe, indicating that family

members who felt very differently about the therapist

often showed this extreme difference in their beha-

vior. Results were somewhat more modest in the

Spanish sample, however, in which only 40% of the

splits matched the in-session behavior, with only

20% of the splits congruent in severity. Importantly,

in both samples the noncongruent splits occurred

generally because individual family members’ self-

reported bonds with the therapist were more dis-

crepant (i.e., severe) than their in-session behavior

suggested. These results, which are consistent with

Beck et al.’s (2006) case studies, suggest that, in a

given session, differences in the quality of family

members’ bonds with the therapist may be detect-

able but the extremity of clients’ feelings tends not to

be readily apparent.

Looking at where the splits occurred, we found

that the majority of split alliances took place early in

therapy (i.e., the third session, when the quality of

the therapeutic relationship is still developing).

Notably, it was not consistently either the adolescent

or the parent(s) whose EC score was the lowest in

the family. In terms of changes over time, only three

families with mild splits in Session 3 had a deterior-

ating pattern, and a fair proportion of the moderate

and severe splits improved with time (43% of U.S.

families, 50% of Spanish families). Consistent with

this finding, split alliances did not invariably lead to

dropout; in both samples, the majority of families

with split alliances did not terminate treatment

prematurely. Notably, though, dropout occurred

most often in families whose splits were relatively

more severe.

Consistent with alliance research in general, the

present clients’ self-reported emotional connections

to their therapist were, on average, fairly (U.S.

sample) to very (Spanish sample) strong, whereas

the average observed (behavioral) EC ratings in both

samples were only moderate. These results are

consistent with previous findings (Shelef, Diamond,

Diamond, & Little, 2005) that clients reported

feeling more positively about their personal relation-

ships with the therapist than their behavior indi-

cated. It may be that in session some clients do not

want to show how much they like the therapist,

especially if they were coerced into treatment by

other family members. Alternately, it is possible that

self-reported alliances are inflated as a result of social

desirability (Shelef et al., 2005), although our clients

knew that their therapists would not have access to

the questionnaire data.

The relatively more positive mean scores on the

SOFTA-s tell only a part of the story, however.

Without examining each family individually, we

might overlook the fact that a fair number of families

in both samples had moderately or severely split

alliances in at least one session. This finding, which is

consistent with three prior studies (Beck et al., 2006;

Heatherington & Friedlander, 1990; Mamodhoussen

et al., 1995), is notable given the cultural, linguistic,

contextual, and demographic differences across sam-

ples. The present U.S. families, all low income, were

seen in a not-for-profit community clinic, whereas the

Spanish clients, seen in a university training center,

tended to be middle class and somewhat older. The

Spanish therapists had relatively more experience and

consistently used a brief systemic approach, whereas

the U.S. therapists were less experienced and notably

more eclectic in orientation. Whether the somewhat

larger proportion of severe splits in the U.S. sample

was due to any of these differences is not possible to

discern. Nonetheless, the similar results across sam-

ples are of importance both theoretically and

practically.

In general, our results are consistent with previous

reports, suggesting that variability in family mem-

bers’ alliance perceptions is commonplace. How-

ever, the present study improved on the earlier

investigations in several ways. This was the first

research to assess the degree of consistency between

self-reported and behavioral split alliances. Like

prior researchers (Beck et al., 2006; Heatherington

& Friedlander, 1990; Knobloch-Fedders et al.,

2004; Mamodhoussen et al., 2005), we operationally

defined a split in terms of standard deviation units,

but we also took into account the relative strength of

clients’ perceived bonds with the therapist (i.e.,

where each family member’s SOFTA-s score was in

relation to the sample mean). This methodology

allowed us to determine congruence in level of

severity, a characteristic that seems to be meaningful

for comparing self-reported and observed splits and

for identifying patterns of split alliances over time.

Use of our methods in future research can permit a

close examination of what actually takes place in

sessions when moderate and severe splits improve or

when they result in premature dropout. We also need

to consider clients’ ratings on all four SOFTA

dimensions in order to determine whether emotional

connection to the therapist is a unique predictor of

outcome. It may well be that, despite having a split

alliance with the therapist, families that stay in

treatment are highly engaged in problem solving,

find the therapeutic setting to be a safe environment,

and share a common vision for family change.

140 C. M. de la Pena et al.

There are at least three explanations for the fair

number of self-reported split alliances that were not

detected by the behavioral ratings. First, there is an

important methodological difference between self-

reported and observed alliances: When asked to

complete an alliance questionnaire, clients consider

their cumulative experience of the therapist, whereas

observers only rate what they see in a single session.

Second, the items in these two measurement formats

differ, despite being based on the same operational

definitions. Whereas clients report on their global

impressions of the therapist, the observational ratings

are more conservative estimates of the alliance. That

is, to maximize interrater reliability, only clearly

evident behavioral manifestations of the alliance are

recorded, and subtle signs of positive and negative

emotional connection are not tallied. Third, clients’

feelings about the therapist may well be the result of

covert processes that could not be captured by any

behavioral rating system. That is, emotional bonds

are not always created through direct communication

but rather through more complex, indirect interac-

tion. Because each family member observes the

therapist interacting with all the other family mem-

bers, a mother may feel a bond with the therapist not

because of how the therapist treats her but because

she sees the therapist showing support and concern

for her troubled child.

Another limitation has to do with the fact that

although EC scores estimate each client’s individual

bond with the therapist, in any family context,

individuals’ thoughts, emotions, and behaviors are

interdependent (Bolger & Shrout, 2007). To avoid

this problem, it would be preferable to analyze data

from individual clients within families, within ses-

sions, and within therapists, but finding large enough

samples for a hierarchical linear analysis can be

problematic. The modest overlap between self-re-

port and observer scores suggests that researchers

need both sources of information on the alliance to

fully understand what is taking place in conjoint

couple and family therapy. One implication for

future research has to do with the problem of

designating an appropriate unit of analysis. Our

findings indicate that it is worthwhile to define a

family-level variable (e.g., mild, moderate, or severe

split) by comparing the behaviors or responses of

individual family members. Future research that

compares families in their development and patterns

of split alliances is the one possible next step in this

line of inquiry.

Another important next step is to examine the

extent to which therapists can detect split alliances.

Therapists and external judges may see the alliance

similarly because they must both rely on observable

behavior to assess family members’ levels of emo-

tional connection. However, given their direct and

subjective involvement with the family, therapists’

perceptions may be more similar to the clients’

experience. On the other hand, therapists’ percep-

tions may not be accurate if they have strong

personal reactions to the family that interfere with

their ability to discern their clients’ feelings correctly.

Finally, what about individual therapist effects? All

but one therapist in each sample experienced a split

session. With future research, we may discover

whether split sessions have more to do with therapist

behavior or with preexisting differences within fa-

milies (e.g., varying motivation levels, relational

capacities). Given the frequency of therapists with

split sessions, we suspect that splits have more to do

with what each family brings to treatment, especially

because previous researchers found that splits could

be predicted from couples’ levels of marital (Kno-

bloch-Fedders et al., 2004; Mamodhoussen et al.,

2005), individual (Mamodhoussen et al., 2005), or

family-of-origin (Knobloch-Fedders et al., 2004)

distress.

When we consider all the reasons why clients’

feelings toward their therapists may not be reflected

in behavior, it is important to know that a fair

number of split alliances can be detected from in-

session behavior. When becoming aware of a split,

therapists should not lose sight of two points. First,

the level of severity may not be apparent. It is likely

that some family members privately feel more

extremely positive or negative than they appear.

Second, the reason for clients’ discrepant feelings is

likely to be different for every family. In one case, the

therapist may be favoring one client over another,

leading the latter to feel resentful; in another case,

there may be a ‘‘hostage’’ who is merely paying lip

service to the therapist to avoid recrimination from

other family members. Discerning the cause of a

split alliance may shed light on hidden family

patterns that, if attended to within the therapy,

may well also improve the family’s private life.

Acknowledgements

This article was presented at the 2008 Annual

Conference of the Society for Psychotherapy Re-

search in Barcelona. We are grateful to the many

research assistants who rated videotapes and to

Shaina Bernardi and Laurie Heatherington for their

helpful comments on a previous version of this

article.

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